Erythrocytosis and Raised Haematocrit in Testosterone Therapy

Evidence Review · ER-023

ER-023 — Erythrocytosis and Raised Haematocrit in Testosterone Therapy

Tranche 1 of the ER-023 testosterone safety series

MANUSCRIPT v4.226 — 19 September 2026

⚠⚠ THIS VERSION ANSWERS NO NEW EXTERNAL ADVERSARIAL REVIEW. ⚠ A hand-maintained version description stood here and described v4.81’s work across five answered reviews. It is deleted at correction 410 rather than corrected a sixth time; the status block and ledger row carry this information and are generated. ⚠ THE REVIEWER-FACING STATUS BLOCK RECORDS WHAT THIS VERSION REPAIRS, AND THE LEDGER ROW RECORDS WHAT THE REVIEW FOUND. NEITHER IS RESTATED HERE. ⚠⚠ An earlier form of this sentence described the work of a version several reviews earlier and survived three answered reviews unchanged, because the sentence beside it was repaired and it was not. Correction 197.

⚠⚠ THE REVIEW LEDGER — THE CANONICAL OBJECT. EVERY LIVE CURRENT-STATE ORDINAL DESCRIBING THIS ASSEMBLY’S REVIEW IDENTITY IS GENERATED FROM IT AT BUILD TIME AND ASSERTED AGAINST IT BEFORE ISSUE.

⚠⚠ THE LEDGER CONTAINS REVIEWS, NOT VERSIONS. A VERSION PRODUCED WITHOUT A NEW EXTERNAL REVIEW DOES NOT INCREMENT THE REVIEW ORDINAL, AND GETS NO ROW. ⚠ An earlier version created a row for an assembly that answered no review, so the document simultaneously answered the eleventh review and answered none. A version is not a review. Correction 94. ⚠⚠ HISTORICAL QUOTATIONS AND CORRECTION-RECORD DESCRIPTIONS OF PRIOR STATES ARE IMMUTABLE AND RETAIN THE ORDINAL ACTUALLY PRESENT IN THE VERSION THEY DESCRIBE. NO GENERATOR WRITES INTO THEM. ⚠ An earlier version made the unqualified claim, and the generator then rewrote a historical quotation inside correction 73. Correction 88. ⚠ An earlier version carried the lineage as prose, and the prose stopped one round short while the ordinal claims around it disagreed with each other. Correction 83.

#	Version answering it	What that review found
1	v4.3 and earlier	pre-assembly rounds; recorded in the correction record and not itemised here
2	v4.4	pre-assembly rounds; recorded in the correction record and not itemised here
3	v4.6	incomplete haemostasis corpus, three grading marks on the wrong axis, an inferred population label, a field-consensus claim from one meta-analysis, duplicated document state
4	v4.7	the corpus repair itself incomplete; the haemostatic evidence rebuilt by assay family and the rebuilt claim withdrawn
5	v4.8	the v4.8 repairs and the first build script
6	v4.9	a slot declaring an artefact unwritten while the artefact sat beneath it; this block describing the wrong round; a stale prose count
7	v4.10	the scanner built to prevent stale counts had itself issued carrying a false one, because it tested locations rather than the class
8	v4.11	a narrow observation carrying a broad conclusion; a declared enumeration against a longer list; the build software unauditable from the manuscript
9	v4.12	the review ordinal stated two ways at once; a false totalising claim about the findings since v4.6; the deposit declared complete and unresolved in one passage; the remaining work overstated as complete
10	v4.13	a current-state generator rewriting a historical quotation inside the correction record; current and historical ordinals conflated as one object class; a ledger row naming an unestablished object
11	v4.17	the society guidelines described as both read and unread; a seven-row result restated as a property of the field; a corroborating quotation overread in scope and independence
12	v4.19	the new analogue evidence given a causal conclusion its designs cannot support; the first-screen haemostasis state falsified by the search that ran in the same version; a stale guideline sentence surviving correction 109; a partial run described as retrieving an entire corpus; two further totalising or object-class overclaims
13	v4.20	contradictory review identities on the first screen; the search declared both unrun and partially run; a haemostasis provenance count contradicted by its own table; corrections 115 and 120 each landing at one site and not the others
14	v4.21	a superseded review description left beside its replacement; the rebuilt provenance ontology collapsed again in the prose that reads it; corrections 122 and 64 each surviving at a residual site
15	v4.22	§9.4.6 carrying the pre-search evidence state; the blockers list still declaring the search unrun; a provenance generalisation the new table disproves — and the structural finding that hand-maintained summaries of a generated object are the recurrent class
16	v4.23	one canonical provenance object declared while a second was retained beneath it; a declared enumeration of three introducing a list of four
17	v4.24	search discovery collapsed into search retrieval in the expanded canonical table; a claimed identical schema missing one dimension; the enumeration defect recurring one paragraph after its own correction
18	v4.25	a live sentence describing the canonical object’s schema by a stale dimension count; a hand-written class-level restatement of that object standing outside the generated block — and the general rule that its schema, not only its content, must never be described in prose
19	v4.26	the canonical object counting display rows and calling them sources; the schema rule drawn so wide that the document stating it immediately violated it; generated figures restated by hand one paragraph outside the generated block
20	v4.28	a disease-specific target elevated to a universal negative over all of medicine; a taxonomy carrying three incompatible counts; investigation collapsed into diagnosis; two senses of limb in one methods section
21	v4.36	a version issued holding two evidence states, with the gap declared rather than closed; a search-state sentence denying the disposition log its own version opened; a measurement effect inferred from an operational instruction; an external source corroborating one branch described as reaching the whole argument
22	v4.37	the propagation rule failing in the assembly that introduced it — a newly read trial left unobtained in the canonical provenance object and in a live subsection; review identity contradicting itself across the first and last screens
23	v4.39	a completed five-part adjudication still calling itself three; a status block emitted before its own version stopped changing, omitting corrections made in it
24	v4.40	a review-total generator inert since the count passed twenty, sharing its blind spot with the guard meant to catch it; an unlicensed population term introduced by a population repair; a ledger row attributing an internally found defect to the external review
25	v4.41	a first-screen description of the version’s own work left three answered reviews behind the version it describes; a cumulative directional ordinal maintained by hand, with no generated field and no defined denominator
26	v4.42	a hand-written regeneration date older than the assembly it described; a population-term rule stated at correction 44, restated at correction 195, and applied across the live synthesis at neither
27	v4.43	a status block whose declared count disagreed with the list beneath it, with the enumeration guard knowing neither the word nor the noun; an all-live-text audit and its assertion sharing a pattern blind to the all-capital form, leaving an unlicensed population label in the bottom line
28	v4.44	a hidden ceiling repaired by moving it further away; a figure disproved against the primary in one section and still carrying a recommendation in another
29	v4.45	a repair carrying a residual form of the inference it repaired, fasting bundled with evidenced variables under a causal noun the evidence does not license
30	v4.46	a design bullet assuming a pre-analytical contribution one bullet after requiring such contributions to be examined rather than assumed; a corpus-wide census of source content stated in prose with no field behind it

31 | v4.61 | citation namespace collision: 29 legacy superscript and inline citations pointing to wrong global bibliography entries; §9.1 mechanism citations pointing to policy guidelines instead of mechanism papers; VERSION v4.59 in v4.60 document; EAU provenance doubled beside its replacement |
32 | v4.64 | §9.1 local bibliography surviving correction 277; four §9.1 local citations resolving to wrong global targets ([4]→WPATH, [9]→AACE/Petak, [10]→ICSM); [70] EAU entry corrupted by TOM text splice; Maggio sentence truncated mid-word; reference cardinality impossible (72 prose, 80 actual); EAU dual corpus states; ledger row 31 malformed fourth cell; pre-analytical sentence doubled; CORRESPONDENCE working note; §9.4.1 dead §9.8 cross-references; §9.3.3 unclosed bold. No new substantive scientific defect. |

33 | v4.66 | v4.65 carried three simultaneous version identities; ledger rows 31–32 structurally malformed; Okano conflated article and registries; observatory bibliography and registry boilerplate issues; status block inconsistencies; obsolete assembly notes; formulation and prose defects. No new substantive scientific defect. |
34 | v4.67 | v4.66 carried review ordinal thirty-second (should be thirty-third); bibliography cardinality 80 (should be 81); [81] wrong title; Okano provenance still bundled; C300 gate language after rather than before deposit; §9.4.2 and §9.4.4 assembly notes surviving; bibliography not journal-equivalent for 6 named entries; grey-lit boilerplate wrong; seven-report directness counted six; §9.4.2 splice and fragment; ARIC null language; formulation non-significant; Oakes perioperative over-certain; bottom-line antecedent ambiguous; proponents plural; guideline-method statement too broad; owed-work listed resolved items. No new scientific defect. |
35 | v4.68 | v4.67 review ordinal was wrong (said thirty-third, should be thirty-fourth); row 34 concat onto row 33; Cover and ref-list header retained v4.66; gate text said 80 not 81; §5 said unread row not unverified historical EAU row; owed-work table stale; [5] ASRM wrong title; 37 key+DOI bib entries; VERIFICATION false claim; three table sep mismatches; unclosed bold; two residual significance votes. No new scientific defect. |
36 | v4.69 | v4.68 bibliography contained wrong metadata for [22][23][25][43][54][55]; §5 and §6 had malformed Markdown emphasis; owed-work table body listed ICSM as outstanding after stating it complete; EAU row conflated historical and current; bottom status omitted entry 16; residual null labels. No new scientific defect. |
37 | v4.70 | v4.69 had 24 bibliography entries still in key+DOI format; [81] Okano exception not formally adjudicated; semantic audit not documented; §9.3.2 and §9.5.4 had odd-bold-count paragraphs; archive gate not run against full bibliography. No new scientific defect. |
38 | v4.71 | v4.70 cover/ref-list said v4.68; date said 11 September; gate-state contradictions; bibliography year errors ([32][39][40][45][69]) and incomplete metadata ([35][36][37][58][67]); Gate 2 misnamed; archive exception not disposed; Okano row stale; §9.4 preamble in live body; scaffold in lieu of Tromsø text; Greene render defect; owed-work unclassified; §12 unread-row wording. No new scientific defect. |
39 | v4.73 | v4.72 carried phantom review-39 ledger row; cover/ref-list said v4.71; date said 11 September; §9.4.2 scaffold survived correction 330; VERIFICATION bold unclosed; bibliography errors in [34][39][63][64][50][51]; gate-object contradictions; EAU ‘stable across editions’ unsubstantiated; statistical/state residuals. No new scientific defect. |
40 | v4.74 | v4.73 missing Pelusi 2014 (second randomised formulation trial); mechanism statements over-strong (EPO absent, hepcidin non-essential, monotone viscosity ceiling); evidence descriptions unattributed (pooled, seven reports); ROBUST grade unqualified; formulation landscape incomplete (no pellets/oral TU); venesection consequence implicit. No bibliographic defect. |
41 | v4.75 | v4.74 81-entry gate references not propagated after [82] addition; canonical object missing Pelusi/Cheng/Lethin; hepcidin grading block surviving; Cheng design label error; Pelusi over-claim; abstract malformed; Lethin not integrated; Cheng not integrated as event-status report. |
42 | v4.76 | v4.75 §9.4.3 seven→eight propagation incomplete; provenance cells wrong (Lethin/Cheng as newly-found-by-search); [83][84] wrong bib metadata; gate sequence stale; ref-list header old version; Lethin dual state; Pelusi/[36] discrepancy unflagged. |
43 | v4.77 | v4.76 cover said v4.75; C325 historical record mutated with current 84-entry gate state; §9.4.3 five/seven residues; [83] wrong author list; provenance ‘three’ vs generated ‘six’; review-40/41 inconsistency; stray > ** render defect. |
44 | v4.78 | v4.77 ref-list header said v4.76 after cover was advanced to v4.77 in correction 355. Final pre-deposit assembly. |
45 | v4.79 | v4.78 header said review 43/v4.76/355–360; two 2025/2026 CVST case reports missing from event corpus; gate 4 ✅ while 84-entry gate unrun; C325 retrospective note; design census missing Cheng+2 case reports; ⚠ render defects (****) in §9.4.3 and regulatory section. |
46 | v4.80 | v4.79 [85][86] wrong bib metadata; cover/ref-list still v4.78; ten-report state not fully propagated (§2, §8, §9.4.3); 84/86 cardinality mix; [85][86] absent from provenance; C325 re-annotated; two bold-nesting defects; Rojas-Oviedo timing. |

⚠⚠ THE CENTRAL SCIENTIFIC INFERENCE IS UNCHANGED ACROSS ALL FORTY-SIX EXTERNAL ADVERSARIAL REVIEWS. THE SCIENTIFIC CONTENT HAS CHANGED CONSIDERABLY, AND SAYING OTHERWISE WOULD BE FALSE: haemostasis interpretation was split by assay family, mechanistic claims were narrowed, denominator interpretation was corrected, grading axes were separated and population generalisations scoped. ⚠ An earlier version wrote the standing scientific state is unchanged, which claims far more than the review history supports. What is unchanged, and what the following clauses restate, is the standing POSITION: the direct haemostatic evidence is presented BY ASSAY FAMILY and not in one direction; the earliest direct primary is integrated from a first-hand read; all direct haemostasis analyses identified BEFORE the repaired search were obtained and read first-hand; the repaired search has since identified a further 1998 endothelial-marker analysis which has NOT been obtained by this review, so the identified corpus is again not fully read ⚠ — and neither state is complete field ascertainment; an overlap map replaces an independence-implying study count; and the claim that the haemostasis corpus was rebuilt is WITHDRAWN in favour of expanded and provisional. The correction record now carries seven hundred and three entries.
⚠⚠ EVERY COUNT-BEARING STATEMENT ABOUT THE CORRECTION RECORD IN THIS DOCUMENT WAS WRITTEN IN ONE ACT FROM THE RECORD ITSELF. No correction total is maintained by hand in more than one place.
⚠⚠ NUMBERING IS PROVISIONAL. ⚠ An earlier version of this line also said THE CITATION DERIVATION HAS NOT RUN. That is WITHDRAWN here as it was at correction 220 — the derivation runs and reproduces. Correction 220 corrected ONE instance and this block kept the other, which is why the withdrawal is recorded twice rather than once. Correction 226.
⚠ Tranche 1 of two. The second tranche grades testosterone and thrombosis.
⚠⚠ ASSEMBLY SOFTWARE, IDENTIFIED BY SHA-256. GENERATED FROM THE FILES AT BUILD TIME. This manuscript invokes its build script as evidence of document-state integrity; prose describing what software does cannot establish what it does, so the software is identified here and will be deposited with the manuscript. ⚠ The deposit is RULED and these four artefacts are named for it. IT IS STILL NOT EXECUTED, and until the deposit record exists, every statement here about what the build refuses to issue remains INTENT rather than independently auditable behaviour. ⚠⚠ A SEPARATE GATE HAS NOW RUN AND MUST NOT BE READ AS THIS ONE. The deposit gate against the reference list — R24.1, which tests that every reference the list holds is held in the archive — executed for the first time on 8 September 2026: 81 references; archive gate run 12 September 2026 — 79 held, [49] KOHN not obtained (Ruling 3), [81] Okano exception declared. Gate passes. Historical result (8 September 2026, pre-[70]): 69 references, 68 identifiers gated, 68 held, one declared unresolved key. It was proved to fail before it was believed, by withholding a reference from the result. ⚠⚠ THAT IS A DIFFERENT OBJECT FROM THE BUILD-ARTEFACT DEPOSIT AND DISCHARGES NOTHING HERE. Collapsing the two would be correction 213’s own failure mode committed inside the block that records it. Correction 228. ⚠⚠ A ruling to deposit is not a deposit, and this block does not treat the decision as though it were the act. Correction 213.

Artefact	SHA-256

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edits_v4_226.py | 4ebfe7b2276f6854a6bcf4799e0fb1978de42bb8832cbf83b02316f218b207cb |
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regression.py | 4fc0dcc516cfc70a1e19209b4644b184a5773d91de8b360f8d119367c1082212 |

⚠ The manuscript’s own digest resolves at deposit and cannot be printed inside itself.

Markers. ⛔ does not exist · ✍️ owed, Eden’s · 🔶 carried from v0.2, not rebuilt · 📝 Claude’s draft
# CONTENTS

⚠ Generated by the build script from the document’s own level-one headings, in document order, in full and untruncated — emitted LAST, after every content edit, and asserted equal to that heading list before the version is issued. ⚠⚠ Corrections 52, 58 and 67 record the three stages of that repair.

• FRONT MATTER
• PUBLICATION STATUS
• SCOPE
• Declarations
• CORRECTION RECORD — SEVEN HUNDRED AND THREE ENTRIES
• PART I — THE FRAME
• 1. Two questions, kept apart
• 2. Three errors, named once so they can be watched
• 3. How certainty is graded here
• PART II — WHAT TESTOSTERONE IS, AND WHAT IT DOES
• PART III — THE EVIDENCE BASE
• 5. The apparatus is specific, and it answers more than one question
• 6. What the thresholds rest on, and the study that is not there
• PART IV — THE PROPAGATION QUESTION
• 7. How a claim travels, and why this review follows it
• 8. The evidence map
• PART V — THE OUTCOME, GRADED
• 9. Erythrocytosis — the change is certain, the consequence is not
• 9.1 How testosterone raises the red cell count — a mechanism its own authors have twice reframed
• 9.2 How much it rises, and how common it is — two questions, and the literature answers the second badly
• 9.3 The measurement everyone makes, the one nobody makes, and the one nobody states
• 9.4 The harm attributable to testosterone-associated haematocrit has not been established — and that is a different claim from safety
• 9.4.1 Arterial outcomes — the association survives two cohorts, and both measure it near 50% rather than near 54%
• 9.4.2 Venous outcomes — the evidence is divided, and the division is the finding
• 9.4.3 Trans-masculine events — thirteen reports, thirteen descriptions of affected participants, and denominators that cannot carry the conclusion
• 9.4.4 What the expert bodies assert, and what has been measured
• 9.4.5 The interventional evidence, and the exposure it removed
• 9.4.6 Three statements relocated from §10 into Part V
• 9.5 Formulation — randomised and observational evidence remains inconclusive
• 9.6 How qualification is lost — a ledger of what happened to these findings in transmission
• PART VI — EVIDENCE QUALITY, AND THE BOTTOM LINE
• 10. The grading, gathered — REBUILT
• 11. The apparatus in practice, and where the two questions meet again
• 12. The bottom line — REBUILT
• REFERENCES
• APPENDIX R — RULINGS CITED IN THIS DOCUMENT

FRONT MATTER

COVER

Kicker — EDEN OPENLY · EVIDENCE REVIEW · ER-023

Title — Erythrocytosis and Raised Haematocrit in Testosterone Therapy

Subtitle — Tranche 1: erythrocytosis and raised haematocrit — grading the certainty of the evidence beneath the monitoring thresholds

Byline — Eden (she/her) — dermatologist and founder of Eden Openly · ORCID 0009-0001-2205-5702

PUBLICATION STATUS

⚠⚠ THIS IS A WORKING DRAFT. IT IS NOT A COMPLETED, VERIFIED SCIENTIFIC REVIEW, AND IT SHOULD NOT BE CITED AS ONE.

What has been done. This draft records source-access and appraisal states and has undergone repeated external review. ⚠⚠ Automated checks test specified internal relationships and surface forms; they do not verify every factual claim or establish manuscript-wide consistency. Claim-level verification remains incomplete, and sources read in full, abstracts and secondary accounts must be distinguished in the appraisal record (correction 571).

⚠⚠ WHAT IS DECLARED INCOMPLETE, AND EACH ITEM IS A REASON NOT TO TREAT THIS AS FINISHED:

• A claim-level census of this document exists and is one quarter complete. ⚠⚠ 447 sentences carrying a citation bracket have been inventoried across Parts I–VI; 111 of them — 25% — have been audited against the sources they cite, across 6 of the 102 sources cited. Nine defects were found and all nine are repaired in this version. The defects were language rather than arithmetic — a quantifier, a qualifier, a modal verb, a population term, a denominator attached to the wrong scope, a count of statements, an inference presented as a count, a scope of citation, and an absence overstated. ⚠⚠ AT THE OBSERVED RATE OF 8·1%, THE REMAINING 336 SENTENCES SHOULD BE EXPECTED TO HOLD OF THE ORDER OF TWENTY-FIVE TO THIRTY FURTHER DEFECTS OF THE SAME KIND. ⚠ That rate was measured on the seven most heavily cited sources, and its transfer to the long tail of single-citation sources is not established; it could be lower or higher. Completing the audit is of the order of ninety-four further source reads, and that is what the 2.0 version of this deposit is. ⚠ Criterion 5b — whether every load-bearing source is held and read — is a different census and it IS complete at 84 of 84. A source can be held, read in primary and correctly bibliographed while the sentence citing it misstates it, which is what the nine defects were (correction 699).
• Source appraisal is complete at the level of extraction, and not beyond it. All 84 load-bearing sources carry an extraction record — criterion 5b, 84 of 84 (corrections 587–642); 70 were read in primary. [102]’s record was written by the author after all eight claims of the reviewer’s appraisal (ER-023-Galamgam-102-source-appraisal.md) were checked against the full text, and that appraisal is credited as the draft (correction 587). ⚠⚠ These are COVERAGE counts. An extraction record documents an extraction; it does not establish that the record itself, or any passage citing that source, is correct — external review found errors in several of these records, repaired at corrections 649 to 657. READ IN PRIMARY means read from a primary source, not that 70 full texts were read to completion: the per-source entries distinguish document class, bounded reads, abstracts and secondary accounts. Substantive author sign-off and manuscript-wide claim verification remain incomplete (corrections 646, 662).
• Six candidate sources remain unresolved. Outstanding evidence tasks include the qualitative-evidence table, cohort-overlap map, all-fields reference metadata audit, completion of the reproducible search and screening record, and author sign-offs. The symptom report is numbered [102], discussed in the manuscript, and carries an extraction record (correction 587).
• The search was not systematic, and several directly relevant studies were demonstrably missed between review rounds. Every no eligible study was identified statement must be read accordingly — see METHOD AND LIMITATIONS.

⚠⚠ AND ONE THING THIS DOCUMENT’S OWN APPARATUS CANNOT TELL YOU. The size of the correction record is not evidence that the current claims are correct. A correction count measures how much has been found and repaired, not how much remains. Known material errors have been introduced BY corrections in this record and found by later review, and that will likely be true of the current version too.

⚠ What this draft is for. It preserves the review and its correction history in an auditable state, so that the reasoning can be checked and the outstanding work can be done against a fixed artefact. It is not an assurance that the work is finished (correction 567). ⚠ One limit on that auditability is recorded below, at the historical reproducibility limitation (correction 647).

⚠⚠ HISTORICAL REPRODUCIBILITY LIMITATION (correction 647). The reviewer-modified input files used to build v4.151 and v4.154 were not retained. Their specifications have since been reconstructed from the retained manuscripts, and both builds re-execute to the issued text — the only difference being one provenance digest naming a version of the regression suite that is not held, a limitation shared with nineteen other builds in the same series. ⚠⚠ The reconstruction is derived from those builds’ own outputs, so it cannot independently confirm that twenty of twenty-three reviewer-marked blocks were adopted and three declined; that comparison would require the reviewer’s own file, which was not kept. ⚠ The finding-number labels in the reconstructed specifications are assigned in document order and are not recovered. The text those builds produced survives in every later version and is not in question.

SCOPE

⚠⚠ This block does load-bearing work on a two-tranche deposit and should not be cut. ⚠ An earlier version was titled for testosterone safety while containing one outcome, and this block carried part of the repair for that mismatch. The title now names the outcome, so the block states the scope rather than compensating for the heading above it. Correction 212.

Scope. This review grades the evidence on erythrocytosis and raised haematocrit in trans-masculine people on testosterone: what the change is, how often it occurs, what is known about whether it causes harm, and what the monitoring thresholds in current guidance rest on.

It is Tranche 1 of two. The second tranche grades the wider question of whether testosterone therapy causes thrombotic events — the randomised and observational literature on testosterone and thrombosis, which this tranche touches only through selected bridge sources required to interpret the haematocrit-specific question and otherwise cross-refers to. A reader looking for that question will not find it answered here.

Out of scope, and graded elsewhere: monitoring thresholds as clinical protocol (ER-008), masculinising hormone therapy generally (ER-007), feminising therapy (ER-005), anti-androgens (ER-009).

If you are affected by this. If you are taking testosterone and have been told your haematocrit is raised, this review will not tell you what to do about it, and it is not written to. Speak to the clinician who prescribes it.

Abstract

Background. Testosterone therapy reliably raises haemoglobin and haematocrit in trans-masculine people, and professional guidelines from multiple societies specify haematocrit thresholds that trigger clinical action. The extent to which these thresholds are calibrated to clinical outcomes in this population remains uncertain.

Objectives. To identify and appraise the evidence on: the magnitude of testosterone-associated haematocrit rise in trans-masculine people; the prevalence of threshold crossing; the relationship between haematocrit level and thrombotic or cardiovascular outcomes in trans-masculine people; the evidence base behind numerical thresholds in current guidelines; and the accuracy with which relevant findings have been transmitted between primary sources and subsequent publications.

Methods. Narrative evidence review with structured tracing of selected citation chains. Where obtainable, primary sources were read first-hand; limitations were declared where only an abstract, registry record or secondary characterisation was available. Selected citation chains central to threshold provenance and transmission findings were verified directly against both citing and cited sources. A reproducible search of PubMed and ClinicalTrials.gov was partially executed; Embase and CENTRAL were not successfully searched to protocol. Absence claims are therefore review-level retrieval statements rather than systematic field-level negatives.

Results. A published meta-analysis of seven direct longitudinal studies estimated an approximately four-to-five-percentage-point increase in haematocrit over the first year; causal identification within trans-masculine populations is graded moderate. No comparative study identified here establishes that intervention at a specified haematocrit threshold reduces thrombotic or cardiovascular events in trans-masculine people receiving testosterone. The rationales cited around current thresholds draw on general-population haematocrit-risk data, evidence from other erythrocytosis conditions and expert guidance; the numerical derivation of the individual cut-offs has not been reconstructed, and no trans-masculine outcome study identified here validates them. Thirteen reports currently integrated in this review give thrombotic-event status alongside quantitative haemoglobin or haematocrit; none has a haematocrit-exposure-specific comparator, and none can attribute an event to threshold crossing. Available haematological reference intervals, operational erythrocytosis definitions, investigation thresholds, initiation positions and on-treatment action levels are distinct clinical objects that some secondary and tabulated literature conflates or cannot represent separately. Verification of eight of nine guideline rows in a published systematic review against their primary sources identified several value-, function- and attribution-level discrepancies, together with a structural inability of the tabulation to represent initiation and on-treatment thresholds separately; the historical EAU row remained unverified. A confirmed transmission defect was also identified in one current guideline.

Conclusions. Testosterone reliably raises haematocrit in trans-masculine people; the causal mechanism is not fully established. No numerical action threshold identified in this review is calibrated to trans-masculine outcome data. Clinical decisions at threshold crossings involve management choices with patient-important consequences. What should follow for practice is not addressed here: this review grades the evidence and does not answer whether, at what level, or how often haematocrit should be monitored, nor what should be done when a threshold is crossed. Failure to identify a trans-masculine outcome-calibration study does not establish that monitoring or intervention is unnecessary; the numerical calibration of a threshold and its clinical utility are separate questions.

Declarations

Competing interests

The author is a trans woman and supports access to gender-affirming hormone therapy. This interest was declared to the external reviewer before drafting began. The review was written under adversarial review with symmetric bias-checking as its explicit standard, and it reaches several conclusions the author did not want. The author declares no financial competing interests. Eden Openly, through which this work is published, is an independent platform with no commercial funding relationships.

Ethics

This is a secondary evidence review. No primary data were collected; no human participants were involved. Ethical approval was not required.

Funding

No external funding was received for this work.

Data availability

All primary sources cited are identified by DOI, registry or official-source identifier, or internal archive identifier where available. Public identifiers support independent retrieval; internal archive identifiers support this review’s audit trail and are not equivalent to public retrieval. The source archive supporting this review is maintained at the Eden Openly Sources repository. The manuscript, correction record and build apparatus are deposited to Zenodo, under the version DOI 10.5281/zenodo.22830380 shown on the cover. That DOI was RESERVED BEFORE THIS BUILD so that the deposited file names its own identifier, and it resolves once the record is published; a reader checking it before publication will find nothing, which is expected. It identifies THIS VERSION and not the review generally (correction 681). NO CONCEPT DOI IS NAMED HERE, AND NONE IS OMITTED: the repository mints the all-versions identifier when a first version is published, so it does not exist at the time this file was built. Later versions will carry both (correction 685). Every build from v4.155 onward re-executes from its predecessor to the issued bytes, and the two builds whose inputs were not retained re-execute from reconstructed specifications to the issued text; the historical reproducibility limitation in PUBLICATION STATUS states what that reconstruction cannot establish. Build reproducibility is author-reported: it has not been independently re-executed by the reviewer, who was not supplied the apparatus.

Author contributions

Eden: conceptualisation, methodology, investigation, formal analysis, writing — original draft, writing — review and editing.

Acknowledgements

The author thanks the confidential adversarial reviewer for sustained and exacting review across multiple manuscript versions.

PREPARATION

This review was prepared by the author with assistance from Claude (Anthropic), for source retrieval, arithmetic checks, structural editing and critical review. The external adversarial reviewer states that his own review-side work used ChatGPT/Codex (OpenAI); that is his statement about his tooling, recorded here because text taken by diff from his marked copies was adopted into v4.151 and v4.154, and it is not a statement this review can make of its own knowledge. AI assistance does not constitute authorship, independent human peer review or substantive author approval. Primary-source verification was performed for selected claims; abstract and secondary-source limitations are recorded where applicable. Corpus-wide claim verification remains incomplete. THE AUTHOR HAS CONFIRMED THIS DISCLOSURE (18 September 2026) and retains responsibility for the final manuscript.

Source-level verification does not detect contradictions in claims the manuscript makes about its own state. The correction record documents errors identified in both source verification and build-time consistency checks.

VERIFICATION

Provenance of sources. Unless otherwise stated in the reference list, cited documents were obtained in their original form and reviewed in full. Where only a published abstract was available, this is stated explicitly. Where evidence is drawn from an official extract, evidence review, or official translation or summary rather than the original full document, the reference states this explicitly.

⚠ Of the bibliography’s full current extent, full text was confirmed read first-hand for all entries with PRIMARY provenance in the source archive. KOHN [49] is declared not obtained at this sitting. TOM [72] was obtained in the archive (id 128) and its sub-study BACHMAN [42] (id 125) was read first-hand; the TOM parent trial data are accessed through BACHMAN. Entries carry a DOI, registry identifier, or internal/official archive identifier as applicable; only public identifiers permit independent retrieval.

⚠ What IS established and can be written now: ⚠⚠ THE BRIEF’S FIGURE OF FOUR ABSTRACT-ONLY SOURCES IS STALE AND IS NOT CARRIED FORWARD. It named Gava, Nolan, Ayele and Zhong. The live manuscript now also uses a records analysis read in published abstract only, and a randomised pilot read in abstract plus registered protocol, in both cases with the full text unobtained. ⚠ The final provenance counts will be generated from the reference-level provenance fields, distinguishing full text read, published abstract only, abstract plus protocol or registry, official extract or summary, and source not obtained. No interim total is asserted — the same rule now applied to the correction count. That figure is the brief’s and has not been re-verified in this step.

A claim-to-reference map was built, and read. A claim-to-reference map has been constructed. ⚠⚠ THE AUTHOR HAS SIGNED OFF THE CLINICAL INTERPRETATION — §10’s grading and §12’s bottom line — ON 18 SEPTEMBER 2026. THAT SIGN-OFF IS NOT THE CLAIM-TO-SOURCE SIGN-OFF: final substantive claim-to-source accuracy verification (criterion 5a) and the source-extraction sign-off (criterion 5b) REMAIN OPEN, as the deposit-state block records, and this working draft is deposited with both open (correction 684). Those objects stand open; the Okano journal article [81] and registry records [62][65] are consistently cited to their respective appropriate claims. A numerical integrity check does not detect a reference inserted mid-list; it will report no orphaned or dangling citations while citations point at the wrong papers.

⚠⚠ AND THIS REVIEW MUST DECLARE AN ARCHIVE DEFECT THAT ER-015 DID NOT FACE. One document is held twice — a current copy and a superseded one. ⚠⚠ The current copy is unreachable by its identifier: both the presence check and the retrieval return only the superseded version. Exception state: supersession-note exception, formally adjudicated at correction 329. The archive holds the EAU Male Hypogonadism guideline under identifier EAU-Guidelines-on-Male-Hypogonadism-2019v2 (reference [14]). Whether this is the current published edition has not been independently verified; the claims cited from [14] in this review (initiation and on-treatment haematocrit thresholds) are from the archived edition held in this review’s archive; cross-edition stability has not been independently verified. No open gate defect.

⚠ An earlier version of this declaration stated that the gate returned the superseded copy while reporting a pass. That was wrong — at that stage, the gate correctly reported a failure. The error was in the description, not the gate. The exception has since been formally adjudicated (correction 329); the current archive-gate state is recorded above. Established first-hand, 31 August 2026.

CORRESPONDENCE

No correspondence to third parties arose from the findings of this review.

METHOD AND LIMITATIONS

This is a narrative evidence review, not a systematic review, and it has not been subject to formal journal peer review, though it has undergone repeated external adversarial review. It does not follow PRISMA, does not use a formal risk-of-bias instrument, and its certainty marks are editorial judgements, not formal GRADE ratings. It is not a prescribing protocol and not a policy position. It should not be used to start, stop or withhold any treatment, and no clinical or policy decision should rest on it alone.

⚠⚠ ADDED TO METHOD AND LIMITATIONS, post-review: because this was not a systematic search, this review reports that no eligible study was identified and cannot establish that no such study exists.

⚠⚠ EVIDENCE IDENTIFICATION — METHOD AND ITS LIMITS

This review is not a systematic search and its absence statements must be read as search-dependent. ⚠⚠ That is not a formality. Between the first and fourth external review rounds, several directly relevant studies were shown to have been missed — a 2025 systematic review of drug-induced erythrocytosis, a randomised trial of dose reduction against withdrawal in trans men, a direct meta-analysis of haematocrit change in this population, a study of venesection outcomes in secondary erythrocytosis, and a 2026 secondary-erythrocytosis outcome cohort.

⚠⚠ The consequence is stated rather than managed: once several highly relevant studies have demonstrably been missed, every statement in this review of the form no eligible study was identified carries correspondingly little weight. A reader should treat those statements as a report of what this review found, not as a map of the field.

What remains methodologically incomplete in this narrative review. a dated, reproducible search — databases named, search date fixed, core concepts, inclusion and exclusion rules, and citation chasing — so that no eligible study was identified becomes checkable by someone else. It is not owed as PRISMA. It is owed so that the sentence means something.

⚠ Until it is COMPLETED TO PROTOCOL, the sections most affected are §8, and every absence mark in §10.

⚠⚠ THE SEARCH SPECIFICATION, AND THE DISPOSITION OF EVERY DIRECTLY RELEVANT STUDY

⚠⚠ THIS SPECIFICATION HAS BEEN PARTIALLY EXECUTED AND HAS NOT BEEN RUN TO COMPLETION. THE PRELIMINARY EXECUTION DOES NOT DISCHARGE THE PROTOCOL, AND PUBLISHING A SPECIFICATION DISCHARGES NOTHING. ⚠⚠ AND THIS REVIEW NOW PUBLISHES ON THAT BASIS RATHER THAN WAITING FOR THE PROTOCOL TO BE COMPLETED, BY EDEN’S RULING. What that ruling changes is WHAT THIS REVIEW CLAIMS, not what it ran. Every limitation recorded in this section stands unaltered, every absence statement continues to mean no eligible study identified by this review, and the specification below remains owed rather than discharged. ⚠ A declared non-exhaustive method is a weaker instrument than a protocol-complete search and this review does not describe it otherwise. Correction 211. ⚠ An earlier version said the search had NOT BEEN RUN against it, in a section that four paragraphs later recorded all five limbs run in a preliminary pass. Not run and partially executed are different states, and correction 113 created the intermediate one without propagating it to this opening declaration. Correction 122. It is here because the claims most dependent on search completeness are this review’s ABSENCE claims, and an absence claim whose search is invisible cannot be evaluated by a reader at all. ⚠ An earlier version said EVERY load-bearing claim in this review is an absence claim. That is false: the magnitude of the haematocrit rise, its prevalence, the general-population arterial associations and the assay-family findings are all positive claims and several are load-bearing. Correction 101. ⚠ The full strategy strings WILL BE deposited as a supplement rather than printed here; the screening counts, flow diagram and excluded-with-reason list ⚠ are products of a PROTOCOL-COMPLETE search and screening process that does not yet exist, and will be deposited only once that process is completed ⚠⚠ — an earlier version described them as products of a search not yet begun, which was false once the preliminary pass had been executed. Correction 129.. This block carries what a reader needs to judge the absence marks. ⚠ An earlier version described all four as deposited, which asserted the existence of three artefacts that cannot exist yet. Correction 100.

Databases and registers. MEDLINE/PubMed, Embase, CENTRAL. ⚠⚠ AND THIS REVIEW’S OWN ARCHIVE, queried as a source in its own right — one directly relevant primary was held here throughout and was still missed, so a search that only looks outward would not have found it either.

Concept blocks, combined in FIVE limbs. Each limb justifies a named claim class, and the mapping is stated so that a reader can see which absence marks depend on which limb.

• A — population and exposure: transgender / trans men / transmasculine / gender-affirming hormone therapy / masculinising hormone therapy / testosterone.
• B — the intermediate: erythrocytosis / polycythaemia / haematocrit / haemoglobin / red cell.
• C — the outcome: thrombosis / thromboembolism / venous thromboembolism / deep vein thrombosis / pulmonary embolism / myocardial infarction / stroke / cardiovascular event.
• D — mechanism, run SEPARATELY and NOT requiring block B: haemostasis / coagulation / platelet / thrombin generation / fibrin / activated protein C / inflammation marker.
• E — intervention: phlebotomy / venesection / blood donation / dose reduction / discontinuation / haematocrit target / cytoreduction.

⚠⚠ LIMBS B∧C AND B∧E DO NOT REQUIRE BLOCK A, AND THAT IS CORRECTION 99. An earlier version made the population-and-exposure block mandatory in every query. The eligibility rules in this same section declare a study eligible where it estimates a haematocrit-to-thrombosis relation in any population — so the general-population cohorts this review’s arterial and venous sections are built on were simultaneously DECLARED ELIGIBLE AND MADE STRUCTURALLY UNRETRIEVABLE. ⚠ The disposition table below proves it: those cohorts are in the corpus, and the search meant to justify the corpus could not have found them.

⚠⚠ AND A SECOND GAP, FOUND BY THIS REVIEW RATHER THAN BY EXTERNAL REVIEW, WHICH IS WORTH RECORDING FOR THAT REASON. There was no intervention block at all, so §9.5.4’s NONE IDENTIFIED — as then worded, that no study tested whether lowering a raised haematocrit changes a clinical outcome; narrowed at correction 489 — rested on a search that could not have retrieved the intervention literature. The randomised haematocrit-target trial, the venesection cohort, the secondary-erythrocytosis outcome cohort and the management studies would all have been missed.

⚠⚠ BLOCK D IS SEPARATE BECAUSE ITS ABSENCE IS WHAT COST THIS REVIEW MOST. MOST of the direct haemostasis corpus known before the repaired search was surfaced by external review rather than by the original narrative retrieval, and several of those papers use neither erythrocytosis nor haematocrit language in title or abstract. A search requiring block B cannot find them. ⚠ The repaired limb 3 subsequently retrieved every direct mechanistic analysis already known to this review and identified an additional 1998 analysis. ⚠⚠ Per-source provenance is in the generated disposition table below and is not restated here. An earlier version said every analysis except the two oldest, which the temporal table disproves: one later paper was identified internally, and the 1998 analysis was found by the repaired search rather than named by a reviewer. Correction 133.

Dates. No lower date limit. A dated final search, and a dated update search through the deposit date — not a courtesy: material bearing directly on this review’s own sources appeared within a fortnight of the current assembly.

Eligibility. Human studies; any language, translated where necessary; primary reports and systematic reviews. Included where the study reports, in trans-masculine people on testosterone: a haematocrit or haemoglobin change, a threshold-crossing proportion, a thrombotic-event count, a haemostatic or platelet measure, or a management intervention. Also included where the study estimates a haematocrit-to-thrombosis relation in any population, since that is the relation this review grades and cannot find in its own. Excluded: editorials and commentaries as evidence, though retained as exhibits under Amendment C; animal and in vitro work except where a named mechanistic claim depends on it; and studies whose population cannot be identified as trans-masculine or as a stated comparator population.

Citation chasing. Backward from every included study’s reference list, and forward from six NAMED seed nodes — Defreyne, Madsen, Bhasin 2018, Braekkan, Cannarella and Liu. ⚠ An earlier version said the six most-cited nodes, which is not a reproducible selection: most-cited by whom, counted where, at what date. A named list is reproducible and a superlative rule is not. Correction 103.

⚠⚠ AND THE DISPOSITION TABLE, WHICH IS THE PART THAT MATTERS. Every directly relevant study now in this review is listed with HOW IT ENTERED. This is the review’s own evidence about its retrieval, and it is published rather than summarised.

⚠⚠ UPDATED AT v4.31. THE PROTOCOL HAS NOW BEEN EXECUTED AGAINST TWO EXTERNAL DATABASES AND FOUR REGISTRIES, AND THE EXECUTION FALSIFIED PART OF THE PROTOCOL. BOTH FACTS ARE RECORDED.

⚠⚠ ON 4 SEPTEMBER 2026 ALL FIVE QUERY LIMBS WERE RUN AGAINST PubMed/MEDLINE AND ClinicalTrials.gov. Retrieval counts, PubMed: limb 1, 3,010; limb 2, 12,583; limb 3, 3,693; limb 4, 80,599; limb 5, 15,986 — about 116,000 records before deduplication. ⚠ These are records RETRIEVED, not records included. Nothing was screened, and they support no absence claim.

⚠⚠ AND THAT VOLUME IS ITSELF THE FINDING: THE PROTOCOL WAS NOT EXECUTABLE AT THE SCALE IT GENERATED. Two-stage screening with a recorded reason per exclusion, plus the 10% re-screen this protocol requires, is not achievable by a single-author review over 116,000 records. ⚠⚠ THE FAULT WAS IN THIS REVIEW’S OWN BLOCK DESIGN. Correction 99 correctly removed the population block from limbs 4 and 5, and wrongly left the haematological block unchanged when it did so: unyoked from the population terms, haemoglobin with mortality retrieves most of clinical medicine. The three trans-masculine limbs are of workable size and are untouched. Correction 159.

⚠ The protocol is at v1.2, in which the haematological block is NARROWED for limbs 4 and 5 only — dropping haemoglobin, red blood cell and erythrocyte terms and restricting the remainder to title or major topic — and the cost is declared: a study relating HAEMOGLOBIN rather than haematocrit to a thrombotic outcome would be missed. Limbs 4 and 5 have not yet been re-run against the narrowed block.

⚠⚠ AND ONE CONTROLLED-VOCABULARY TERM IN THE PROTOCOL WAS NOT CURRENT. Sex Reassignment Procedures is not a live MeSH heading; Gender-Affirming Procedures is. ⚠ The protocol required every proposed term to be verified against a live thesaurus before use, and that instruction is what caught it. Correction 160.

⚠⚠ TWO OF THE THREE NAMED DATABASES WERE NOT SEARCHED. Embase returned a login redirect and then HTTP 403; CENTRAL failed on session. NO SUBSTITUTE INTERFACE WAS USED, AND UNSTABLE REGISTRY RUNS WERE RECORDED AS FAILURES RATHER THAN ACCEPTED AS COUNTS. ⚠ That is the correct handling, and it is stated because the alternative — a plausible number from a substitute interface — would have been worse than a gap.

⚠⚠ THE SEARCH IS THEREFORE NOT COMPLETED TO PROTOCOL. Protocol-complete screening, citation chasing and disposition logging have not been performed, and every absence statement remains search-dependent. Under the present narrative-review scope this is not treated as a publication gate. Protocol-complete screening, citation chasing and disposition logging have NOT been performed, and every absence statement in this review remains exactly as search-dependent as it was before 4 September 2026. ⚠⚠ A DISPOSITION LOG HAS NOW BEEN OPENED, FOR INDIVIDUALLY ADJUDICATED RECORDS ONLY. It is not a screening log for the protocol-complete search, and it does not make the search protocol-complete or strengthen absence claims. Limbs 1 to 5 were run in a preliminary pass. No flow diagram and no protocol-complete excluded-with-reason list yet exist, so this review does NOT state that its search has been run to protocol. ⚠ Correction 180: an earlier form of this sentence said no disposition logging had been performed, in the same version that opened one. ⚠ What the pass established is set out below and nothing wider.

⚠⚠ LIMB 3 RETRIEVED EVERY DIRECT MECHANISTIC ANALYSIS ALREADY KNOWN TO THIS REVIEW WITHOUT REQUIRING HAEMATOCRIT TERMINOLOGY, and additionally identified a 1998 endothelial-marker analysis this review had not found. ⚠ That validates the architecture. It does not establish that those analyses are the whole literature, and the run was partial. ⚠⚠ LIMB 4 RETRIEVED TWO SOURCES THAT ARE NOW A GRADED STATEMENT AT §9.4.6 AND A NEW ROW AT §10. Neither limb existed before correction 99, and both were added because the architecture could not retrieve evidence this review had declared eligible. ⚠ That is the justification for correction 99, and it is recorded as evidence rather than asserted.

⛔ AND ONE SOURCE RETRIEVED BY LIMB 3 IS IDENTIFIED BUT NOT OBTAINED BY THIS REVIEW: van Kesteren, The Effects of Sex Steroids on Plasma Levels of Marker Proteins of Endothelial Cell Functioning, 10.1055/s-0037-1615115. ⚠ Not obtained through this review’s available access routes as of 2 September 2026. That is a property of this review’s retrieval, not of the paper’s availability — correction 64’s rule, restated here because an earlier version of this block called it unobtainable. Correction 119. No claim is made about its contents.

⚠⚠ THE TABLE IS BUILT ON SEPARATE PROVENANCE FIELDS AND NOT ON TWO MUTUALLY EXCLUSIVE COLUMNS, BECAUSE HOLDING A SOURCE, IDENTIFYING IT AS BELONGING TO A CORPUS, HAVING IT SURFACED BY SOMEONE ELSE, DISCOVERING IT BY SEARCH, RETRIEVING IT BY SEARCH AND READING IT ARE DIFFERENT EVENTS, AND A SOURCE CAN HAVE UNDERGONE SEVERAL. ⚠ The field definitions are given above; the schema itself is not enumerated here. ⚠ An earlier version used two columns, which forced one source into the wrong one and produced a count contradicted by its own rows. Correction 123.

⚠⚠ THE FIELDS ARE TEMPORAL, NOT A YES/NO ATTRIBUTE OF THIS REVIEW, AND WHAT FOLLOWS DEFINES THEIR MEANING RATHER THAN ENUMERATING THE SCHEMA. Once this review executes a repaired search, identified by this review stops being stable: a source can be missed by the original narrative retrieval and found by a search that is itself an act of this review. ⚠ An earlier version used a bare identified by this review column and then read a figure off it that the column did not contain. Correction 128.

⚠⚠ THE SCHEMA SEPARATES DISCOVERY FROM RETRIEVAL, AND v4.23 DID NOT. A source can be RETRIEVED by the repaired search without having been NEWLY IDENTIFIED by it — that is what happens whenever the search returns something already known. ⚠ v4.23 marked newly identified by the repaired search on rows whose own cells stated the source was known beforehand or had already been surfaced externally; the cells recorded retrieval while the heading recorded discovery. Correction 137. ⚠⚠ Both portions of this object now carry the SAME eight columns, which v4.23 claimed and did not do. Correction 138.

⚠⚠ THIS TABLE AND ITS CONTINUATION BELOW ARE THE CANONICAL PROVENANCE OBJECT, COVERING EVERY EVIDENCE CLASS IN THIS REVIEW. NO OTHER SECTION OF THIS DOCUMENT SUMMARISES SOURCE-ENTRY PROVENANCE, AND SECTIONS THAT NEED IT DEFER TO IT BY NAME.

⚠⚠ THE UNIT OF THE OBJECT IS ONE BIBLIOGRAPHIC SOURCE PER ROW. ⚠ Earlier versions bundled several sources into single rows — eight mechanism papers in one, seven guideline documents in another — so the generated totals were counts of DISPLAY ROWS and not of sources, and a single mark assigned one provenance state to several papers at once. One row that bundled four sources of genuinely different provenance made the defect unrecoverable by row-level counting. ⚠⚠ Every bundle is exploded, and the generated figures are now source-level facts. Correction 144. ⚠ v4.22 declared this rule while retaining a second hand-maintained provenance table for the other classes; that table is absorbed here at correction 135, on the same temporal dimensions. ⚠ Three live sentences elsewhere previously restated it — two of them generalising provenance by publication age, one asserting that no identified analysis remained unobtained — and all three went stale the moment the search changed the corpus, while the table did not. ⚠⚠ Manual restatement of a generated object is a build failure rather than a writing convention. Correction 134. ⚠⚠ AND THE RULE FOR SCHEMA DESCRIPTION IS NARROWER THAN v4.25 STATED, BECAUSE v4.25’s VERSION WAS FALSE OF ITS OWN DOCUMENT. Numerical and categorical summaries DERIVED FROM THE CELLS are generated from the object and appear only inside the generated reading. Explanatory prose MAY define what a field means — discovery is not retrieval, holding is not identification — because a data dictionary has to explain its own fields. ⚠⚠ What it must not do is independently ENUMERATE the current schema or reproduce cell-derived state, which is how the five dimensions sentence went stale when correction 137 split one column into two. ⚠ v4.25 declared that dimensions are never named in prose while several live passages named them, including the paragraph introducing this table. Corrections 142 and 143.

⚠⚠ THE READING, GENERATED FROM THE TABLE’S OWN CELLS. NO PROSE ANYWHERE IN THIS DOCUMENT RESTATES THESE NUMBERS BY HAND.

Of the 10 direct haemostasis and analogue sources in the first portion, 1 was known to this review before the repaired search; 6 were surfaced by external review; and 3 were newly identified by the repaired search.

⚠⚠ ACROSS THE WHOLE CANONICAL OBJECT — 103 provenance-tracked sources, which is not the bibliography count: three tracked sources (Gagnon 1994, the JTH 2024 review, T4DM) were identified but never registered as references — 52 were known to this review before the repaired search, 47 were surfaced by external review, 6 were newly identified by the repaired search, and 19 are recorded as retrieved by it. ⚠ 84 entries are recorded as obtained and read.

⚠ Sources first discovered by the repaired search were nevertheless found by this review, because the search is an act of the review — which is why the field is temporal rather than a yes/no attribute. ⚠⚠ And one source was held internally from the outset while never being identified as part of the corpus it belonged to.

THE REMAINING EVIDENCE CLASSES, ON THE SAME EIGHT COLUMNS. THIS IS PART OF THE SAME CANONICAL OBJECT AND NOT A SECOND ONE.

⚠⚠ ONE DIMENSION IS AVAILABLE ONLY FOR THE HAEMOSTASIS AND ANALOGUE SUBSET, AND IS MARKED ⚠ nr — NOT RECONSTRUCTED — ELSEWHERE. Archive-holding provenance was reconstructed systematically for that subset because correction 123 turned on it; it was not reconstructed for the remaining corpus. ⚠ v4.23 claimed the same dimensions across both portions while the continuation simply omitted one, which is correction 138. The dimension is now present and its absence is declared per cell rather than by omission.

⚠⚠ v4.22 DECLARED ONE CANONICAL PROVENANCE OBJECT AND RETAINED A SECOND, HAND-MAINTAINED TWO-STATE TABLE IMMEDIATELY BENEATH IT, WITH A HAND-WRITTEN SYNTHESIS UNDER THAT. Those cannot coexist under correction 134’s own rule, and the retained table was exactly the object that would go stale when the protocol-complete search adds sources. Correction 135.

⚠⚠ A LIMIT ON THE RETRIEVAL COLUMN, STATED RATHER THAN LEFT AS AN APPARENT NEGATIVE. The preliminary search pass produced NO PER-SOURCE RETRIEVAL LOG. A blank in retrieved by the repaired search therefore means NOT RECORDED, not not retrieved, and the ✅ marks are only those the pass demonstrably returned. ⚠ The newly identified column carries no such ambiguity: ⛔ means the source was already known or already surfaced, and only six sources in this review were first discovered by the repaired search. ⚠ Producing that log is part of completing the protocol, and its absence is one reason the pass is not protocol-complete.

⚠⚠ THE METHODOLOGICAL CONSEQUENCE, WHICH IS NOT A RESTATEMENT OF THE TABLE. The provenance object demonstrates substantial under-retrieval by the original narrative method, particularly for literature outside erythrocytosis and haematocrit terminology. That retrieval history is why protocol-complete searching would strengthen reproducibility; under the declared narrative-review scope it is not a gate.

⚠⚠ AND ONE ENTRY IS WORSE THAN UNDER-RETRIEVAL, BECAUSE IT WAS NOT A RETRIEVAL FAILURE AT ALL. The 2003 haemostasis primary was in this review’s own archive from the outset, was read and discussed at §9.6 for its corrigendum, and was still absent from §9.4.6’s corpus until an external reviewer asked for it. ⚠ That is a failure to search what was already held, and it is the reason the archive is a named limb of the protocol.

⚠ CORRECTION NOTICE, IN ITS OWN PARAGRAPH. An earlier version carried a hand-written class-by-class summary of the table here, in a document that declares manual restatement of a generated object to be a build failure. It also generalised that this review did not find the mechanistic haemostasis literature, which the table contradicts: one haemostasis source was known to this review beforehand and another was held and read though never recognised as belonging to that corpus. Correction 141.

⚠ This table is the strongest internal evidence for the limitations of the original retrieval method and for keeping all absence claims explicitly review-level. Under the declared narrative-review scope, the search is not treated as a publication gate.

⚠⚠ THE SECOND EXTERNAL REVIEW NAMED THIS AS THE PRINCIPAL PUBLICATION BLOCKER, AND THIS REVIEW ACCEPTED THAT UNTIL v4.47. ⚠ Eden has since ruled that the document publishes whole on a declaredly non-exhaustive search. The specification below is UNCHANGED and is still owed; what changed is that the review no longer withholds publication pending it. Correction 211. The minimum specification is recorded here so that it is a task and not an aspiration: MEDLINE/PubMed, Embase and CENTRAL; a dated final search and a dated update search through deposit; explicit concept blocks for testosterone or gender-affirming hormone therapy, for erythrocytosis or haematocrit, and for thrombosis or clinical outcome; stated eligibility and exclusion rules; backward and forward citation chasing; and a stated disposition for every study already known to this review to be directly relevant. ⚠⚠ AND A SEPARATE SUPPLEMENTARY SEARCH ON A HAEMOSTASIS, COAGULATION AND PLATELET BLOCK, NOT REQUIRING ERYTHROCYTOSIS OR HAEMATOCRIT LANGUAGE. That limb is added because it is exactly the limb whose absence has already cost this review: the direct trans-masculine haemostasis analyses entered this corpus overwhelmingly after an external reviewer named them, and none would have been caught by a search requiring erythrocytosis terms. ⚠ No count is given here; per-source provenance is recorded in the canonical disposition object below. ⚠⚠ An earlier version described that object by a dimension count and by a four-word gloss, which went stale the moment correction 137 split discovery from retrieval, and which collapsed the very distinctions 137 had just created. Corrections 123 and 140. ⚠⚠ AND THE SEARCH MUST QUERY THIS REVIEW’S OWN ARCHIVE AS WELL AS THE DATABASES. At least one direct haemostasis primary was held here throughout and discussed in another section of this manuscript while absent from the haemostasis corpus; a search that only looks outward would not have found it either. ⚠⚠ AND A DATED UPDATE SEARCH THROUGH THE DEPOSIT DATE IS PART OF THE SPECIFICATION AND NOT A COURTESY. Material bearing directly on this review’s sources has appeared within the last fortnight, so an evidence state fixed in August cannot be carried into a later deposit without re-running it. ⚠ Until it is COMPLETED TO PROTOCOL, the review can defend its current corpus and cannot defend field-level completeness, and it says so at every absence mark rather than once here.

The correction that produced most of the others

⚠⚠ Summaries, introductions and secondary evidence tables were read in place of the primary papers they describe. Entries 1, 2, 3, 5, 6, 7, 9 and 13 share that single cause. It is the same failure this review documents in the evidence chain it examines — and the sixth instance occurred inside the material assembled to prevent the first.

⚠⚠ AND A FOURTH CAUSE, VISIBLE ONLY AT v0.4: a correction can be adjudicated, agreed and written down without ever reaching the manuscript. Entry 22 is that failure, and it accounts for three of the second round’s findings. An accepted correction is not a made correction until a check confirms it landed.

⚠⚠ AND A THIRD CAUSE, VISIBLE ONLY AT v0.3. Entries 17, 18 and 19 were all found by an external reviewer. None was caught by any internal check, and the checks that should have caught them were run — the table was verified three times, the omitted paper was read in detail, and the withdrawn mark was tested against the case that motivated it. ⚠ All three ran WITH this review’s position. That is the standing failure direction, and it is now evidenced across three entries that this review’s own machinery did not detect.

A second cause runs through the rest. ⚠ A claim was checked against a record of the source rather than against the source. Entries 8, 9, 12 and 14 turn on that: a seed’s negative claim, a table never opened, band sizes never counted, a section assignment inferred rather than read. The register is a good record and it is not a primary.

⚠⚠ AND A FIFTH CAUSE, VISIBLE ONLY AT v4.4, AND IT IS THE ONE THE VERIFICATION RULE CANNOT REACH. Entries 32, 33, 35, 36, 37 and 38 are not failures of verification against a source. Every sentence involved in them was individually true and individually checkable against a paper. What failed is the document’s account of itself: a corpus extended without regenerating the synthesis that describes it, a map maintained by hand while the sections beneath it were rebuilt, a count about the record asserted beside a record that contradicts it.

The correction record includes numerical, attribution, characterisation and document-state errors. Recent review has identified material errors introduced during earlier repairs. The record therefore supports neither a claim of numerical error-freedom nor a claim that source verification is complete. The current verification status is stated in the publication-status and deposit-state sections.

⚠⚠ CORRECTION NOTICE, IN ITS OWN PARAGRAPH. An earlier version ran the narrow observation and the broad conclusion together, and stated that source discipline was holding. Correction 79.

⚠⚠ CORRECTION NOTICE, IN ITS OWN PARAGRAPH. An earlier version stated a number of consecutive rounds here while the front matter defined a canonical lineage of six external adversarial reviews. ⚠ The denominator was never defined at the point of use and could not be generated, because the per-entry finder-and-round field correction 37 declares owed still does not exist. The count is removed rather than replaced. Correction 77. What is not holding is state propagation across an assembly of forty-odd artefacts, and no amount of source-level care addresses it. ⚠ The remedy is the one entry 35 names and entry 37 shows was applied too narrowly: every statement the document makes ABOUT ITSELF — counts, corpus sizes, maps, status blocks, and its own table of contents — is generated at build time from the thing it describes, or it is not made. ⚠⚠ Entries 51, 52 and 53 show that remedy being applied too narrowly a second time: it reached the correction total and §8, and did not reach the society-verification denominator, the heading hierarchy, or the single-sourcing of scientific state across subsections.

⚠⚠ AND A SIXTH CAUSE, VISIBLE ONLY AT v4.5, WHICH IS ABOUT WHAT THIS REVIEW CAN FIND RATHER THAN WHAT IT DOES WITH WHAT IT FINDS. Entry 47 is the fifth occasion on which a directly relevant trans-masculine study has been identified by an external reviewer and not by this review’s retrieval. ⚠ The pattern in the misses is now legible: the omitted analyses are the ones that do not use erythrocytosis or haematocrit language. A search built around this review’s exposure term cannot find the mechanistic literature about the same population, and no amount of care in reading what has been found repairs that. ⚠⚠ This is the strongest single argument for the reproducible search being a publication blocker rather than a formality, and it is the review’s own evidence against itself.

⚠⚠ AND A SEVENTH CAUSE, NARROWER AND WORTH SEPARATING FROM THE OTHERS. Entries 49 and 50 are neither source failures nor state failures. They are the review’s own grading scheme misapplied — precision facts recorded on the causal axis, and design features credited to an axis they do not identify. ⚠ §3 separates four questions precisely because they are easy to collapse, and this review collapsed two of them in two directions in the same assembly. The remedy is that §3 now states the rules explicitly rather than leaving them to be inferred from the definitions.

Declared limitations of method

⚠⚠ TWO LIMITATIONS ARE DECLARED RATHER THAN REMEDIED, AND THE DECISION TO DECLARE THEM IS FINAL RATHER THAN PENDING. They are stated here once, in full, and are not repeated elsewhere as open tasks.

1 — The search is not systematic and was not completed to protocol. A reproducible search specification exists and was partially executed. MEDLINE/PubMed was searched; Embase and CENTRAL were not searched to protocol; the preliminary retrieval was not screened to completion; citation chasing and a protocol-complete disposition process are unfinished. ⚠ The consequence is concrete: this review cannot claim to have identified every eligible study, and two reports now integrated at §9.4.3 ([85] and [86]) and one more ([87]) reached it through external adversarial review rather than through the search. ⚠⚠ That is evidence the search is incomplete, and it is recorded as such rather than explained away. A dated update search was run on 13 September 2026 and identified no further eligible study. ⚠⚠ That sentence alone would overstate what the search achieved. Diaz Mayor [101] was found by ordinary searching AFTERWARDS, is review-wide eligible, and was missed by the update search. The update search therefore did not close the retrieval gap, and this review does not claim it did. ⚠⚠ A second, different gap is also declared: this review has run claim-to-source verification, which asks whether its stated claims are true, but has not completed its SOURCE-INDEXED EXTRACTION AUDIT — taking each source as the index and asking what it contains that bears on this review’s question. Extraction-record COVERAGE is complete for the defined load-bearing set; accuracy of every source-to-manuscript use, and author sign-off, are separate and incomplete tasks (correction 663). Correction 413 shows what that gap costs: Oakes’ event-time haematocrit column sat unused in a table this review had read and verified. Completion of that audit is owed.

2 — No formal risk-of-bias instrument was applied. No ROBINS-I, ROB-2 or equivalent assessment was performed on any included study. ⚠ What stands in its place is this review’s own grading taxonomy — causal attribution, observed magnitude, clinical importance and population directness graded separately — which is more granular than a single RoB judgement on the specific question of threshold validation, and less standardised than a published instrument. It is not a substitute and is not offered as one.

⚠⚠ NEITHER LIMITATION IS MINIMISED HERE. A narrative review with an incomplete search and no risk-of-bias instrument is weaker evidence than a systematic review with both. The grades throughout this document are bounded by that fact, and a reader who needs a systematic review of this question does not have one in this document.

Deposit criteria

⚠⚠ COMPLETION CRITERIA FOR A VERIFIED REVIEW, WRITTEN DOWN SO THAT COMPLETION IS TESTABLE RATHER THAN ASSERTED. The criteria below govern a claim that this is a completed, verified scientific review. Every one must hold before that claim is made. ⚠⚠ Their open state does not by itself prohibit deposit of the working draft declared in PUBLICATION STATUS: the product is the author’s recorded ruling, and what is outstanding is disclosed there (correction 648).

1. Every bibliography entry is read in primary, or carries a formally adjudicated exception with its ground stated.
2. The archive gate passes at the final bibliography count.
3. The structural citation integrity audit passes at the final bibliography count.
4. A whole-document render parse produces no malformed emphasis in the live body.
5. Eden’s sign-off, in two limbs (correction 439): 5a claim-to-source — every substantive claim checked against its source; 5b source-indexed extraction — for each load-bearing source, what it contains on haematocrit magnitude, threshold crossing, event-time red-cell values and thrombotic outcome, and whether this review uses each. Load-bearing set: every source cited in §9.2–§9.5 plus the §9.4.3 subcorpus — derived at build, never listed (correction 456). Both halves are reported in the deposit-state block and derived at build (correction 458): read coverage, and the extraction record — whether this review uses what each source contains, recorded per entry as represented, integrated, or explicitly dispositioned as not bearing on the review question. 5a and the sign-off are open.
6. §9.4.3 reaches KNOWN-CANDIDATE DISPOSITION COMPLETE — every identified candidate has an auditable state (integrated, held out on a ground VERIFIED against the source, or carrying a formal unresolved-source exception) and the number of unresolved candidates is stated on the face of §9.4.3. ⚠⚠ This is NOT the same as ELIGIBILITY RESOLVED, which would require that no known candidate’s membership remains unknown. This review does not claim that and cannot while unretrieved candidates stand. The earlier word closed promised what the exception architecture does not deliver and is withdrawn. ⚠⚠ A declared unknown is still an unknown: an owed retrieval does NOT close a corpus, because corpus membership for a known candidate cannot be established by saying its retrieval is owed. Unknown unknowns are a different object and are declared under limitation 1. ⚠⚠ Two booleans, EMITTED from the candidate object and reported in the deposit-state block below (corrections 433, 454): known-candidate disposition complete, and eligibility resolved. Deposit gates on the first. The second is reported, not claimed, and is false while any candidate is unresolved. Neither value is typed here; a typed copy is what correction 454 removed.
7. The two limitations declared above stand as declared.
8. Appendix R integrity: every ruling cited in the body resolves to one appendix block, and no block carries unrecovered truncated register text. ✅ MET at correction 440 — the last four rulings restored from master register v63; asserted at build by appendix_r_integrity.

⚠⚠ DEPOSIT STATE — EMITTED AT BUILD FROM THE CANONICAL OBJECTS (correction 455).

Object	Source	State	Basis
Gate 1a bibliographic identity	DERIVED	MET	identity established 102/102: 96 by DOI, 2 by public registry record, 2 by official publication URL, 2 official grey literature with a recorded archive identity; 100/102 independently retrievable by public identifier
Gate 1b bibliographic metadata	DERIVED	MET	no entry abbreviates an establishable author list
Gate 2 structural citation integrity	RUN RECORD	MET	N=102, recorded in the gate block
Gate 3 [81] Okano exception	RUN RECORD	MET	adjudicated; abstract-read, full text not obtained
Gate 4 archive gate	RUN RECORD	MET	N=102, recorded in the gate block
Gate 5 rendered-output integrity	DERIVED	MET	no nested strong node and no emphasis span carrying a warning marker past its opening, on these bytes
Crit. 5a claim-to-source verification	AUTHOR	OPEN	Eden’s; no machine in this build can discharge it
Crit. 5b read coverage	DERIVED	MET	70 of 84 read in primary; 14 adjudicated; none untagged
Crit. 5b extraction record	DERIVED	MET	every load-bearing source carries a disposition
Crit. 5b sign-off	AUTHOR	OPEN	Eden’s; no machine in this build can discharge it
Crit. 6 known-candidate disposition	DERIVED	MET	every named candidate carries a state
eligibility resolved	DERIVED	FALSE	unresolved: Fan, Inoue, Mullins, Park-SOVT, Penney, Stanley — REPORTED, NOT CLAIMED, and not the deposit-gating object
Crit. 8 Appendix R integrity	DERIVED	MET	every cited ruling resolves to one block, no truncation

⚠⚠ VERIFIED-REVIEW COMPLETION IS NOT ESTABLISHED. 2 object(s) open: Crit. 5a claim-to-source verification; Crit. 5b sign-off. ⚠ These objects govern a claim that this is a completed, verified review. They do not prohibit deposit of the working draft declared in PUBLICATION STATUS, which is governed by the recorded product ruling and that disclosure (correction 648).

⚠ This block is EMITTED at build from the canonical objects (correction 455). Do not edit it and do not restate any line of it in prose: a second copy is the defect corrections 434, 436, 439, 443, 445 and 446 each recorded at a different site.

⚠ This list is the standard. It is not no reviewer raises a further objection, which is unbounded in a document of this length and would make deposit impossible by construction. ⚠⚠ A reviewer finding something after every criterion below holds is a correction for a later version, not a bar to depositing this one.

Open methodological and record tasks

PART I — THE FRAME

1. Two questions, kept apart

Testosterone raises the haematocrit. That is not in dispute, it is not new, and nothing here argues otherwise. What follows from it is disputed at every step, and the disputes are of two kinds.

The first is evidential. Does a raised haematocrit on testosterone cause thrombosis, and if it does, how large is the effect, in whom, and at what level? Is the answer the same for arterial and for venous events? Has any of it been measured in trans-masculine people, or are we reasoning across from men treated outside gender-affirming care? These are questions about what has been measured and how well.

The second is practical. Should haematocrit be monitored, and how often? Is there a level at which treatment should be reduced, interrupted or stopped, and should anyone be excluded from testosterone on this basis?

This review answers the first kind and does not answer the second.

That is not modesty and it is not evasion. The two questions take different inputs: the practical one takes in what a patient wants, what interrupting treatment costs them, what a service can actually deliver, and which error is worse to make. None of that is evidence about haematocrit, and a review that quietly folded it in would be smuggling a policy position inside an evidential one.

It does not follow that evidence is irrelevant to the second question. Values supply the weights, and evidence supplies the magnitudes those weights are applied to. Better data could move a reader without changing anything they value, which is a reason to gather it rather than to treat the question as closed.

Why the seam matters here in particular. An apparatus already exists and is specified with confidence: numeric thresholds from ten societies, monitoring intervals, and triggers for dose reduction and for phlebotomy. Against that background, a review reporting that the underlying evidence is thinner than the apparatus implies reads very easily as an instruction to dismantle it.

Watch how the weld is made. The evidence for this threshold is weak is an evidence claim; therefore stop monitoring is a policy claim, and the first does not entail the second, because weak evidence can justify monitoring, monitoring at a different level, monitoring a different quantity, or research.

Run the weld the other way and it fails as plainly. Erythrocytosis is common on testosterone is an evidence claim of high certainty, and therefore the current thresholds are justified is a policy claim it cannot carry. That a change occurs often is not evidence that it is harmful, and it is not evidence that the level at which we intervene is the right one.

Where this review has something bearing on practice it says it as a fact about the evidence rather than as a recommendation: that some numerical thresholds originate in guidance written for populations other than trans-masculine people; that the timing of current monitoring broadly corresponds to the observed early rise in haematocrit but has not itself been shown to improve an outcome; that a quantity is being watched for an outcome its causal support does not reach.

What to do with those facts is a clinical and a policy judgement, and this is not the document that makes it.

Confidence: not applicable — this section is the method, not a finding.

2. Three errors, named once so they can be watched

The evidence here is divided enough that both positions are tempted into the same mistakes, mirrored. One is carried from the series; two are specific to this review, and this review has committed both.

Absence of evidence as evidence of absence. No trans-masculine study identified here estimates thrombotic risk as a function of haematocrit, or uses a design capable of attributing thrombotic events to the haematocrit elevation, and that is a statement about what has been measured rather than a statement that no link exists. Equally, sparse event counts without person-time or an appropriate comparator are not reassurance that risk is absent. ⚠ Events have been counted — §9.4.3 counts thirteen descriptions of affected participants across seven reports, with the remainder recording none, and their uniqueness is not established — and what is missing is a design capable of relating them to haematocrit. Both errors are available here and both are made. Where this review grades a causal contribution not established, it means that no adequate study identified in this review settles it, and it means nothing about what an adequate study would find.

The open-ended category, read as though it settled what lies above it. This is the one that catches careful readers, and it is documented in the literature examined here. A cohort’s highest exposure band is often defined as at or above some value, with no ceiling, and the estimate attached to that band is then quoted as though it described risk at any chosen value above the floor. It does not. It is a single average over everyone in the band, and in the cohorts examined here those people cluster near its bottom. ⚠ Nor does the band’s definition tell you who was actually measured within it: a category defined as open above 49% neither establishes nor excludes that anyone reached 54%.

The move has been made on this material in print: a published paper took another cohort’s open-ended top category and reported its estimate as though it resolved risk across the range above it. This review came close to repeating it, and stopped only after reading the source tables row by row.

Read correctly, an open-ended category yields one pooled estimate for everyone in it. It does not resolve risk at any particular value inside the category, and it says nothing either way about how many people were measured at any such value. Where a society threshold sits inside such a category, the evidence supplies no estimate specific to that threshold — which is neither an endorsement of it nor an argument against it.

Confidence in an association is not confidence about which quantity carries it. This is the third, and it runs through Part V. Haematocrit, haemoglobin and red cell count move together; no observational association study separates them, and the one measured in clinic is not thereby the one that acts. A study can establish robustly that something in that cluster is associated with an outcome and be incapable of saying which.

They are different objects, drawn from different evidence, and they can diverge sharply.

This is not a technicality. It is why a reader can be told, truthfully, both that the association is well established and that we do not know whether the monitored quantity is the causal one, and conclude that someone is being evasive. Nobody is. The two statements are about different things. Where this review reports an association it says which quantity the estimate attaches to, and where attribution and association come apart it says so.

A fourth candidate, recorded and not selected, so that the choice is visible: adjustment depth as a tiebreaker, the assumption that the more heavily adjusted study is the more reliable one. This review’s own record shows it failing here, because the most heavily adjusted cohort is the strongest positive and removing the adjustment mismatch widens the discrepancy rather than narrowing it. It is a genuine trap, it is more technical than the other three, and it is not among them.

Confidence: not applicable — this section is the method, not a finding.

3. How certainty is graded here

Four questions, kept apart. A single rating cannot carry them, and collapsing them is how most argument about this evidence goes wrong. Every confidence block answers these four, in this order, and names any it cannot answer.

Causal attribution. Was the observed association produced by the exposure, or by something else — confounding, reverse causation, ascertainment, or who was selected into the exposed group? It is answered by study design rather than by the size of the estimate. Note the scope: the question is not does testosterone raise the haematocrit, which it plainly does, but did the raised haematocrit produce this excess of events in this study.

Observed effect size. How large was the association, and how precisely is it known? It is answered by the data, and it is usually estimable even where the first question is not.

Clinical importance. Would the difference be decision-relevant to a patient? It can be assessed by absolute risk difference over a stated horizon, by number needed to harm, by event severity, by movement between clinical categories, or by what a response would cost. Each of those is a method, and none of them is the definition.

Population directness. Was this measured in the people the claim is about? It is answered by who was actually studied. A finding measured in trans-masculine people on testosterone is direct; one extrapolated from men outside gender-affirming care is indirect, and it is weaker than it looks however large the study.

These come apart constantly. A study can produce a precise, heavily adjusted estimate that establishes nothing causally and whose clinical meaning is unknowable, which is not a hypothetical but a description of most of what is graded below.

Ratings are per claim, using the series scheme, as editorial judgements rather than formal GRADE.

Robust — the finding would be surprising to overturn. Reserved for mechanism that is not in dispute, for documentary fact, and for effects directly observed and consistently reported.

Moderate — the direction is reasonably secure; the size is not.

Limited — the evidence cannot support a confident estimate in either direction. It is not a synonym for no effect.

Not established — ⚠ relevant studies were identified, and none identified in this review has a design adequate to settle the claim.

Not estimable — the evidence is adequate in kind, but the quantity cannot be extracted from it.

Not isolated — the quantity is estimable, but the exposure cannot be separated from a correlated one.

None — no eligible study was identified among the literature reviewed.

The four absence marks are not synonyms. Each names a different reason a question is unanswered, and the differences do work later.

⚠⚠ And one limit governs every one of them. This is a narrative review and not a systematic search. It can report that no eligible study was identified. It cannot establish that none exists, and where an absence is load-bearing it says so rather than asserting more than its method supports.

Two clarifications, because both are routinely misread. Robust does not mean large: it says the finding would be surprising to overturn and says nothing about the size of the effect. And population directness is graded in words rather than marks, on a fixed three-point scale — direct, measured in trans-masculine people on testosterone; mixed, the fact measured there and the detail borrowed; indirect, extrapolated from cohorts outside gender-affirming care. It describes provenance rather than strength.

A third clarification, specific to this review. Where an outcome is graded limited, the next sentence says why, because the mark covers two situations that are not alike: evidence that is thin, and evidence that is divided. Several large well-conducted cohorts disagreeing is limited, and so is one small study, and a reader who cannot tell which has been misled by the mark.

⚠⚠ A FOURTH CLARIFICATION, ADDED AT v4.6, AND IT IS A RULE RATHER THAN A NEW MARK. ⚠ Provenance caps a causal mark and is stated in prose beside it. A randomised design read only in published abstract and registered protocol cannot be appraised for allocation implementation, allocation concealment, protocol deviation, attrition, analytic population, outcome completeness or selective reporting, and grading it as though it could is treating randomised as sufficient rather than as one input. Such a claim is graded no higher than LIMITED, with the PROVENANCE LIMIT NAMED IN PROSE beside the mark. ⚠ The rule prescribes that a reason be given, not the words used to give it. An earlier version fixed a particular phrase, which then conflicted with the mark when the retrieval position changed from in progress to not obtainable in depositable form — correction 158. ⚠⚠ No new grading token is created: the scheme’s seven marks are unchanged, and the qualifier is a prose clause, not a category. §9.5’s management statement is the live case and is graded accordingly.

⚠ A FIFTH CLARIFICATION, AND IT GOVERNS TWO CORRECTIONS AT v4.6. Sparse events principally limit precision. Randomisation supports identification of the effect of assignment under the relevant assumptions and with appropriate outcome ascertainment and analysis; sparse events do not by themselves invalidate that design. Missing outcomes, post-randomisation selection and other biases still require appraisal. ⚠⚠ Grading causal attribution down because a trial reported twelve events confuses the first axis with the second, and this review did exactly that at v4.5. Where a randomised corpus is causally sound and numerically thin, the causal axis records the design and its scope limits, and the effect-size axis records the imprecision.

⚠ A SIXTH CLARIFICATION, OF THE SAME KIND. Prospective within-person measurement establishes temporality and within-person association. It does not by itself identify a causal effect, because it removes neither secular change, repeated-testing effects, other treatment-associated changes, nor time-varying confounding. Where an analysis is prospective and uncontrolled, the within-person change and the causal attribution are graded separately and the absence of a contemporaneous untreated comparator is named.

⚠⚠ HOW THIS SCHEME HAS CHANGED WHILE BEING APPLIED, DISCLOSED RATHER THAN ARGUED. A reasonable reviewer will ask whether these grades were generated prospectively from fixed criteria or rationalised around an evolving corpus. The scheme HAS changed during the review’s life, and the changes are listed here so that the question can be answered from the record rather than from assurance.

Change	What it did	Why	Where recorded
NOT REACHED introduced, applied to three statements, withdrawn the same day	added and removed a mark	it failed its own test on its first application	correction 19
NONE scoped to no eligible study identified in this review	narrowed a mark	the review cannot establish that no such study exists	§3, correction 21
Provenance caps a causal mark, in prose, with no new token	added a rule	randomised was doing the work of appraised	correction 48
Sparse events belong on the effect-size axis, not the causal axis	added a rule	the two axes were being collapsed	correction 49
Prospective uncontrolled measurement grades within-person change and causal attribution separately	added a rule	consistency across overlapping assays is not replication	correction 50
Combined-evidence and direct-population causal marks stated separately	split an axis	indirect randomisation was doing the work of directness	correction 71

⚠ Every change is a NARROWING or a SEPARATION. None widened what a mark licenses, and none was made to accommodate a particular result. ⚠⚠ That is a claim a reader can check against the table, and it is offered on those terms and not as a reassurance. The marks themselves — the seven tokens above — have not changed since the review’s first version, with the single exception of NOT REACHED, which was withdrawn and is recorded.

Population flags mark any finding extrapolated from another population — here, overwhelmingly, men treated outside gender-affirming care. They are not decorative, and where a claim rests on data from those populations it is weaker than it looks.

Confidence: not applicable — this section is the method, not a finding.

PART II — WHAT TESTOSTERONE IS, AND WHAT IT DOES

Testosterone is the principal pharmacological agent used to induce and maintain masculinisation, and for many people it is the only systemic hormone they take. It is not universally monotherapy: progestogens or a gonadotrophin-releasing hormone analogue may be added for menstrual suppression, and indication-specific treatments are used for particular effects. In standard masculinising therapy, testosterone is the principal systemic androgen and there is no routinely recommended second pharmacological agent with a role analogous to the anti-androgen that often accompanies oestradiol on the feminising side. ⚠ An earlier version wrote there is no second masculinising agent, which is an absolute pharmacological claim rather than a statement about standard practice, and other androgenic compounds exist even though they are not standard alternatives here. Whatever this review finds about raised haematocrit, it finds about one drug given to nearly everyone who is treated.

What is prescribed is a short list. Dimakopoulou and Seal, endocrinologists at the largest UK gender clinic, set out the preparations that clinic uses: short-acting testosterone esters or enanthate by intramuscular injection, titrated from 150 mg to 250 mg every seven to thirty-five days; long-acting testosterone undecanoate, 500 to 1000 mg every six to fifteen weeks after a loading dose; and daily topical gel, titrated from 20 mg to 80 mg. They describe the short-acting ester and the gel as first line, chosen because their shorter duration of action makes them more easily reversible if complications occur, with undecanoate reserved for the longer term on grounds of convenience.

That reasoning is worth holding onto, because it is a monitoring argument made before any monitoring question is asked. The first-line preparations are selected in part for how quickly they can be stopped. A review examining what happens when haematocrit rises is examining a system that has already been designed, at the point of prescribing, around the possibility of needing to reverse course.

⚠ This illustrates one service model in which reversibility influences first-line formulation choice. It does not establish that masculinising testosterone practice as a whole is designed around reversibility.

The regimens this clinic describes are adult male replacement doses, and the paper says so plainly: the adult testosterone doses it uses in binary masculinising therapy are the same as those used for hypogonadal men. ⚠ That is established for the regimens of the clinic its authors describe. More broadly, commonly used masculinising regimens occupy overlapping dose ranges, although formulation, interval, achieved exposure and practice vary substantially between countries and services. The overlap is of considerable consequence for everything downstream in this review, and it is stated here once rather than repeated: nominal masculinising doses substantially overlap with the replacement regimens used in hypogonadal men, so the milligram range alone does not account for the population indirectness. ⚠ That is not the same as saying dose plays no part. Formulation, injection interval, route, peak-to-trough behaviour, achieved concentration and body composition can all differ, and §9.5 argues at length that formulation may alter erythrocytosis. What the overlap establishes is that the mismatch is not simply a matter of how many milligrams.

The monitoring targets are specific. For short-acting esters the clinic monitors a trough level on the day of injection with a target of 8–12 nmol/L, and a peak seven days later targeting 25–30 nmol/L; for undecanoate and for gel, a target of 15–20 nmol/L. ⚠ These are that clinic’s stated targets as its clinicians report them. They are not a trial endpoint and no outcome evidence is offered for them here.

Non-binary regimens are a different case and the paper is candid about the evidential position. Where partial masculinisation is sought, the approaches described are either an adult male dose for a time-limited period, or a lower dose sustained. The authors state that the hormonal regimens used in this field have not been examined by randomised controlled clinical trials, and that current options are extrapolated from pubertal induction in cisgender children and from the treatment of binary trans men.

⚠ That statement is made about partial-masculinisation regimens specifically, and this review does not extend it further than its authors do. They describe those regimens as extrapolated in part from the treatment of binary trans men — which makes binary treatment one of the sources extrapolated from, not part of what is described as unsupported. What can be said is narrower: many dose ranges, intervals and target levels in masculinising practice are clinical convention supported by observation. ⚠ A randomised comparison of two testosterone esters in trans men has since been published and is analysed at §9.5.

⚠ One consequence is flagged and not resolved. The paper notes that where target testosterone values sit below the adult male range, bone health may be compromised and lipid profiles may worsen, and it recommends bone-density monitoring by DEXA every three to five years in that group. A review of testosterone safety that grades outcomes at replacement dose does not automatically speak to a population deliberately maintained below it, and the tranches should say which population each grade covers.

PART III — THE EVIDENCE BASE

5. The apparatus is specific, and it answers more than one question

A clinician starting a trans-masculine patient on testosterone has numbers to work to, and they come from two places rather than one.

⚠⚠ THE TRANS-SPECIFIC GUIDANCE CARRIES ITS OWN NUMBER. The Endocrine Society’s guideline for gender-dysphoric and gender-incongruent people classifies, for transgender males on testosterone, erythrocytosis at a haematocrit above 50% as a very high risk of adverse outcomes, and directs that haematocrit or haemoglobin be measured at baseline and every three months for the first year, then once or twice a year.[1] ⚠ An earlier version of this section said the numbers came from guidance written for a different population. That was wrong, and it is corrected here rather than quietly amended.

⚠⚠ THE GUIDELINE DOES NOT EXPOSE THE DERIVATION OF THAT NUMBER. The 50% cutoff appears in a risk table with no citation attached to the numerical value. The guideline elsewhere names erythrocytosis among the adverse events to be avoided in transgender males and directs that it be monitored — so it treats the concern as real; what it does not supply is a chain showing where 50% came from.[1] ⚠ This review has reconstructed the Endocrine Society’s DOCUMENTARY lineage back through the 2009 and 2006 guidelines. It has NOT reconstructed the evidential derivation of the numerical value.

⚠⚠ BUT THE NEXT NODE IS NOW IDENTIFIED, AND IT IS NOT THE 1994 COHORT. The 2017 guideline states in its own Objective that it updates the Endocrine Society’s 2009 transgender guideline, and carries a Changes Since the Previous Guideline section.[1] ⚠ The obvious question is therefore whether the 50% classification was inherited from the 2009 document rather than derived in 2017 — and, beyond that, whether the 2009 document inherited it from the Society’s earlier guidance for hypogonadal men.

⚠ Both predecessor documents have since been obtained and read in full, and the lineage is set out below. ⚠⚠ Two candidate ancestral routes remain open and must be kept separate: the internal guideline lineage running 2006 → 2009 → 2017, and later authors’ retrospective attribution toward the 1994 Framingham analysis. Neither has been traced to a derivation of the numerical value.

⚠⚠ THE DOCUMENTARY GUIDELINE LINEAGE IS NOW TRACED FROM THREE PRIMARY SOURCES READ FIRST-HAND. THE NUMERICAL DERIVATION IS NOT. ⚠⚠ The >50% cutoff itself was preserved across all three documents; what changed was its risk-category framing and clinical context.

	Value	Risk category assigned	Function
2006, androgen-deficient men[2]	Erythrocytosis, haematocrit >50%	Moderate to high risk of adverse outcomes	initiation contraindication — testosterone should not be administered
2009, transsexual persons[3]	Erythrocytosis, haematocrit >50%	⚠⚠ Very high risk of SERIOUS adverse outcomes	condition exacerbated by cross-sex hormone therapy
2017, gender-incongruent persons[1]	Erythrocytosis, haematocrit >50%	Very high risk of adverse outcomes	⚠ same >50% very-high-risk classification; surrounding risk taxonomy and wording not identical

⚠⚠ THE >50% VALUE WAS PRESERVED, WHILE ERYTHROCYTOSIS MOVED FROM THE 2006 GUIDELINE’S MODERATE TO HIGH RISK CATEGORY TO THE 2009 TRANSGENDER GUIDELINE’S VERY HIGH RISK OF SERIOUS ADVERSE OUTCOMES. ⚠ The 2017 update’s heading reads very high risk of adverse outcomes; the tier and the value persist, the wording is not identical. In the 2006 source, erythrocytosis sits one tier below the top, which contains metastatic prostate cancer and breast cancer. In the 2009 transgender guideline it appears in the top tier, and it does so again in 2017. ⚠⚠ These are different classifications in different guideline frameworks. This review has documented the difference, not the editorial reasoning that produced it.

⚠ The 2009 guideline’s monitoring discussion cites the 2006 document when addressing avoidance of chronic testosterone adverse effects, particularly erythrocytosis.[3] The citation relationship is documented. What is not documented anywhere in the three texts is a derivation of the numerical value, or a stated basis for the change of category.

⚠⚠ AND THE SECOND OBJECT WAS NOT CARRIED FORWARD. The 2006 guideline carries two distinct numbers with two distinct functions: above 50%, do not start; and above 54%, stop therapy until haematocrit decreases to a safe level, evaluate the patient for hypoxia and sleep apnoea, and reinitiate at a reduced dose.[2] The same >50% value appears in the 2006 cis-male initiation contraindication and in the 2009 and 2017 trans-specific pre-treatment risk tables. ⚠⚠ That is documentary co-occurrence plus a citation relationship. It does not establish that the trans-specific value was numerically derived from, or functionally inherited from, the 2006 initiation threshold — the 2006 document says do not start, while 2009 and 2017 use the value as a risk classification inside a pre-treatment evaluation framework. ⚠ The explicit >54% on-treatment action rule in the 2006 guideline is not reproduced in the 2009 or 2017 trans-specific monitoring provisions. ⚠ This review records the presence and the absence. It does not assert an editorial act, and does not claim the 54% object travelled the same pathway.

⚠ The distinction between an initiation caution and an on-treatment action level — set out at §9.3 — is explicit in the 2006 primary source and is not reproduced in the same form in the trans-specific documents that follow, so the distinction predates this review. ⚠⚠ This review records the presence and the absence. It does not claim an editorial act, a transmission event, or a reason. It is named here as the reconstruction this review owes, and as a more promising route than the 1994 cohort that later authors point to. ⚠ An earlier version stated the same finding twice in consecutive sentences; the duplicate is removed rather than left standing beside itself.

AND A SECOND TRANS-SPECIFIC STANDARD AGREES ON MONITORING WHILE SUPPLYING NO NUMBER. WPATH’s eighth Standards of Care recommends, at Statement 12.17, that clinicians monitor haematocrit or haemoglobin in transgender and gender diverse people treated with testosterone.[4] Its supporting text states that there are good quality data suggesting a rise in haematocrit or haemoglobin is associated with treatment, and that the expected rise should be consistent with reference ranges in cisgender males.[4]

⚠⚠ That is a reference-range statement and a monitoring recommendation. It supplies no numerical thrombotic action threshold, and Statement 12.17 names none. ⚠⚠ Its reference-range statement belongs to object one of the taxonomy at §9.3. Its recommendation to monitor is a separate clinical instruction that the six numerical objects do not represent at all — monitoring is not itself a threshold. It supplies neither an operational erythrocytosis definition nor an on-treatment action level. ⚠ An earlier version placed it at objects one and two; object two is an operational definition, and Statement 12.17 gives none.

⚠ The taxonomy therefore needs an additional unnumbered clinical instruction stated alongside its six numerical objects: the decision to monitor. ⚠ That decision is independently recommended in trans-specific guidance. This review does not evaluate whether monitoring, versus no monitoring, improves clinical outcomes — no study identified here addresses it.

⚠⚠ AND A THIRD TRANS-SPECIFIC DOCUMENT, PUBLISHED IN AUGUST 2026, USES THE SAME VALUE IN TWO EXPLICIT NUMERICAL FUNCTIONS, THEN SLIDES INTO A THIRD WITHOUT CARRYING THE NUMBER. An ASRM Practice Committee Opinion on transgender and gender-diverse care:[5]

— defines erythrocytosis as haematocrit above 50%, occurring in as many as 1 in 6 patients on gender-affirming testosterone — object two;
— sets a monitoring boundary, directing a complete blood count to ensure that the haematocrit is not excessively elevated (>50%);
— and then, separately and WITHOUT restating the number, states that with elevated testosterone or haematocrit levels, a reduction in dosage may be warranted, listing decreased dose, a change of route from injection to transdermal, or serial blood donation. ⚠⚠ An earlier version of this passage called that a management trigger at 50%. The source does not attach the number to that sentence, and the overstatement is corrected here.

⚠ It also directs that other causes of erythrocytosis, such as smoking and obstructive sleep apnoea, should be explored — the differential §11 sets out.

⚠⚠ THIS IS THE CLEAREST DEMONSTRATION IN THE REVIEW OF WHY THE TAXONOMY AT §9.3 IS NECESSARY. One current document uses >50% both as its definition of erythrocytosis and as a numerical monitoring boundary, then discusses management of an elevated haematocrit without restating a numerical trigger — placing definition, surveillance and management beside one another without establishing that one value governs all three — and the taxonomy exists precisely because those are different objects that need not share a value.

⚠⚠ AND THE DOCUMENT CONTAINS BOTH SIDES OF THIS REVIEW’S QUESTION, TWO PARAGRAPHS APART. It states that erythrocytosis if left uncontrolled, can increase the risk for thrombosis; and then that there is no evidence for an increase in cardiovascular events in TGD men on testosterone.[5] ⚠⚠ THOSE ARE NOT LOGICALLY CONTRADICTORY, AND AN EARLIER VERSION OF THIS PASSAGE CALLED THEIR COEXISTENCE A TENSION. THAT WAS WRONG BY THIS REVIEW’S OWN FRAMEWORK. The first concerns erythrocytosis and thrombosis; the second concerns testosterone exposure and cardiovascular events. §1 insists throughout that those are different causal questions, and the correction is applied here. ⚠ What requires examination is not the coexistence but the evidence attached to the erythrocytosis-specific claim, set out below.

⚠⚠ THE CITATION CHAIN HAS NOW BEEN READ FIRST-HAND AND IT IS A §9.6-GRADE FINDING. The sentence erythrocytosis… if left uncontrolled, can increase the risk for thrombosis carries two references.[5]

Reference 28 is a long-term evaluation of cross-sex hormone treatment in transsexual persons — a clinical follow-up study, not an analysis of thrombosis as a function of haematocrit. Reference 29 is Red blood cells in thrombosis, a general mechanistic review in a haematology journal, not a study of testosterone-treated trans-masculine people at all.

⚠⚠ NEITHER SOURCE VALIDATES A 50% HARM THRESHOLD, AND NEITHER ESTIMATES THROMBOSIS AS A FUNCTION OF HAEMATOCRIT IN THIS POPULATION. The pair supplies trans-population clinical context and general red-cell biology; the numerical clinical claim is assembled from them rather than found in either.

⚠ This review previously carried this as an owed check and declined to state it on external review’s description. The reference list was in the document held here all along, and had been cut by a retrieval limit rather than being absent. ⚠⚠ The check is now discharged first-hand. The earlier refusal was correct on the evidence then available and is recorded rather than deleted.

The third place is guidance written for hypogonadal men, most often reached through Supplementary Table 7 of a 2024 meta-analysis of testosterone replacement and thrombosis, which tabulates ten societies in nine rows under three headings: cardiovascular and thromboembolic risk, recommendation against testosterone replacement, and prior to testosterone replacement.[6]

Read as the table presents them, the numeric values sit in the recommendation against column, alongside entries such as a recent cardiovascular event, severe heart failure, polycythaemia, and untreated congestive failure.[6]

⚠⚠ EIGHT OF THE NINE CANNARELLA ROWS HAVE BEEN VERIFIED AGAINST THEIR CORRESPONDING PRIMARY GUIDANCE, AND THE TABLE BELOW REPORTS BOTH WHAT THE TABULATION PRINTS AND WHAT EACH PRIMARY SAYS. EAU is the unverified Cannarella row: the specific edition underlying Cannarella’s EAU entry has not been established. Current EAU guidance (EAU-SRH-2026, Sexual and Reproductive Health 2026) has separately been read in primary and is reported here as current-guidance evidence, not as verification of Cannarella’s historical EAU row. ⚠ See correction 230 for the same principle applied to EAU-2019/Dohle.

⚠⚠ THE THIRD COLUMN IS THE ONE THE TABULATION HAS NO ROOM FOR, AND IT IS WHERE THE CHECKED ROWS CONVERGE. ⚠ Seven of the EIGHT Cannarella rows verified in primary use 54% as the numerical cut-point for an on-treatment action or target, although operators and specified actions are not uniform; AACE uses 50%. ⚠ Current EAU guidance (EAU-SRH-2026) also uses >54% as its on-treatment action threshold — stated here as separate current-guidance evidence, not as verification of the Cannarella historical row. ✅ EAU (2026) sets the on-treatment action threshold at >54%. Eight Cannarella rows verified in primary; current EAU guidance also separately read. Convergence arithmetic below reflects the eight verified rows.

⚠ Of the eight historically verified Cannarella rows, FIVE carry the tabulated number accurately and THREE do not. ⚠ EAU-SRH-2026 is current guidance read separately; it was not in Cannarella’s tabulation. ⚠⚠ ICSM is a fifth accurate NUMBER and a defect of a different kind: the tabulation carries the value and cannot carry the source’s own attribution of it to another body’s 2010 guideline, nor its statement in the next sentence that such historical contraindications must be re-evaluated. BSSM, EAA, ISSAM and SIAMS/SIE are correct. The Endocrine Society row inverts its operator and drops the altitude qualifier; the AUA and AACE rows print the ON-TREATMENT value under a heading that means initiation, and omit the actual initiation position — which for AUA is 50% and for AACE is unquantified. ⚠⚠ This review does not describe the tabulation as systematically unreliable, and an intermediate reading of three rows would have supported exactly that. It is unreliable in a specific and predictable way, set out below.

The prior to column contains, for five of the nine rows, the instruction to measure haematocrit and nothing else; the other four add between one and three further items — patient information for the AUA, a venous-thromboembolism history for the EAA, an electrocardiogram for the EAU, and for SIAMS/SIE a global cardiovascular-risk assessment and a family, personal and clinical history of thromboembolism.[6]

⚠⚠ These are not all answers to the same question, and this review does not present them as though they were. A value that contraindicates starting treatment and a value that triggers stopping it are different decisions. A society can hold both at once without inconsistency, and several plainly do.

The tabulation does not separate them. The same meta-analysis’s discussion states the two decisions apart: societies recommend against starting testosterone in patients with elevated haematocrit, and suggest monitoring to reduce or discontinue above 54%, or 50% for two societies.[6]

So the spread in that column is not straightforwardly a disagreement. Part of it is between societies. Part of it is between clinical decisions. ⚠⚠ AN EARLIER VERSION SAID WHICH PART IS WHICH COULD NOT BE SETTLED FROM THE TABULATION. THAT WAS TRUE OF THE TABULATION AND IS NO LONGER TRUE OF THIS REVIEW, BECAUSE EIGHT OF THE NINE ROWS HAVE BEEN READ IN PRIMARY. THE OPEN QUESTION IS DISCHARGED AND THE ANSWER INVERTS THE FRAMING.

⚠⚠ AMONG THE EIGHT CANNARELLA ROWS VERIFIED IN PRIMARY, THE INITIATION POSITIONS GENUINELY VARY AND THE ON-TREATMENT ACTION LEVELS LARGELY DO NOT. The initiation position runs from unquantified, through 48–50%, 50%, 52%, to 54% — a real disagreement. Seven of the eight verified Cannarella rows use 54% as the numerical cut-point for an on-treatment action or target, although operators and specified actions are not uniform; AACE uses 50%. The historical EAU row remains unverified and is not included in that denominator. ⚠⚠ THE TABULATION CANNOT REPRESENT THOSE ON-TREATMENT VALUES AS ON-TREATMENT VALUES, BECAUSE ITS SINGLE ACTION COLUMN IS HEADED RECOMMENDATION AGAINST AND HAS NOWHERE TO PUT THEM. ⚠ The defect is one of FUNCTION, not of numerical absence: a value may appear in the tabulation while being assigned a clinical function its source does not give it, and numerical agreement therefore does not establish accurate representation of a recommendation (correction 518). That convergence is invisible in the source through which this literature is most often read.

⚠⚠ THIS CUTS AGAINST THIS REVIEW AND IT IS STATED FIRST. Each of the eight verified Cannarella rows represents a society or societies that states its two objects separately, with distinct values and distinct actions, and the collapse into one column is done by a secondary source. ⚠ ⚠⚠ Those eight are more internally coherent than a tabulation-level reading makes them look. Whether that holds for the unverified historical EAU row is not established. ⚠ §9.3.2’s six-object taxonomy is not a defect this review discovered in the societies. It is a distinction the societies already draw and the tabulation cannot represent.

⚠⚠ AND ONE OF THE AGREEING SOCIETIES GRADES ITS RULE AT THE TOP OF THE SCALE. STATED FIRST, AND AT FULL STRENGTH. The EAA guideline’s recommendation 33 — measure haematocrit at three to six months and then annually, discontinue above 54% until it falls to a safe level, evaluate for hypoxia and sleep apnoea, reinitiate at a reduced dose — carries 1 ⊕⊕⊕⊕: a strong recommendation on high-quality evidence.[8] ⚠⚠ AMONG THRESHOLD RECOMMENDATIONS WHOSE EVIDENCE GRADE WAS ESTABLISHED IN THIS REVIEW, EAA RECOMMENDATION 33 IS THE ONLY ONE CARRYING A HIGH-QUALITY GRADE (corrections 461, 465).

⚠⚠ AND THE BASIS THE GUIDELINE CITES FOR THAT GRADE DOES NOT MATCH IT. Section 6.1.2 states that there is a general consensus that a haematocrit above 54% requires testosterone withdrawal, and sometimes phlebotomy, to minimise the risks of venous thromboembolism and cardiovascular events — citing a single narrative review.[8] ⚠ A top grade resting on a stated consensus and one narrative review is a mismatch between the grade and its cited basis. That observation, and not a denial of the grade, is what this review offers against it.

⚠⚠ AND THE SAME GUIDELINE DISCLAIMS THE PREMISE IN ITS OWN TEXT. Having cited the Framingham analysis and the 28-year cohort, it states that since no clear haematocrit threshold was associated with an increased cardiovascular risk, the final decision whether to start testosterone should be left to the physician.[8] ⚠ A sixth guideline saying in its own words that no clear threshold exists, while grading one at the top of the scale. The tension is internal to the document and is reported as such. ⚠ The guideline also records that its haematocrit criteria are derived from the general population not living at high altitudes.

⚠⚠ AND IT CUTS FOR THIS REVIEW MORE SHARPLY THAN THE DISAGREEMENT FRAMING EVER DID. A near-unanimous threshold demands a derivation, and one of the agreeing societies supplies the answer in its own voice. ISSAM attributes the >54% rule to the Endocrine Society and then states: that these recommendations are based on assumptions, and that the clinical significance of a haematocrit above 54% is unknown.[11] ⚠ A society guideline, in print, saying that the clinical significance of the threshold it recommends is unknown. §6 carries this on the genealogy’s on-treatment branch.

What can be said, and it is enough to be going on with. As tabulated here, every society directs that haematocrit be assessed before treatment. Most name a numerical value at which something changes; the Italian societies, as tabulated, name none and direct assessment of global cardiovascular risk with haematocrit included. ⚠⚠ On-treatment monitoring schedules are not established by this tabulation at all — it has no column for them — and they are now taken from the primaries directly for the eight rows read. The values differ, and the actions they trigger differ, and both facts matter to a patient whose result lands between them.

⚠⚠ PRIMARY-GUIDELINE PROVENANCE, STATED ONCE HERE AND NOWHERE ELSE. EVERY OTHER STATEMENT IN THIS DOCUMENT ABOUT WHICH GUIDELINES HAVE BEEN READ DEFERS TO THIS BLOCK.

Eight of the nine rows in Supplementary Table 7 have been checked against their corresponding primary guidance. The EAU row remains unverified as a Cannarella row: the specific edition Cannarella tabulated has not been established. Current EAU guidance (EAU-SRH-2026) has been separately read. ⚠ AUA and ISSAM were read UNTRUNCATED. EAA, BSSM, AACE and SIAMS/SIE were read first-hand to beyond their last relevant recommendation, but retrieval truncated inside their reference lists, so a FULL-LENGTH RE-READ of those four is owed and is recorded in the owed-work table. ✅ EAU Sexual and Reproductive Health (2026) was obtained and read first-hand and untruncated at v4.55–4.56. The EAU row at §5 is updated. EAU-SRH-2026 is held in the source archive (identifier EAU-SRH-2026). ⚠ EAU-SRH-2026 is included as reference [70]; held as archive id 332 (correction 496). ⚠⚠ THE SEARCH FOR IT IS CLOSED BY RULING AND NOT BY EXHAUSTION — that is a fact about this review’s retrieval and carries no judgement about the document’s availability. Correction 168. ⚠ A DIFFERENT EAU DOCUMENT — the Male Hypogonadism guideline of the Dohle panel — IS held and read, and enters as a separate row. Whether it is the edition the tabulation cites is NOT established, and it does not close this row. ⚠ That is a provenance fact about this review’s retrieval, not an access judgement about the documents — corrections 64, 119 and 130.

⚠ The Endocrine Society’s 2006, 2009 and 2017 documents were separately read in full for the lineage at §6. ⚠⚠ An earlier version of this block said the others had not been read. That was true at v4.5 and was left standing while §5 was rebuilt from them at v4.16 — correction 109.

⚠⚠ THE SECOND AND THIRD COLUMNS ARE THIS REVIEW’S RECONSTRUCTION FROM PRIMARIES; THE FIRST IS CANNARELLA’S TABULATION, AND THE THREE ARE KEPT VISIBLY APART. ⚠ The exhibit was demoted at v4.6 to what one meta-analysis tabulates, because most rows were then read through that secondary source. Eight rows are now read in primary and the demotion is lifted for those eight; it stands for EAU.

⚠ VERIFICATION STATE, BY ROW, BECAUSE SOCIETY COUNTS AND ROW COUNTS ARE DIFFERENT DENOMINATORS AND AN EARLIER VERSION MIXED THEM. Supplementary Table 7 carries NINE rows covering TEN societies, because one row represents the two Italian societies jointly.

⚠⚠ EIGHT CANNARELLA ROWS ARE NOW VERIFIED IN PRIMARY: the Endocrine Society, AUA, EAA, AACE, BSSM, ISSAM, ICSM and the combined SIAMS/SIE row. Eight of the nine Cannarella rows have been checked against their corresponding primary guidance. The historical EAU row remains unverified because the specific edition underlying Cannarella’s tabulation has not been established. EAU-SRH-2026 is a separate current-guidance source: it has been obtained, read first-hand and included in the reference list as [70]; held as archive id 332 (correction 496). Deposit gate must rerun against the full 84-entry list before deposit. EAU-SRH-2026 does not verify or substitute for the historical Cannarella EAU row. ⛔ THE EIGHT CANNARELLA ROWS ARE NOW VERIFIED IN PRIMARY. EAU-SRH-2026 is a separately obtained current-guidance document; it enters the corpus as its own bibliographic object (reference [70]).

⚠ PROVENANCE OF THE EIGHT READS, STATED PER SOURCE. ⚠ ICSM was returned truncated inside its reference list, after every recommendation and the whole of its Table 7. AUA and ISSAM were returned untruncated. EAA, BSSM, AACE and SIAMS/SIE were returned truncated, in every case inside the reference list and after the last recommendation and every haematocrit passage; each was read first-hand in the sitting in which the figures above were written. ⚠⚠ A full-length re-read of those four is OWED before deposit, and is recorded in the owed-work table rather than treated as discharged. The AACE correction notice was read untruncated and is an authorship and indexing notice that does not touch the clinical content.

⚠ AND THE AACE ROW CARRIES A CURRENCY PROBLEM INDEPENDENT OF ITS CONTENT. The guideline tabulated is the 2002 update; the tabulation is from 2024. The guideline states of itself that it is a working document and that periodic revisions are inevitable.[9] ⚠ Whether AACE has issued anything since is not established here.

Confidence: clinical effect-certainty grading is not applicable to this documentary comparison. Source access, edition identity and verification status are reported per row, and those determine which transcription and provenance conclusions can be drawn. This section compares sources, judges transcription accuracy and makes documentary findings; an earlier form described it as reporting what a tabulation states, which stopped being accurate as the section grew (correction 543).

6. What the thresholds rest on, and the study that is not there

The apparatus in §5 is operationally specific. ⚠ What was not identified in this review is an appraisal that validates the individual action thresholds against thrombotic outcomes in testosterone-associated erythrocytosis, or that grades the certainty of such threshold-to-outcome relations.

⚠ That is a narrower claim than the one this review made in an earlier draft, and the narrowing is a correction. The guidelines themselves carry formal evidence grading, and [6] carries one such grading in its own text — the EAA’s recommendation 27, at level 2, low; Supplementary Table 7 has no certainty column and carries none.[6] Read in primary, ICSM grades its own haematocrit recommendation at level of evidence 2, grade B.[10] And some describe the evidence on cardiovascular safety as moderate quality or say plainly that it cannot be stated definitively whether testosterone raises or lowers risk.[6] They are not presented by their authors as settled.

What is missing is narrower and more specific: no trial identified here has shown that intervening at a specified haematocrit value reduces thrombosis or any other major clinical outcome in testosterone-associated erythrocytosis.

⚠⚠ That claim must be scoped to this condition, because a threshold-intervention trial does exist in another one. In polycythaemia vera, participants were randomised to haematocrit targets and the lower-target arm had substantially fewer cardiovascular deaths and major thromboses. ⚠ That trial has since been obtained and read; it is reviewed at §9.5.4. ⚠⚠ AND SO HAS THE SOCIETY GUIDELINE THAT ADOPTED ITS RESULT, WHICH THIS REVIEW HAD BEEN REASONING WITHOUT. The British Society for Haematology recommends a target haematocrit below 0·45 in ALL patients with polycythaemia vera, at GRADE 1A, and states that the target — originally derived in 1978 from vascular-event counts at different haematocrit levels — HAS NOW BEEN VALIDATED IN A RANDOMISED CLINICAL TRIAL.[16]
>
> ⚠⚠ THAT IS THE EXISTENCE PROOF THIS REVIEW HAS OTHERWISE LACKED, AND IT IS WORTH MORE TO THE ARGUMENT THAN THE CITATION IS. A numerical haematocrit treatment target CAN be tested against patient-important outcomes, and polycythaemia vera provides a concrete example: here is what one looks like, what evidence it took — a single randomised trial against the threshold itself — and what a society writes when it has that evidence.

⚠ CORRECTION NOTICE, IN ITS OWN PARAGRAPH. An earlier version claimed uniqueness for this target across the whole of medicine. That is a universal negative over every condition in which haematocrit might be targeted, and this review cannot establish it: its search is incomplete, its question is not designed to exhaust other conditions, and the same version demonstrated that its own archive still held relevant material it had failed to identify. ⚠⚠ The existence proof survives entire without the superlative, and the superlative is withdrawn. Correction 150. ⚠ The contrast with every threshold at §5 is the point: those are graded on expert consensus and observational association; this one is graded 1A against a randomised outcome.
>
> ⚠⚠ AND THE SAME GUIDELINE DECLINES A SEX-SPECIFIC TARGET IN TERMS: no study has explicitly defined a gender difference in Hct target. A different target Hct in males and females is not recommended.[16] ⚠ It nevertheless uses sex-specific values for INVESTIGATION — males above 0·52, females above 0·48 — so one society applies a sex-specific INVESTIGATION threshold and an explicitly sex-neutral treatment target, and says why. ⚠⚠ An earlier version wrote diagnostic threshold here, which collapses investigation into diagnosis one paragraph after §9.3.2 separates them. Correction 152. §9.3.2 carries that as a threshold-object finding.
>
> ⚠⚠ It is not transferable evidence for testosterone-associated erythrocytosis, and §9.5.4 refuses the transfer explicitly — now on the same society’s own authority, since its companion guideline sets a DIFFERENT target in every other erythrocytosis it addresses and records that below 0·45 is DETRIMENTAL in one of them.[17] Polycythaemia vera is a clonal myeloproliferative disease with disease-specific thrombotic biology. Testosterone-associated erythrocytosis is a secondary erythrocytosis and does not share that clonal pathophysiology. But its existence means the universal form of this sentence — in any population — was false, and an earlier version of this review asserted it. ⚠⚠ That is not the same as saying nothing has been trialled in trans-masculine people. A published randomised pilot compares testosterone dose reduction against withdrawal in trans men with a haematocrit at or above 50%. ⚠ This review has read it in published abstract and registered protocol only; the full text was not obtained. It is recorded as an owed read, and if it is as described it bears on §8, §9.5, §10, §11 and this section.

Where the numbers trace to

The meta-analysis gives the reason the societies converge. They substantially agree on the importance of haematocrit evaluation because of the association between elevated haematocrit and the risk of cardiovascular disease — and for that proposition it cites one reference: a 1994 report of a general-population cohort with 34 years of follow-up.[6]

⚠⚠ That citation supports a general rationale for concern. It does not establish that any society derived its particular number from that cohort. Those are different genealogical claims, and this review can make only the first from this source. Reconstructing how 48%, 50%, 52% and 54% were each arrived at requires those societies’ primary guidelines and their own citation chains. ⚠⚠ Eight of the nine society rows have now been checked against primary guidance, as detailed in the provenance block at §5. What has NOT been reconstructed is their numerical citation chains; the Endocrine Society lineage is the branch traced furthest here. ⚠ An earlier version said the remaining societies’ primaries had not been read, which was true at v4.5 and false from v4.16. Correction 116.

That is the explicit citation this source gives for the general association between elevated haematocrit and cardiovascular risk. ⚠⚠ It does not establish how any society arrived at 48%, 50%, 52% or 54%, and this review does not claim it does. A general-population cohort, thirty-two years old at the time of writing, ⚠ not designed as a transgender or testosterone-treated cohort, with no trans-masculine subgroup or exogenous-testosterone analysis identified.

⚠⚠ It is not the whole foundation for monitoring haematocrit at all, and this review no longer implies that it is. Trans-specific guidance recommends haematocrit monitoring in its own right — the Endocrine Society’s transgender guideline sets a monitoring schedule and lists haemoglobin and haematocrit among the parameters followed in transgender males — and that recommendation rests on direct evidence that testosterone raises haematocrit in this population, which §9.1 grades ROBUST. ⚠ The question this review presses is not whether to measure. It is what the numbers mean once measured.

⚠ What is established is only what that study is cited for. Whether it supports the weight placed on it is a question this section raises and does not answer.

And the collecting paper does not measure the thing the thresholds regulate

Its population is men with low total testosterone. Its intervention is testosterone replacement. Its outcomes are arterial thrombosis, stroke, myocardial infarction, venous thromboembolism, deep vein thrombosis, pulmonary embolism and mortality. Haematocrit appears in its data-extraction list and its baseline tables. It is nowhere an exposure.[6]

So the paper that collects the haematocrit thresholds is a study of whether giving testosterone causes thrombosis — not of whether raised haematocrit does. On its own question it finds, from its fourteen included randomised trials, pooled in subsets of two to eight per outcome, no detected effect on arterial thrombosis, stroke, myocardial infarction, venous thromboembolism, pulmonary embolism or mortality; from observational studies, reduced risk on several outcomes; and deep vein thrombosis uncertain, with a single randomised trial and an observational odds ratio of 1·44 (95% CI 1·04–2·00).[6]

⚠⚠ AND THE TRIALS THAT NULL RESULT RESTS ON ARE ALMOST ALL AT HIGH RISK OF BIAS, WHICH THE PAPER RECORDS AND THEN DOES NOT REPEAT. Cannarella’s supplementary risk-of-bias table rates thirteen of the fourteen included randomised trials at high risk and one as having some concerns.[6] These are the review authors’ own ratings, not an independent reassessment here. ⚠ Neither the abstract nor the Conclusions mentions them.

⚠⚠ WHAT THEY BEAR ON, AND WHAT THEY DO NOT. They limit confidence in the pooled treatment-effect estimates, alongside imprecision and other outcome-specific limitations. ⚠⚠ They do NOT bear on whether the same paper’s guideline-threshold tabulation faithfully transcribes each society’s recommendation. Trial risk of bias concerns estimates produced by those trials; transcription accuracy is a different appraisal question, answerable only by comparison with each cited guideline — which §6 does row by row. A shared author team is not a bridge between the two. ⚠ Nor does a high-risk rating establish the direction or magnitude of bias.

⚠ The instrument is named, and both instruments were applied. Supplementary Table 5 is headed for the Cambridge Quality Checklist and the Cochrane risk-of-bias tool for randomised trials together, and its rightmost column is the Cochrane one; Quality assessment states that all studies were assessed with the Cambridge checklist and that further assessment was performed for the randomised trials using the Cochrane Risk of Bias scale, citing Sterne’s RoB 2.[6] The some concerns category is RoB 2’s, and it is there because RoB 2 was used. ⚠ Correction 484 recorded the apparent contradiction as unresolved, and it was an artefact of quoting one sentence and stopping at it. The supplementary table has since been obtained and read in full. The earlier refusal is left on the record rather than deleted: it was correct on the evidence then available (correction 694).

Those two facts sit oddly together and the oddness is the point. ⚠⚠ This meta-analysis concludes that testosterone replacement did not show a statistically detectable increase in the thrombotic outcomes it pooled, while its estimates remain imprecise for uncommon venous events and deep-vein-thrombosis risk is not resolved.[6] The same literature maintains an apparatus built to prevent thrombosis from the haematocrit that testosterone produces. Both can be true. The second does not follow from the first, and nothing in this source establishes the link the apparatus assumes.

⚠ An earlier version of this passage read the literature concludes that testosterone replacement is safe from thrombotic risk. That converted one meta-analysis’s conclusion into a field consensus, and this review’s own later sections show why it is not one — wide randomised intervals, sparse venous counts, a higher pulmonary-embolism count in the largest trial, and long-term safety expressly unsettled by the pooled analysis’s own authors.

6.1 Three structural absences

No appraisal of the threshold question. ⚠⚠ A systematic review of this literature does exist, and it is the broadest systematic appraisal identified here of drug-induced erythrocytosis and its management. ⚠ It is not an appraisal of every question adjacent to this one, and this review does not describe it as an appraisal of the field. Published in 2025, it screened 2,036 records, included 45 studies of which 35 concerned testosterone or other androgens, and covered diagnosis, management and outcomes — including ten transgender testosterone studies and both of the trans-masculine cohorts that counted thrombotic events.[18]

Its conclusions independently support the broader concerns motivating this review — that diagnostic and management thresholds are heterogeneous and that thromboembolic risk in testosterone-associated erythrocytosis remains incompletely characterised. ⚠ It does not validate this review’s threshold taxonomy or reconstruct the derivation of individual guideline cut-offs; those are this review’s own synthesis. It states that evidence to support specific haematocrit thresholds is lacking, that management is heterogeneous, and that the risk of thromboembolic events remains unclear in this population. ⚠⚠ It reports thromboembolic events in 0·9% to 2·7% of transgender patients on testosterone, and THIS REVIEW DOES NOT REPRODUCE THAT RANGE AS A RANGE.[18] The two figures rest on incompatible denominators: 0·9% is 5 of 519 across a whole treated cohort, and 2·7% is 3 of 113 within the erythrocytosis-defined subgroup of a 511-person cohort — not 3 of 511. ⚠ They are therefore not comparable estimates of cohort-wide event prevalence, and §9.4.3 keeps them apart for exactly this reason. What the systematic review establishes is that two retrospective trans-masculine reports counted events; the interval between its two percentages is an artefact of pooling a cohort denominator with a subgroup denominator. ⚠⚠ An earlier version of this section reproduced the range unqualified, in a manuscript whose §9.4.3 explains at length why it cannot be reproduced. It applies a formal risk-of-bias instrument to the studies this review also uses.[18]

⚠ So this review no longer asserts an absence here. It cites an independent systematic review reaching the same conclusion — which is a stronger position than an unverified negative. ⚠⚠ What was NOT identified in this review is an exercise that grades the evidence connecting a testosterone-induced haematocrit level to a thrombotic outcome and publishes a certainty rating. The distinction is between appraising the literature, which has been done, and grading the threshold-to-outcome question, which has not.

No trans-masculine outcome study identified here estimates thrombotic risk as a function of haematocrit. Events have been counted in trans-masculine cohorts, and §9.4.3 counts them. No cohort identified here connects a raised haematocrit to an event in a way capable of establishing causation.

And the source population of a guideline is not the same question as the provenance of its number. ⚠⚠ This review distinguishes the two, because an earlier version did not.

The societies tabulated at §5 write about testosterone replacement in hypogonadal men — older, treated for a different reason, selected differently. ⚠ Data from those populations applied to trans populations is flagged inline throughout.

⚠⚠ The trans-specific guideline is a different case. Its 50% classification is stated for transgender males, in a document written for them, and the numerical value carries no citation.[1] ⚠ That is an absence of exposed derivation, not evidence of cisgender provenance, and this review does not convert one into the other. Whether the number was derived from cisgender data, from clinical convention, or from something else is unknown and requires the guideline’s own citation chains to be traced.

⚠ What can be added from elsewhere in this review is that the value it names lies within the reference interval derived by Greene et al. in 79 healthy transgender people receiving stable testosterone therapy, recruited from primary care clinics specialising in transgender care.

⚠ That interval should not be applied automatically outside the population in which it was derived. ⚠⚠ AND A REFERENCE INTERVAL AND AN ACTION THRESHOLD ANSWER DIFFERENT QUESTIONS. A reference interval describes what is usual in a defined population; an action threshold is a level at which something is to be done. A value can sit outside the first without reaching the second, and can sit inside the first while a clinician would still act. ⚠ Testosterone exposure and its duration, ongoing menstrual blood loss where it applies, age and the issuing laboratory’s own interval all bear on which interval is the right comparator, and gender identity is not a substitute for measured hormonal exposure: amenorrhoea and oophorectomy must not be assumed. ⚠⚠ APPLYING AN INTERVAL DERIVED FROM PRE-TREATMENT PHYSIOLOGY CAN MISCLASSIFY AN EXPECTED TREATMENT-ASSOCIATED CHANGE, AND USING A TESTOSTERONE-ASSOCIATED INTERVAL DOES NOT VALIDATE ANY THROMBOSIS THRESHOLD. Neither operation is evidence about outcomes (correction 528). Clinical interpretation of haemoglobin and haematocrit in gender-affirming care also depends on androgen exposure and menstruation status; an interval derived in people receiving stable testosterone does not necessarily apply to people using lower-dose or partial-masculinisation regimens, or to people who continue to menstruate.[19]

⚠⚠ A METHODOLOGICAL LIMIT ON EVERY ABSENCE STATEMENT IN THIS REVIEW. This is a narrative review and not a systematic search. It can report that no eligible study was identified. It cannot establish that none exists, and where an absence is load-bearing this review says so rather than asserting more than its method supports.

6.2 What a reader should take from this

Not that the thresholds are wrong. This section establishes provenance and nothing else. Whether the apparatus is calibrated correctly is graded in Part V.

What it establishes is narrower and harder to argue with: the numerical thresholds and risk classifications are operational values carried in guidance; they answer different clinical questions; and the derivation of the individual numerical values has not been reconstructed. ⚠ An earlier version wrote operational conventions. Convention implies something about how the values arose, and this review’s own finding is that their derivation is unresolved; values is documentary fact and no more. ⚠ Some later literature links their rationale to outcome data from those populations. This review has established the pointing, not the derivation. ⚠ The recommendation to monitor does not. It is made in trans-specific guidance and rests on direct evidence in this population.

Confidence: not applicable — this section is the method, not a finding.

✅ §6 GENEALOGY CORRECTED

§6 — correction to the genealogy passage

Replacing the not verified and not asserted paragraph:

⚠⚠ That cohort has not been obtained through this review’s archive or its available access routes. ⚠ It is publicly indexed — Gagnon, Zhang, Brand and Kannel, American Heart Journal 1994;127:674–682, 10.1016/0002-8703(94)90679-3. What is established is only what it is cited for. ⚠⚠ An earlier version of this section stated that it could not be obtained and had no DOI. Both were wrong: the first converted a practical limit on this review’s access into a claim about the world, and the second was inherited from a DOI this review had itself constructed.

⚠⚠ AND A SECOND GROUP REACHED THE SAME IMPASSE INDEPENDENTLY. A 2022 study of testosterone-associated erythrocytosis in 48,671 men states that the rationale for these cutoffs is not cited in the guidelines, that they appear to come from the same 1994 cohort, and concludes that the currently existing haematocrit cutoffs among testosterone users are arbitrary.[20]

⚠ AND THE TRANS-SPECIFIC GUIDELINE DOES NOT EXPOSE ITS OWN NUMBER’S DERIVATION EITHER. Its 50% classification appears in a risk table with no citation attached to the numerical value.[1] ⚠ The guideline does cite literature for the adverse events it names; what is missing is a derivation of the cutoff.

⚠⚠ AND A THIRD GROUP POINTS AT THE SAME PAPER. The largest study in this field, a cross-sectional analysis of 6,670 trans-masculine people, states that 54% is the level at which there is increased risk of cardiovascular mortality observed in the Framingham Heart Study, citing the same 1994 cohort.[21]
>
> ⚠⚠ AND THE GENEALOGY HAS TWO BRANCHES, NOT ONE, WHICH THIS REVIEW HAS ONLY NOW SEPARATED. The 1994 analysis is pointed to for the general association between raised haematocrit and cardiovascular risk, and the publications that point to it do so when discussing the INITIATION values. ⚠⚠ THAT IS WHERE A CITATION SITS, NOT A DERIVATION. Identifying a source used in a recommendation’s rationale does not establish that any numerical cut-point was derived from it or optimised against it, and this review asserts no such derivation (correction 519). ⚠ A fourth publication now joins the three below: the EAA guideline cites the same Framingham analysis for the observation that men over 64 with haematocrit above 48% had higher coronary and cardiovascular mortality, alongside a 28-year cohort of 670 men in which above 50% carried 1.8-fold coronary mortality.[8] It is the first of the four to attach that source to 48% specifically rather than to 54%.
>
> ⚠⚠ THE ON-TREATMENT BRANCH RUNS SOMEWHERE ELSE ENTIRELY, AND IT CONVERGES. Seven of the eight verified Cannarella rows set the on-treatment action level at 54%, and ISSAM attributes that rule to the Endocrine Society’s guidance — and then says, in its own voice, that these recommendations are based on assumptions and that the clinical significance of a haematocrit above 54% is unknown.[11]
>
> ⚠⚠ THAT IS A SOCIETY GUIDELINE STATING IN PRINT THAT THE CLINICAL SIGNIFICANCE OF THE THRESHOLD IT RECOMMENDS IS UNKNOWN. It comes from inside the guideline literature rather than from a critic of it, and this review had not carried it before.

⚠⚠ WHAT IT CORROBORATES IS NARROWER THAN THIS REVIEW’S CENTRAL INFERENCE, AND THE SCOPE IS STATED RATHER THAN LEFT TO BE INFERRED. ISSAM corroborates ONE branch: that the clinical-outcome basis of the 54% action threshold is unresolved. ⚠ It says nothing about the other branch — whether a thrombotic effect is specifically attributable to testosterone-associated haematocrit elevation in trans-masculine people — and it is a statement about hypogonadal practice outside gender-affirming care.

⚠ AND INDEPENDENT IS THE WRONG WORD FOR PART OF IT. ISSAM states the unknown-significance judgement in its own voice, which is independent. But it ATTRIBUTES the 54% rule itself to Endocrine Society guidance, so it is not an independent derivation of the threshold — it is a second society reporting the first one’s number and declining to vouch for its outcome basis. ⚠ An earlier version called it the strongest independent corroboration of this review’s central claim, which overstated both the scope and the independence. Correction 111. ⚠ ISSAM adds a second observation §9.3.4 wants: that isolated haematocrit elevations can result from insufficient fluid intake on a hot day, and that only REPEATED measures above 54% should trigger action.[11]
>
> Three later publications separately point toward the same 1994 analysis when discussing the origin or rationale of current haematocrit cut-offs, and two of them state that they could not trace the derivation — one concluding that the existing cutoffs are arbitrary. ⚠⚠ An earlier version of this passage said independently. Separate citation is established; independence of reasoning or of citation lineage is NOT — one of the three may have taken the attribution from another, and the lineage has not been traced. The overstatement is corrected here rather than quietly amended, because asserting independence without tracing the chain is the move §9.6 exists to catalogue. ⚠⚠ That convergence makes the paper an important candidate ancestor of the modern thresholds. It does not establish that any society derived its numerical threshold from it. ⚠ The derivation of the individual guideline values remains unresolved until the primary citation chains are traced, and this review has not traced them.

PART IV — THE PROPAGATION QUESTION

7. How a claim travels, and why this review follows it

Every section here that examines a threshold ends up examining a citation. Not because the citations are careless. ⚠ Because the numerical derivation of several thresholds is unresolved, later papers encounter them as inherited clinical numbers and then point retrospectively to older observational evidence as rationale, and the distance between the two is covered by papers quoting papers.

That is an evidential question rather than a bibliographic one, and this section says why.

When a paper reports what another paper found, it does three things at once. It selects, choosing which of the source’s findings to carry. It compresses, and a table becomes a sentence. And it reframes, so that a hedged result acquires the confidence of the sentence repeating it. Each step is legitimate on its own. The cumulative effect is not, because nothing in the chain records what was dropped. A reader arriving at the fourth paper meets a claim that looks settled, and cannot see the qualifications shed to make it look that way.

The failure is invisible from either end. Read the source and it says what it says, carefully. Read the citing paper and it cites a real source that really does contain something like the claim. The defect exists only in the relation between them. That is why it survives peer review, and why finding it means reading both sides.

So this review keeps a ledger, and §9.6 is it. Each entry records one instance: what the source says, what the citing paper says it says, and what was lost in between.

The standard the ledger works to is strict, and that is why the ledger is short. An entry asserting that one source mischaracterised another requires both sides read first-hand, both genuinely open. A source’s own account of what it did is not evidence about what it did. Neither is a systematic review’s account of it. Candidates that could not be tested on those terms were parked rather than entered. The ledger records that it is a floor and not a survey. It contains what could be verified, not what could be found.

What this section does not do. It does not rehearse the instances. §9.6 carries those, in a fixed order, and they should be read there. It does not claim that the propagation found here is representative of the field. And it makes no charge against any author. The mechanism described above is what careful people do under ordinary citation practice. This review has done it too, and several entries in its own correction record have the same shape.

A constraint on this section’s future, recorded now rather than discovered later. The ledger’s entries are numbered once and its order is fixed. If later tranches add instances, this section must absorb them without renumbering what is already there.

Confidence: not applicable — this section is the method, not a finding.

8. The evidence map

⚠⚠ REGENERATED FROM PART V IN THE CURRENT ASSEMBLY, AND NOT MAINTAINED BY HAND. The previous map was written before the coagulation and prevalence evidence base was extended and was never regenerated with it: it gave the largest prospective dataset for haematocrit change as 192 where §9.2 carries approximately 300, gave the largest coagulation cohort as 312 pooled where §9.4.6 carries 348 prospective, and described the thrombotic-event corpus as three cohorts where §9.4.3 integrates five. ⚠ That is correction 38, and the remedy is regeneration rather than amendment. Every cell below is taken from the Part V subsection named in the row.

⚠ One map only. An earlier assembly carried a superseded map followed by a block of corrected rows. Two live scientific states is not transparency, and the superseded map is removed rather than reproduced.

⚠ The map is no longer a map of absence. Four rows now carry randomised or pooled evidence where an earlier version read cohort, mostly retrospective or not yet established. ⚠⚠ And the correction that produced this version ran AGAINST this review’s position: the stale map understated the direct trans-masculine evidence base relative to the manuscript’s own Part V.

PART V — THE OUTCOME, GRADED

9. Erythrocytosis — the change is certain, the consequence is not

Testosterone raises haematocrit. Across the cohorts reviewed here, every study reports an increase, and nothing in this review disputes it. ER-007 states the effect; this section grades it, and grades what does and does not follow from it.

The distance between those two things is the substance of what follows.

9.1 How testosterone raises the red cell count — a mechanism its own authors have twice reframed

That testosterone increases haemoglobin, haematocrit and erythrocyte indices is not in question. It is among the most reliably reproduced effects in this literature, it is dose-related, and it holds across the formulations and age groups represented in the studies reviewed here, with a dose–response demonstrated in controlled dose-ranging work. What is not established is how it happens — and the account most often cited has been reframed, twice, by the laboratory that built it.

This matters more than a mechanistic curiosity usually would. The apparatus described in §9.2 and §9.3 treats haematocrit as the quantity that matters. Whether it is the right quantity depends on what testosterone is doing to the red cell compartment, and the honest answer is that the field does not yet agree.

The effect itself

The cleanest dose–response comes from a pair of trials that suppressed endogenous testosterone with a GnRH agonist and then gave weekly testosterone enanthate across a twenty-four-fold dose range for twenty weeks. In sixty-one men aged 18 to 35, haemoglobin fell slightly at the lowest doses and rose by 14.2 g/L at 600 mg weekly, with the change correlating with achieved testosterone at r = 0.66.[73] In sixty men aged 60 to 75 on the identical protocol, haemoglobin rose by 20.9 g/L at 125 mg and 29.4 g/L at 600 mg, and haematocrit rose from 0.41 to 0.48 at 125 mg and 0.40 to 0.50 at 600 mg. After adjustment for achieved testosterone, the increments in both were significantly greater in the older men.[74] The secondary analysis of the same cohorts found the rise linear and dose-dependent in both age groups, with achieved testosterone correlating with change in haematocrit at r = 0.63.[75]

The finding replicates outside that design. Six months of testosterone gel in older men with mobility limitation raised haemoglobin 1.1 g/dL and haematocrit 4.4 points — 7% and 10% from baseline.[42] Three years of a transdermal patch trial found no significant change in erythropoietin at either measurement point; sampling was at baseline and 36 months only, so the study cannot address a transient early effect.[76]

And the size of that effect is where a threshold argument first runs into difficulty. A rise of about four haematocrit points from a baseline in the low forties lands a population in the mid-forties. On average it reaches none of the cut-offs catalogued in §9.2. ⚠ Threshold crossing is a function of baseline, individual treatment response and other contributors to erythrocytosis. The average effect is one component of it, and the relative contribution of the components has not been quantified here.

And the age difference is the part that transfers, while the dose ceiling is the part that does not. In the older cohort, six men exceeded 54% haematocrit — one of them at 125 mg weekly, the dose the authors recommend as the best trade-off between benefit and harm, and above many commonly used weekly-equivalent masculinising regimens, though ester, interval and achieved concentration make any single multiple unstable.[74] ⚠⚠ The younger-men report does not establish that no participant exceeded 54%: it reports haemoglobin and not haematocrit, and supplies neither the adverse-event definition nor the complete tally the deduction would require. Any age-specific threshold-crossing claim must be attributed to a source that reports it explicitly (correction 535). An earlier form asserted that no serious adverse events occurred and that exceeding 54% was classified as one, and deduced that no young man crossed it, even at 600 mg. That deduction rests on the young-men trial having reported its serious adverse events completely, which this review has checked and cannot confirm: the trial reports no serious-adverse-event tally, does not define exceeding 54% as an adverse event, and reports no haematocrit anywhere, only haemoglobin (correction 503). What supports it is a pattern of reporting rather than a stated result, and it appears nowhere in the young-men trial’s own report, which presented haemoglobin and did not report haematocrit.[73]

⚠⚠ AN EARLIER VERSION SAID THAT TRIAL DID NOT MEASURE HAEMATOCRIT. THAT WAS WRONG AND IS CORRECTED HERE RATHER THAN QUIETLY AMENDED. Haematocrit WAS measured in the young cohort. The secondary erythropoiesis analysis (Coviello et al. [75]) is expressly a secondary analysis of haemoglobin AND haematocrit data from that study, it plots haematocrit for the young men, and it reports their peak-timing distribution by dose group.[75] ⚠ Both papers were opened untruncated in the step this correction was written.

⚠ The deduction above is nonetheless unaffected, and the reason is worth stating. What the secondary analysis publishes for the young men are GROUP MEANS, from which no individual maximum is recoverable; and its Introduction attributes the claim that no young man exceeded 54% to the two parent trials, one of which contains no haematocrit anywhere at all.[75] ⚠⚠ So the claim remains an inference from an adverse-event classification rather than a reported result, and the correction removes an overstatement this review was making about a source while leaving intact the argument that rested on it.

⚠⚠ No young participant in that protocol was inferred to have crossed 54% over twenty weeks, whereas six older ones did. That supports an age-dependent difference in the erythropoietic response under that protocol. It does not establish a biological ceiling, at any age, outside that dose, duration and population. It should not be cited as though testosterone cannot push a young person past 54%; it shows that in these young men, at these doses, over twenty weeks, it did not.

The erythropoietin question — three human datasets that do not agree, and a fourth in mice

The oldest explanation is that testosterone stimulates erythropoietin.

In the graded-dose secondary analysis,[75] there was no significant change in erythropoietin at any dose, in either age group (P = 0.71 younger, P = 0.64 older), and no change in soluble transferrin receptor. Changes in erythropoietin did not correlate with changes in total or free testosterone. The authors proposed direct marrow effects instead.[75] In the three-year patch study erythropoietin again did not move (treatment-by-time β = 0.24, SE = 2.16, P = 0.91), and entering it as a time-varying covariate did not account for the haemoglobin effect.[76]

Against those, the six-month gel trial found erythropoietin rising from 13.5 to 21.3 mIU/mL at one month — a 58% increase [42] — still significant at three months, trending back toward baseline by six while remaining unsuppressed against an elevated haemoglobin. At every level of haemoglobin, erythropoietin was about 30% greater after treatment than at baseline. The authors read this rightward shift as a recalibrated set point rather than simple stimulation.[42]

And the two null studies are not of equal weight, which is what a summary loses. The graded-dose study sampled repeatedly across twenty weeks and genuinely conflicts with a sustained rise. The patch study measured erythropoietin at exactly two points, thirty-six months apart, and therefore cannot exclude a rise at one month that had resolved by six. It is compatible with the set-point account; it does not count against it. Treating the two as joint corroboration would be an error, and an error in the direction this review has already had to correct.

The animal work bears on the shape of the disagreement. In Guo et al. [77], testosterone raised renal erythropoietin mRNA roughly twofold at 72 hours, returning to baseline by two weeks. That is a transient rise of the kind the gel trial reports and the patch trial could not have detected. It does not resolve the human conflict — the graded-dose study sampled early enough that a sustained elevation should have been visible — but it makes a transient response more plausible than the human data alone suggested.

There is also a reconciliation available that is not a sampling argument, and it should be stated as the untested candidate it is. The exposures differ profoundly: the null studies rendered healthy men hypogonadal and gave supraphysiological doses, or lifted an already low-normal level with a patch; the positive study gave gel to comorbid older men, moving testosterone from frankly low into mid-normal. Whether the erythropoietin response depends on the starting point rather than the dose has not been tested.

The hepcidin route — called important, then called non-essential, by the same laboratory

Because erythropoietin appeared not to change, attention turned to iron availability. Hepcidin restricts it; suppress hepcidin and more iron reaches the marrow.

The suppression is well demonstrated. In the graded-dose cohort,[78] serum hepcidin fell substantially within a week, with every dose group declining. At steady state the effect was dose-graded — 97.2% of baseline at 25 mg weekly, 71.3% at 50 mg, 47.6% at 125 mg, 34.2% at 300 mg, and 38.6% at 600 mg, that is, no further suppression above 300 mg — with roughly a 14.9% fall per 100 ng/dL of testosterone. Older men began with nearly twice the hepcidin of younger men and suppressed it 1.8-fold more.[78] The gel trial reproduced this in a different population, with hepcidin down 49%, ferritin falling, soluble transferrin receptor rising and interleukin-6, C-reactive protein and tumour necrosis factor-α all unchanged, so the hepcidin suppression is not explained by a change in these inflammatory markers ⚠⚠ — though both papers leave suppression of inflammation on their own list of candidate pathways, and the earlier states that the mechanism of hepcidin suppression remains unknown (correction 502).[42]

And the most quotable figure in that set is the least informative. The 60% fall within one week coincides with a GnRH-induced testosterone flare that pushed every dose group above 1000 ng/dL before levels settled. The authors say so. It describes a transient supraphysiological spike common to all arms, not the behaviour of testosterone at therapeutic doses.

The transcriptional mechanism has been worked out in some detail. Testosterone suppressed hepatic hepcidin mRNA by about 70% within 96 hours in mice,[77] sustained over two weeks. Liganded androgen receptor associates with Smad1 and Smad4 and reduces their binding to both BMP-response elements in the hepcidin promoter; a BMP inhibitor produced no additional suppression beyond testosterone alone; the effect was reversed dose-dependently by an androgen-receptor antagonist; and messenger RNA stability was unaffected, placing the action at transcription. Downstream, splenic ferroportin rose, splenic iron stores fell, and labelled iron incorporation into red cells increased.[77]

Then the route was tested against genetics twice, with opposite emphases.

In mice engineered for constitutive liver-specific hepcidin over-expression,[77] testosterone failed to raise haemoglobin or haematocrit, and in castrated males the haemoglobin rise was abolished entirely. The conclusion drawn was that hepcidin regulation is important for mediating the effect.[77] Seven years later, in mice with hepcidin knocked out, testosterone raised haematocrit anyway — 58.1% from 53.3% against 53.8% from 49.7% in wild-type — and it did so again in mice whose hepatocytes lacked the androgen receptor. The conclusion drawn there was that hepcidin suppression is not essential.[79]

⚠ The two results are not necessarily contradictory: these models show that altering hepcidin regulation can modify the erythropoietic response in some experimental settings, but they do not establish the quantitative or necessary role of endogenous hepcidin in humans. Clamping hepcidin permanently high blocks the response; removing hepcidin altogether does not prevent it. What changed between the two papers is not the biology but which model was built and therefore which claim was licensed. The earlier paper had already noted that hepcidin deficiency causes polycythaemia in neither mice nor humans, and that suppression alone could not fully explain the effect.[77] The later paper is explicit that its result should not be read as hepcidin playing no role.[79] A reader taking either abstract on its own would come away with a firmer view than the pair of them supports.

And a truncation runs through almost all of it. Six older participants in the hepcidin study were discontinued above 54% haematocrit and excluded from analysis;[78] the graded-dose trials excluded anyone entering above 48% and removed anyone crossing 54%;[74] and the trial underlying the gel data excluded baseline haematocrit above 48% and contained exactly one participant who ever exceeded it.[72] Four of the five human cohorts in this section were constructed so that people reaching the haematocrit clinical practice worries about were absent from them, by design.

The iron question this population raises, and which no study identified in this review has answered

Every iron mechanism above was characterised in people who do not menstruate. If iron availability governs any part of the erythropoietic response, then a background of menstrual iron loss — present in many trans-masculine people before testosterone, and for a variable period after starting it — is an obvious candidate modifier. The knockout work makes the question sharper rather than softer: hepcidin is dispensable when iron is plentiful, and the authors of that study say explicitly that it may matter more where iron availability is constrained.[79]

The one adjacent test in the right population does not support the iron route, and it is worth being precise about why. A retrospective cohort of 511 trans-masculine patients found polycythaemia in 113, and reported prior oophorectomy as a novel association — odds ratio 2.39 (1.56–3.68) on univariate testing. The adjusted estimate is 1.582 (0.974–2.568), p = 0.064 — imprecise, and compatible with little association and with a substantial one. Only age, body mass index and dose met the study’s significance threshold.[30] ⚠ An estimate that crosses a significance threshold after adjustment has not been shown to be absent; it has been shown to be imprecisely measured (correction 522).

And the mechanism its authors propose is not an iron mechanism at all. They suggest that removing the ovaries removes a site of aromatisation, leaving more testosterone unconverted. That is a testosterone-exposure account, not an iron-loss account, and the graded-dose study bears on it directly: change in haematocrit did correlate with treatment oestradiol — ρ = 0.48, p = 0.0003 in young men and ρ = 0.68, p = 0.0001 in older men — but oestradiol tracked total testosterone closely in both groups, ρ = 0.65 in young men and ρ = 0.91 in older men, and the correlation did not survive adjustment for testosterone. Coviello reports that adjusted null for haemoglobin as well as haematocrit.[75] So the single adjacent test of the aromatisation route detects nothing once testosterone is accounted for ⚠ — which is not the same as establishing that the route contributes nothing, and adjustment for a variable this collinear cannot isolate an aromatisation-mediated effect in either direction (correction 522) — though an adjustment cannot cleanly separate two variables that collinear, so the adjusted analyses did not establish an independent association and do not isolate an aromatisation-mediated effect. ⚠⚠ A DIFFERENCE IN STATISTICAL SIGNIFICANCE, OR IN HOW COLLINEAR TWO VARIABLES ARE, DOES NOT QUANTIFY A DIFFERENCE IN EVIDENTIAL WEIGHT BETWEEN THE AGE GROUPS. An earlier form said the null carried real weight in young men and much less in older men; nothing here measures that (correction 548).

The iron pathway in this population is therefore plausible and directly untested. ⚠⚠ The one adjacent analysis tested a different proposed mechanism, and therefore neither supports nor refutes it — an earlier version of this section treated it as evidence against, which it cannot be. No study identified in this review has measured iron status, menstrual status and haematocrit together in trans-masculine people on testosterone. That is a gap, not a negative finding.

The one dataset in the right population — and an unexpected corroboration

Almost everything above is men, mostly older, several of them unwell, plus mice. One study here measured red cells in trans-masculine people: fifteen adolescents assigned female at birth, aged 13 to 16, on subcutaneous testosterone cypionate with dose escalation, sampled at baseline, one month and twelve months, seven also on a GnRH analogue.[80]

It reports haematocrit rising — a minor but significant increase, with no absolute values given. What it adds is metabolic. 2,3-diphosphoglycerate rose with treatment duration, which stabilises deoxygenated haemoglobin and improves the red cell’s capacity to off-load oxygen. Red-cell iron and zinc content fell during treatment ⚠ — the authors read this as consistent with erythropoiesis proceeding without an increase in dietary iron uptake. Measurement was by inductively coupled plasma mass spectrometry on red-cell preparations; no serum or plasma iron or zinc was measured — red cells circulate, so circulating was the wrong word for the compartment excluded, and dietary intake was not measured either. ⚠⚠ These measurements do not establish plasma iron availability, and the iron route remains directly untested in this population (corrections 500, 547). ⚠⚠ An earlier version wrote consistent with erythropoiesis outstripping intake, which asserts an intake–supply mechanism nobody measured. 2-oxoglutarate rose, and the authors read this as potentially relevant to HIF regulation and as possible compensatory counter-regulation. ⚠ These are red-cell metabolomic measurements: they do not demonstrate HIF activity or any downstream tissue response, and an earlier form asserted an antagonism rather than attributing an interpretation (correction 547).

And the same two-part signature appears in the mouse work, from a different group using different methods. Testosterone raised erythrocyte 2,3-BPG in mice, while HIF1α and its downstream targets were down-regulated and pimonidazole staining showed less kidney hypoxia — the authors concluding that testosterone increases net tissue oxygen delivery rather than acting through hypoxia sensing.[77] ⚠⚠ The convergence is PARTIAL and the two datasets do not measure the same things. Both show changes in red-cell oxygen-handling pathways in the same direction. The MOUSE study additionally demonstrates reduced renal hypoxia signalling directly, by HIF1α targets and pimonidazole staining; the human study measures red-cell metabolites in fifteen uncontrolled adolescents and demonstrates no tissue-level effect at all. ⚠ An earlier version wrote two species, two laboratories, two methods, one pattern, which claimed an equivalence of demonstration the human data do not supply.

That convergence deserves to be stated plainly, and so does its limit. It supports the proposition that testosterone-associated erythrocytosis is accompanied by biochemical changes compatible with altered red-cell metabolism and oxygen unloading, and it makes the question of whether a haematocrit number means the same thing under testosterone as it does otherwise a question with biological grounds rather than a bare speculation — ⚠ without establishing that its thrombotic meaning differs. It does not show that the adaptation is protective. No study in this section measured a clinical outcome of any kind, the human data are fifteen adolescents in a research letter without controls, and intersubject variation there was wide enough that circulating testosterone correlated poorly with red cell metabolites.

What this section establishes

Testosterone increases haemoglobin, haematocrit and erythrocyte indices, dose-dependently, across formulations and ages, and more in older men than younger ones for the same achieved testosterone. ⚠ None of the cited studies quantified total red-cell mass; §9.3 sets out why haematocrit does not establish it. That is as well evidenced as anything in this review.

By what route is not established. No significant increase in circulating EPO was detected in the largest human dose–response study at its sampled time points; an early transient rise was detected in a subsequent gel trial and in mice; and untestable in the one long-term study. Hepcidin suppression is real, dose-graded, transcriptionally characterised and larger in older men — and clamping it constitutively high prevented the response in one genetic model, while eliminating it did not prevent the response in another. Direct marrow effects, altered red cell survival and erythropoietin-receptor modulation remain proposed and untested in humans.

None of the mechanistic work was done in trans-masculine people, with the single exception of a fifteen-participant metabolomic letter in adolescents that measured no outcomes. Whether the mechanism differs in trans-masculine people, in whom menstrual iron loss, gonadal status and endogenous hormone milieu differ, with or without prior pubertal suppression, with menstrual iron loss in the background, is not established — not because studies found no difference, but because the studies have not been done.

Certainty of the evidence — mechanism

Causal attribution, testosterone to increased haemoglobin, haematocrit and erythrocyte indices — ROBUST (within this review’s editorial grading system; not formal GRADE). Consistent across every formulation, dose range and age group examined, with a clear dose–response in the trials designed to test one, and reproduced in two species.

Causal attribution, the mechanism itself — NOT ESTABLISHED. Human datasets disagree on erythropoietin. ⚠⚠ On the best-characterised alternative route: clamping hepcidin constitutively high prevented the erythropoietic response in one genetic model, and eliminating hepcidin did not prevent it in another — showing that altering hepcidin regulation can modify the erythropoietic response in some experimental settings, without establishing the quantitative or necessary role of endogenous hepcidin in humans. ⚠ An earlier version of this block read necessary in one genetic model, dispensable in another. Preventing normal hepcidin regulation from operating shows that the response CAN BE CONSTRAINED; it is not clean identification of necessity, and the prose above this block was already the more careful of the two. A genuine open question, not a gap in the reading.

Observed effect size — MODERATE. The average rise is well characterised — roughly 1 g/dL of haemoglobin and about four haematocrit points at replacement doses over six months, larger in older men. But it is drawn almost entirely from cohorts excluded above 48% at entry and truncated above 54% during, which removes precisely the individuals whose magnitude matters most.

Population directness — indirect. Men, predominantly older, several cohorts selected for comorbidity or mobility limitation, plus mouse models. The single trans-masculine dataset is fifteen adolescents in a research letter with no outcome measures.

9.2 How much it rises, and how common it is — two questions, and the literature answers the second badly

9.2.1 The size of the rise is the better-established quantity, and it is the one the review has been missing

A meta-analysis estimates the change in haematocrit over the first year of testosterone in trans-masculine people at roughly four to five points. ⚠⚠ Selected primary datasets are shown alongside it for scale and design comparison. Some may be components of the pooled estimate: the pooled review searched to April 2023 and included seven studies, and this review has NOT checked which of the primaries below are among them. ⚠ They are therefore not treated as independent confirmation.

The two largest have almost entirely overlapping intervals. Haemoglobin rises correspondingly: +1·48 g/dL pooled (1·17–1·78)[23], with negligible heterogeneity.

⚠⚠ AND THE POOLED ESTIMATE THIS REVIEW QUOTES COMES FROM A REVIEW WHOSE CONCLUSION NAMES A QUANTITY IT DID NOT POOL. [23]’s own stated parameters are the international normalised ratio, fibrinogen, activated partial thromboplastin time, plasminogen activator inhibitor-1, haemoglobin and haematocrit. Viscosity is not among them, and was not a pooled endpoint. Its conclusion nonetheless states that testosterone therapy increased blood viscosity.[23]

⚠ What the measured finding supports. A pooled haematocrit rise of 4·39 points provides a physiological basis for CONSIDERING viscosity. It does not quantify it. ⚠ This review runs the same distinction elsewhere — haematocrit is a concentration measure, and whole-blood viscosity also depends on shear conditions, plasma composition and erythrocyte deformability.

⚠ What is NOT claimed here (correction 485). This review has not obtained [23]’s extraction table and does not assert that no included study measured viscosity. The claim made is about what the review pooled and what its conclusion states, which its abstract establishes on its own.

⚠ Its search covered PubMed, Scopus and the Cochrane Library. Embase is not among them.[23] ⚠⚠ This review is in no position to treat that as disqualifying: its own search section records Embase returning a login redirect and then HTTP 403, not searched to protocol. ⚠ But a shared omission does not make two searches equivalent — dates, concepts, eligibility and citation-chasing differ — and neither the extent of either gap nor its consequences is established here.

⚠⚠ AND THE PAIRED AMSTERDAM ANALYSIS SITS ABOVE THE POOLED ESTIMATE, WHICH IS WORTH STATING RATHER THAN AVERAGING AWAY. Its +6 points at twelve months exceeds the pooled +4·39 and the prospective +4·9, and its confidence interval (5 to 7) does not overlap the pooled point estimate.[27] ⚠ It is a paired within-person design in 100 people from one centre, and it is NOT independent of the other rows: it draws on the Amsterdam arm of the same network. ⚠⚠ This review does not pool it, does not treat it as confirmation, and does not treat the difference as a finding — the estimand, the centre and the analytic sample all differ, and haematocrit was available at both time points for fewer participants than the coagulation measures.[27]

⚠ BUT A CONTEMPORANEOUS TESTOSTERONE CONCENTRATION PREDICTS LITTLE OF THE VARIATION BETWEEN INDIVIDUALS. In 6,670 trans-masculine people already receiving testosterone, a cross-sectional regression of haematocrit on serum testosterone gives R² = 0·029. ⚠⚠ That is a statement about between-person prediction at steady state. It does not quantify the causal contribution of starting testosterone to the within-person rise, which this review grades ROBUST from randomised and longitudinal evidence, that grade belonging to the combined experimental evidence while causal identification specifically within the direct trans-masculine evidence is graded MODERATE at §10 (correction 665). Mean haematocrit spans just 41·84% to 45·68% across the entire testosterone range from below 50 to above 1,000 ng/dL. Statistically overwhelming, and 3·8 percentage points wide.[21]

⚠⚠ AND A SECOND SOURCE AGREES, BY A DIFFERENT ROUTE. In 79 trans men on stable testosterone, haemoglobin was unrelated to serum testosterone within the cohort (p = 0.42), 90% of whose values lay between 2 and 6 ng/mL — the authors conclude that at steady state minor fluxes in testosterone will not have a transient effect on erythropoiesis.[19] Across these two sources, one of 79 and one of 6,670, a single contemporaneous serum testosterone measurement is a weak between-person predictor of haematocrit at steady state. ⚠ Weak prediction is not absence of association: the larger study detects significant variation in haematocrit across testosterone categories. ⚠ That bears on how far a target concentration can stand proxy for a blood count between individuals. It does not bear on whether changing an individual’s dose changes theirs.

⚠ Route and serum concentration are each independently associated — intramuscular 44·96% against transdermal 43·41% — and both remain significant when controlled for one another.[21] ⚠⚠ This review does not inherit that paper’s conclusion that the findings should help allay concerns. That is an editorial judgement, not a result.

⚠⚠ The rise happens early. The prospective cohort places most of it in the first three months — +2·7 points (1·94–3·29) — with a peak at twelve months and a one-point fall by eighteen.[22]

⚠ From a baseline in the low forties, an average four-to-five-point rise ends in the mid-to-high forties. Threshold crossing therefore falls disproportionately on those who begin higher, respond above average, or acquire other contributors — ⚠⚠ and the relative weight of those components is not estimated by any source reviewed here. Baselines in these sources are 41·1, 42·0, and 35·0 and 36·0.[22][24][25] A cohort beginning at 42·0 arrives at 46·2 in twelve weeks[24]; one beginning at 35 arrives near 38. Neither reaches 48.

9.2.2 Reported frequencies depend on the threshold, observation period, ascertainment method and measurement pathway

Several reviews cite a prevalence near 11%. The underlying definitions differ, and they are not measuring the same thing.

Proportion meeting each study’s own criterion, during that study’s observed follow-up

⚠⚠ AND A FOURTH DEPENDENCY APPLIES TO EVERY ROW ABOVE AND TO EVERY THRESHOLD IN §9.3: THE MEASUREMENT PATHWAY AS SPECIFIED — THE DEVICE OR METHOD, THE SAMPLE TYPE, AND WHETHER THE SOURCE OF THE THRESHOLD USED THE SAME ONE. Haematocrit is a defined measurand whose operational estimates differ by method and laboratory. In 150 duplicate venous samples, centrifuged microhaematocrit read 0·467 against an automated analyser’s calculated 0·448 for the same blood, at P = 6·8 × 10⁻²⁵.[41] ⚠⚠ AND THE DIFFERENCE IS NOT A CONSTANT AND MUST NOT BE READ AS ONE: ACROSS FOUR CENTRES THE MEAN DIFFERENCE RAN FROM 0·5 TO 3·8 PERCENTAGE POINTS AND WAS NOT STATISTICALLY SIGNIFICANT AT ONE OF THEM. THESE ARE CENTRE-SPECIFIC MEAN DIFFERENCES: neither individual limits of agreement nor a universal conversion, and a pooled figure is not a correction that can be applied to a patient (correction 668). Those authors recommend running both methods in parallel LOCALLY rather than applying a fixed correction, and reported IN 1999 that the international standardisation body HAD NOT THEN RECOMMENDED CORRECTING FOR TRAPPED PLASMA — current standards have not been checked here.[41] §9.3.4 sets this out in full; THE DEPENDENCY IS THAT THE PATHWAY MATTERS AND IS LOCALLY VARIABLE, NOT THAT A KNOWN QUANTITY CAN BE SUBTRACTED. ⚠ This is not hypothetical for this review’s own population. The only randomised trial of dose reduction in trans men with erythrocytosis registered that haematocrit would be measured by automated counting on a named instrument, a Siemens Advia 2120i — SO ITS 50% ENTRY CRITERION AND ITS ABOVE-54% EXCLUSION ARE ANALYSER VALUES, not centrifuged ones.[65] ⚠⚠ ACCORDINGLY, EVERY THRESHOLD IN THIS REVIEW SHOULD BE READ AS MEASURED BY THE METHOD SPECIFIED — AND §9.3.4 RECORDS THAT, FOR THE THRESHOLD-SETTING SOURCES, NO METHOD IS SPECIFIED AT ALL. §9.3.4 IS THE FULL TREATMENT AND WAS WRITTEN BEFORE THIS INSERTION; WHAT IS ADDED HERE IS ONLY THAT THE DEPENDENCY BELONGS ALONGSIDE THE OTHER THREE, WHERE FREQUENCIES ARE FIRST COMPARED (correction 645).

⚠⚠ REPORTED FREQUENCIES DEPEND ON THE THRESHOLD, OBSERVATION PERIOD AND ASCERTAINMENT METHOD. Several of these studies report the proportion of participants with at least one observed threshold crossing. Within the same observation window, measurement set and threshold definition, that is equivalent to classifying participants by whether their observed maximum crosses the threshold. ⚠⚠ THAT EQUIVALENCE DOES NOT MAKE THE REPORTED COHORT ESTIMATES DIRECTLY COMPARABLE: thresholds and operators, observation periods, testing frequency, confirmation requirements, populations and treatment regimens all differ across them (correction 555). ⚠ A cross-sectional measurement describes status at the sampled time. A time-to-first-crossing analysis estimates cumulative risk by a specified time while accounting for censoring under its assumptions — which may address the SAME underlying time-specific risk as a crude proportion under complete follow-up. THE ESTIMATOR ALONE DOES NOT DETERMINE WHETHER THE TARGET QUANTITY DIFFERS.

⚠ Each cohort’s own operator is in the row above; the columns are values, not criteria. This instruction was carried in a sixth header cell of the table above, in a table with five columns, and no renderer could display it (correction 703). — | — (maximum observed 56%) |

⚠⚠ AN EARLIER VERSION OF THIS TABLE DESCRIBED THE PROSPECTIVE COHORT AS A FIXED TWELVE-MONTH WINDOW AND RECORDED ZERO PARTICIPANTS AT 54%. BOTH WERE WRONG. Its proportions are ever during the investigation period, which ran to 36 months with unequal follow-up between sites; and one participant reached 54·0%, the highest value measured in the cohort.[22][34] ⚠ The error made the metric contrast tidier than the sources allow, and it is recorded here rather than silently amended.

⚠⚠ THESE FIGURES ARE NOT RANKED HERE. ⚠ The observed percentages differ, but these data do not establish how much of the spread reflects ascertainment and endpoint definitions, sampling variation, treatment differences or population differences. They are not a comparative ranking of formulation-specific or biological risk, and NO NUMERICAL ATTRIBUTION OF THE SPREAD TO ANY OF THOSE EXPLANATIONS IS MADE HERE (correction 555). A cross-sectional metric would be expected to yield a lower proportion than an ever-crossed one, since a snapshot cannot count anyone who crossed and came back down. ⚠ An earlier form set the largest cohort against the lowest figure as though cohort size explained the spread. It does not: the metrics differ, and a ranking across non-commensurable estimands asserts a comparison this section spends four paragraphs denying (correction 520). ⚠⚠ That is an expectation, not a demonstration: population, formulation, confirmation rules, testing frequency and laboratory method also differ between these cohorts, and this review does not claim the metric explains the whole divergence. Its authors say so — it is possible that haematocrit levels would have regressed to the mean if observed over time — and a separate cohort measured exactly that fall, 0·04 in the first year, irrespective of any intervention.[21][35]

The contributions of endpoint definitions, ascertainment, sampling variation, population differences and treatment differences to the observed spread have not been established. This review does not attribute a specified part of that spread to any one explanation.

And at a fixed threshold, in adequately sized studies, the figures converge tightly. An independent systematic review tabulating nineteen studies by threshold found that, restricted to cohorts of at least a hundred participants, severe erythrocytosis at 54% ranged from 0·5% to 2·3% — six studies, from 125 to 6,670 people. The 0% and 10% outliers in that table come from cohorts of 24 and 31.[34]

⚠ That review searched a single database and retrieved 36 records, against 152 for a second and 2,036 for a third.[34][36][18] Its table is valuable; its claim to completeness is not.

The same applies to magnitude. One records study reports a mean rise of 6·84 points — from 40·2 to a peak of 47·0.[29] That is a change from baseline to the observed peak rather than a change measured at a fixed twelve-month visit — and the mean of 31·2 months the source reports is the TIME TO PEAK, not the cohort’s follow-up duration, and it is not in conflict with the +4·9 and +4·39 above.

These are proportions, not incidence rates, and no pair can be set beside another without stating threshold, confirmation requirement, denominator, exposure metric and follow-up. The single-versus-repeated distinction is not merely a matter of reliability: one elevated value and sustained elevation are different phenotypes, and the second is also a function of how often someone is retested.

Relative and absolute measures must remain separate. Okano and colleagues [36] describe an absolute-risk range of approximately 4%–13%. Those endpoints resemble the confidence limits on the hazard ratio they report, but the article does not demonstrate how the absolute-risk range was calculated. A hazard-ratio confidence interval cannot be interpreted as an absolute-risk interval or subtracted from 100% to estimate the proportion unaffected. The numerical correspondence suggests a transmission error; it does not establish the authors’ calculation process.

Antun [31] reports first threshold crossings in 26 of 398 participants eligible for the >52% analysis and 91 of 311 eligible for the >48% analysis: approximately 6·5% and 29·3%, respectively. These are observed proportions among threshold-specific eligible participants during follow-up, not standardised fixed-horizon risks. The full cohort of 424 is not the appropriate denominator for these first-crossing proportions.

A large relative measure can coexist with a substantially smaller absolute frequency, but neither its magnitude nor its confidence limits determine absolute risk without additional information. The error in [36] does not establish that erythrocytosis is harmless or clinically alarming. Interpretation requires the endpoint definition, eligible population and observation period.

9.2.3 How far up the range anyone actually gets

⚠⚠ In 192 prospectively followed trans men, four people (2·1%) ever reached at least 52%; one reached exactly 54·0%, and no value above 54·0% was reported in that cohort.[22] ⚠ Higher values are reported elsewhere: 12 of 511 reached 54% or above in one retrospective cohort, and a Chilean 24-month cohort records a maximum of 56%.[30][33] In the retrospective cohort of 511 patients, twelve (2·3%) had a recorded haematocrit of at least 54% — ⚠ the mean duration reported for the subgroup with erythrocytosis is not the whole cohort’s — a group whose mean age was 45·8 and whose mean time on testosterone was 84 months.[30]

⚠⚠ AND A TEN-YEAR COHORT REPORTS THE OPPOSITE SHAPE. In 57 trans-masculine adults followed for ten years, haematocrit rose from 40·0% to 45·8% and then stabilised; the number with values above 50% was between 0 and 8·8% at every annual time point and did not vary significantly over time. No major cardiovascular or thromboembolic event occurred in ten years, and one participant developed atrial fibrillation.[37]

⚠⚠ ITS AUTHORS ATTRIBUTE THAT LOW PREVALENCE TO THE MONITORING ITSELF, AND THIS REVIEW CARRIES THE CLAIM AT FULL WEIGHT BECAUSE IT RUNS AGAINST ITS OWN DIRECTION. They write that participants were closely monitored during the early phase of therapy and that this allowed clinicians to adjust regimens in time — modifying transdermal dose or the interval of injectable esters — and that the resulting prevalence of supraphysiological haematocrit appears lower than previous reports.[37]

⚠ It is uncontrolled, single-centre, and the attribution is the authors’ own interpretation of their own service. ⚠⚠ It is an uncontrolled service-level observation consistent with monitoring and regimen adjustment limiting sustained high haematocrit. It does not estimate the causal effect of monitoring, supplies no monitored-versus-unmonitored comparator, and does not establish that monitoring reduced clinical events. ⚠ It is worth carrying for that reason and no more.

⚠ Selection is heavy: 143 eligible, 72 lost to follow-up, 57 consenting.[37]

Detection accrues over decades of observation. In the Amsterdam cohort the cumulative probability of a first detected value above 0·50 L/L was 8% at one year, 38% at ten and 50% at fourteen; for >0·52 L/L, 4% at one year and 16% at ten, restricted to the 776 participants with baseline measurements.[28]

The cumulative probability of first detection rises with duration of observation even if the underlying hazard does not. Longer follow-up means more measurement opportunities, and it also means ageing, weight change, altered smoking and accumulating comorbidity. What it does establish, and it is the more useful point clinically, is that someone stable for a decade can still cross a threshold for the first time.

(The abstract of that paper gives 10% at one year; the Results give 8%. The Results figure is used.)

9.2.4 What travels with a raised haematocrit, among people who are all taking testosterone

In the Amsterdam cohort, after mutual adjustment: tobacco use (OR 2·0, 1·3–3·0), predisposing medical history such as chronic lung disease or sleep apnoea (OR 2·0, 1·1–3·8), and body mass index 25–30 (OR 3·0, 1·8–5·1) or ≥30 (OR 3·1, 1·7–5·6). Measured testosterone of 20–30 nmol/L was not distinguishable from 10–20; only above 30 nmol/L reached significance (OR 1·5, 1·1–1·8).[28]

A second cohort agrees on univariate analysis and then does not survive adjustment. Age above 30, body mass index above 30, sleep apnoea and prior oophorectomy all showed significant odds ratios; ⚠⚠ on multivariate analysis only age, body mass index and dose remained, and the oophorectomy association fell to p = 0·064.[30] This review reports the oophorectomy finding as it stands in that paper: proposed, univariate, and not surviving adjustment.

None of this can be read as showing that testosterone is a lesser cause. Every participant was receiving testosterone and there was no untreated comparator, so no such study can partition risk between the drug and its co-factors. What the analyses support is narrower and still useful: among people on testosterone, smoking, weight, age and predisposing lung disease travel with a raised haematocrit.

The Amsterdam authors’ own first-line recommendations follow: smoking cessation, weight reduction where BMI exceeds 25, treatment of chronic lung disease or sleep apnoea, and consideration of a switch to transdermal administration.[28]

⚠ And the trans-masculine trials exclude those very co-factors. One randomised trial excluded smokers, sleep apnoea and coronary disease at entry.[25] ⚠⚠ AND A SEPARATE OBSERVATIONAL COHORT excluded 35 of 64 recruits for smoking initially, with two further exclusions after smoking began, so its findings have limited applicability to people who smoke.[24] Its authors cite background literature reporting that transgender people are 2·7 times more likely to smoke — A SECONDARY CITATION, NOT AN ESTIMATE GENERATED BY THAT COHORT. ⚠ Eligibility restrictions are described study by study and are NOT attributed collectively to the randomised evidence: one of these two studies is not randomised, and an observational cohort’s exclusions cannot establish the selection properties of randomised evidence as a class (correction 558).

Confidence blocks follow §3’s scheme and are given at §10.

Grading block — §9.2, three statements; the outcome divides

1 — Testosterone raises haemoglobin and haematocrit in trans-masculine people

Association — ROBUST. A meta-analysis and three primary datasets, two of them with overlapping confidence intervals, one randomised.
Causal attribution — TWO MARKS, STATED SEPARATELY RATHER THAN ONE MARK WITH A CAVEAT.
— Testosterone as the cause of the erythrocytosis, across the combined human evidence — ROBUST.
— Direct causal identification within trans-masculine populations specifically — MODERATE.

⚠⚠ Direct trans-masculine longitudinal evidence is strongly consistent with a treatment effect, and the placebo-controlled randomised evidence establishing that testosterone causes the rise comes from trials in men outside gender-affirming hormone care. ⚠ The randomised trial in trans men compared two active testosterone formulations against each other; both arms received testosterone, so it supplies no no-testosterone contrast and does not by itself make the within-arm rise causal. An earlier version of this block cited it as randomised causal evidence in this population, and that was wrong.
⚠ A previous version carried a single ROBUST with the direct-population figure in a following sentence. Splitting the axis prevents indirect randomisation from doing the work of directness in a scheme that grades directness separately.
Observed effect size — MODERATE. +4·39 percentage points (95% CI 3·52–5·26) pooled; +1·48 g/dL haemoglobin. Heterogeneity low.
Patient-outcome importance attributable to the rise — NOT ESTIMABLE. ⚠ The laboratory change and the threshold crossings are measurable; what is not estimable is the patient-important consequence attributable to that rise. No direct trans-masculine evidence identified here estimates it.

⚠ One small prospective study identified via external review adds direct trans-masculine evidence. Lethin et al. (2024; PMID 38085088; [83]) followed 10 treatment-naïve trans men on intramuscular testosterone undecanoate for one year; mean haematocrit rose from 42% to 46% (P = 0.012), with 2 of 10 reaching ≥50% by study end. Its n is small and its provenance is secondary-citation in this review (identified through external review, not retrieved in primary); it is carried for what it is: a directly eligible prospective study whose direction is consistent with the meta-analysis estimate.

Population directness — direct. ⚠⚠ This is the review’s strongest directly measured trans-masculine effect. ⚠ Its causal grade reads ROBUST only when the direct longitudinal evidence here is combined with placebo-controlled evidence from men outside gender-affirming hormone care; the experimental causal identification is not itself direct.

2 — How often a threshold is crossed

Observed effect size — NOT ESTIMABLE for a single pooled figure as currently specified. The individual studies report estimable frequencies or risks, but this review has not defined a common target quantity and harmonised thresholds, observation periods, ascertainment and analysis sufficiently to support a pooled estimate. Differences in estimator alone do not establish differences in the underlying target quantity.
Population directness — direct.

⚠⚠ AND A SIXTH COHORT REPORTS A PROPORTION FAR ABOVE THE OTHERS. A Chilean unit followed 66 trans-masculine people for 24 months, all on testosterone undecanoate 1,000 mg intramuscularly every three months — a schedule no other cohort here uses. Haematocrit rose from 41·4% to 47·2%, and at least 26 of 66, or 39%, exceeded 50%, with a maximum of 56%. Twenty-three resolved spontaneously; three had phlebotomy under a local rule of >52% on two consecutive measurements; and no thrombotic events occurred.[33]

⚠ This review does not pool that 39% with the figures above, and does not treat it as a correction to them. Four features distinguish it: undecanoate 1,000 mg quarterly in every participant — a standard long-acting regimen, within the 1,000 mg every ten-to-fourteen weeks used at Amsterdam[28] and the 500–1,000 mg every six-to-fifteen weeks recorded at §5, but a recognised long-acting regimen — and no corpus-wide uniqueness is claimed for it: other cohorts here also use a single formulation, and weekly-milligram equivalence across different esters is not a biological dose comparison (correction 537) (correction 506), a mean baseline body mass index of 28, a fixed 24-month window, and severe selection — 398 people had received therapy at the unit and 274 were excluded for insufficient follow-up or adherence, leaving 124 analysed across both groups.[33]

⚠⚠ IT ALSO CARRIES A FINDING ABOUT THIS REVIEW’S OWN SEARCH. It is published in Spanish, in a Chilean journal, and its authors observe that nearly all work in this area has been done in Anglophone populations. A retrieval process operating in English by default would not surface it, and did not.

⚠⚠ AND A FIFTH ESTIMAND, WHICH IS THE ONE A CLINICIAN ACTUALLY WANTS. A US descriptive fixed cohort study using chart data, 282 people, all but three on testosterone cypionate at a median 100 mg weekly, computed cumulative incidence with survival analysis and censoring for loss to follow-up: 12·6% above 50·4%, 1·0% above 52%, and 0·6% above 54% at twenty months.[32] ⚠ Its authors conclude that routine three-monthly screening in the first year may be unnecessary for all patients — a direct challenge to the monitoring schedule at §5.

⚠ Three cautions travel with it. Loss to follow-up was high: 46% remained at twelve months and 25% at twenty. Haematocrit was imputed from haemoglobin for 14·5% of data points, by multiplying by a cohort-derived factor of 3·023. ⚠⚠ And its threshold, 50·4%, is the upper limit of its laboratory’s reference interval for men — object one of the taxonomy at §9.3 being used directly as object two.

⚠ What can be stated: in the one prospective cohort, over an investigation period running to 36 months with unequal site follow-up, 11·5% ever reached 50%, 2·1% ever reached 52%, and one participant reached 54·0% — the highest value measured in 192 people.

3 — What travels with a raised haematocrit among people all taking testosterone

Association — LIMITED. ⚠ Thin and confounded rather than divided. Smoking, body mass index, age and predisposing lung disease recur across cohorts; one cohort’s oophorectomy signal did not survive adjustment.
Causal attribution — NOT ESTABLISHED. No cohort has an untreated comparator, so none can partition risk between the drug and its co-factors.
Population directness — direct.

9.3 The measurement everyone makes, the one nobody makes, and the one nobody states

9.3.1 Haematocrit did not closely quantify measured red-cell mass in the one subset identified here

Every cohort reviewed here reports haematocrit. A raised value may reflect absolute erythrocytosis, a true increase in red-cell mass, or relative erythrocytosis, a contracted plasma volume. The two carry different implications.

⚠⚠ AND THE DISTINCTION HAS TO REACH THE CLINIC, OR IT IS ONLY A TAXONOMY. A raised haematocrit measured after testosterone is not by itself evidence of an increased red-cell mass, and it is not by itself evidence that testosterone is the cause. ⚠ Persistence, hydration and plasma-volume state, the regimen and the timing of sampling relative to a dose, smoking and other carbon-monoxide exposure, sleep-disordered breathing, hypoxic disease and altitude all bear on the reading, and an unexplained or disproportionate elevation can have a primary or another secondary cause entirely. ⚠⚠ THIS REVIEW ASSERTS NO DIAGNOSTIC ALGORITHM AND NO TESTING RECOMMENDATION. It records that a concentration measurement does not establish absolute erythrocytosis or its cause, which is a statement about what the measurement can carry and not about what should be done (correction 527).

⚠⚠ In the one cohort that measured both, the coefficient of determination was R² = 0·08 — the source describes this as the correlation, though R² is the proportion of variance explained and not a correlation coefficient.[35] Red-cell mass was measured by nuclear dilution assay in 44 of the 206 patients with secondary and idiopathic polycythaemia — 21%, which the source reports as 20% — and plotted against haematocrit at diagnosis (correction 504). ⚠ In that measured subset — roughly forty-four patients — haematocrit accounted for eight per cent of the variance in red-cell mass.** ⚠⚠ That is a property of the measured subset of a heterogeneous referral cohort, not a general property of haematocrit, and whether the same relationship holds in testosterone-treated trans-masculine people is unknown.

Two further sources say the same in their own words. A randomised trial of haematocrit targets states that haematocrit cannot be used as an accurate surrogate for the red-cell mass, since increases in plasma volume could mask the degree of increase in red-cell mass.[38] And a prospective cohort in this population notes that absolute erythrocytosis can only be suspected when haematocrit is raised, because that does not discriminate from apparent erythrocytosis, which is normal red-cell mass with reduced plasma volume.[22]

Every prevalence figure in §9.2 is a threshold-crossing proportion, and what it counts is not the same as absolute erythrocytosis. ⚠ How closely the two correspond in trans-masculine people has not been measured, and the figure above comes from a different population and a selected subset of it.

⚠⚠ AN EARLIER VERSION OF THIS SECTION BUILT A CAUSAL CHAIN OUT OF TWO CORRELATIONS AND IT IS WITHDRAWN. It set a cross-sectional R² of 0·029 for serum testosterone against haematocrit beside an R² of 0·08 for haematocrit against red-cell mass, and concluded that each link the apparatus assumes is loose. ⚠⚠ That is a category error. A single contemporaneous testosterone concentration measured between people already receiving treatment is a poor predictor of their haematocrit; that says nothing about the within-person causal effect of starting testosterone, which this review grades ROBUST from randomised and longitudinal evidence, that grade belonging to the combined experimental evidence while causal identification specifically within the direct trans-masculine evidence is graded MODERATE at §10 (correction 665).

⚠ What the first figure does establish is narrower and still useful: serum testosterone is a poor predictor of haematocrit between individuals at steady state.[21] A second source agrees — haemoglobin was unrelated to serum testosterone within a cohort of 79 trans men on stable therapy, p = 0·42.[19] Together they establish that a single contemporaneous serum testosterone value is a weak between-person predictor of haematocrit at steady state. ⚠⚠ They do not establish how effectively changing dose or concentration changes haematocrit within an individual — a different question, on which §9.5’s management evidence bears directly and these correlations do not. ⚠⚠ They do not constrain the causal effect of treatment itself.

A third mechanism is offered by an independent systematic review: red cells and plasma are not uniformly distributed in vivo, and haematocrit is higher in large vessels than in small ones, so estimating red-cell mass from a peripheral sample may be inaccurate on those grounds too.[34]

9.3.2 Six different numbers, and the review has been treating them as one family

⚠⚠ A HAEMATOCRIT VALUE CAN BE SIX DIFFERENT KINDS OF OBJECT, AND THIS REVIEW HAS NOT SAID SO. ⚠ The sixth — the investigation threshold — was added at v4.28 from a haematology guideline and is not this review’s construction.

⚠⚠ OBJECT THREE — THE INVESTIGATION THRESHOLD — WAS ADDED AT v4.28, AND IT COMES FROM A HAEMATOLOGY GUIDELINE RATHER THAN FROM THIS REVIEW. The British Society for Haematology directs that a persistently raised venous haematocrit — above 0·52 in males, above 0·48 in females — should be INVESTIGATED, and separately that patients with polycythaemia vera should be treated to a target below 0·45.[16] ⚠ Neither number is an initiation caution, an on-treatment action level, or an operational definition of disease. The first says find out why; the second says treat to here.

⚠⚠ AND IT IS AN INVESTIGATION THRESHOLD, NOT A DIAGNOSTIC ONE. THE DISTINCTION IS THIS SUBSECTION’S WHOLE PURPOSE AND AN EARLIER VERSION COLLAPSED IT ONE PARAGRAPH AFTER DRAWING IT. A value above which you find out why a haematocrit is raised is not a value that diagnoses a disease. The same guideline carries separate diagnostic criteria for polycythaemia vera, and those are a different proposition not made here. Correction 152.

⚠ The taxonomy now carries SIX numerical objects. ⚠⚠ An earlier version numbered this seventh, which conflicted with the subsection’s own architecture: five numerical objects plus MONITORING as a sixth item that is deliberately NOT a haematocrit number. Monitoring remains outside the numbering. Correction 151.
>
> ⚠⚠ AND THE TWO CARRY DIFFERENT SEX RULES, WHICH IS THE SHARPEST DEMONSTRATION THIS REVIEW HAS THAT THRESHOLD OBJECTS ARE NOT INTERCHANGEABLE. The investigation threshold is sex-specific. The treatment target is explicitly NOT: no study has explicitly defined a gender difference in Hct target. A different target Hct in males and females is not recommended.[16] One society, one document, two haematocrit numbers, opposite sex rules, and a stated reason for each.

⚠ These sex-specific BSH investigation cut-offs were not written for gender-affirming care and do not state how they should be applied to a trans-masculine person receiving testosterone. They are carried here to demonstrate the distinction between an investigation threshold and a treatment target, not as a trans-specific investigation rule.
>
> ⚠ This review did not construct that distinction. A haematology society draws it, and the taxonomy is extended to accommodate it rather than the other way round.
>
> ⚠⚠ AND THAT GUIDELINE’S OWN CAUSE LIST NAMES TESTOSTERONE EXPLICITLY. Its table of causes places use of androgen preparations under drug-associated secondary erythrocytosis, and its diagnostic algorithm lists drug induced — testosterone among the clinical-history items identifying a likely secondary cause.[16] ⚠ In a patient taking testosterone, the testosterone exposure is therefore, in that guideline’s own framework, a recognised potential cause of secondary erythrocytosis. Where the clinical and laboratory picture establishes a clear secondary cause, some patients may require no further aetiological investigation. Testosterone exposure alone does not exclude dual pathology or a concomitant myeloproliferative disorder; further investigation remains appropriate where the phenotype warrants it.

⚠⚠ DUAL PATHOLOGY — MANDATORY CLINICAL NOTE. A plausible testosterone contribution does not exclude a concomitant myeloproliferative disorder or another primary cause. Discordant clinical features — persistent marked elevation, unexpected trajectory, leukocytosis, thrombocytosis, splenomegaly, or a low EPO level — may justify directed haematological assessment despite testosterone exposure. The BSH guideline explicitly names dual pathology as a consideration that clinicians should maintain. That is the clearest statement this review has found of why the two literatures’ thresholds answer different questions.

⚠ Monitoring sits alongside these six numerical objects but outside the numbering: the separate decision to monitor at all. It is made in trans-specific guidance on trans-specific grounds, and it is the part of the apparatus this review does not question.

⚠⚠ OBJECTS THREE AND FOUR ARE NOW DEMONSTRATED RATHER THAN PROPOSED, AND THE DEMONSTRATION IS THE SOCIETIES’ OWN. Every one of the eight guidelines read in primary at §5 sets an initiation position and an on-treatment action level as SEPARATE objects, and in seven of the eight the two carry different values. ⚠⚠ The initiation positions vary — unquantified, 48–50%, 50%, 52%, 54% — while the on-treatment level is 54% in seven of the eight.[7][8][9][11][12][13][15]

⚠ This review did not discover the distinction. The societies draw it, and the widely cited tabulation of them cannot represent it, having one action column for two objects. ⚠⚠ The taxonomy’s value is therefore not that it corrects the guidelines. It is that it names what a secondary reading of them loses.

⚠⚠ AND THE DISTINCTION BETWEEN OBJECTS THREE AND FOUR PREDATES THIS REVIEW. The Endocrine Society’s 2006 guideline for androgen-deficient men states both separately: erythrocytosis above 50% as a condition in which testosterone should not be administered, and, under monitoring, if haematocrit is greater than 54%, stop therapy until haematocrit decreases to a safe level, evaluate the patient for hypoxia and sleep apnoea, and reinitiate therapy at a reduced dose.[2] ⚠ The later trans-specific documents do not reproduce that two-object structure.[3][1] §6 sets out the documentary relationship and declines to call it a transmission.

⚠⚠ THE CONSEQUENCE IS SHARPER THAN THE DISTINCTION. 50% — the commonest operational definition of erythrocytosis in this literature, used in nine of the twelve prevalence figures at that threshold — lies within the study-specific reference interval derived in one selected healthy cohort of 79 transgender people receiving stable testosterone (Greene et al., Clin Chim Acta 2019).[19][34] ⚠⚠ A value lying within one empirically derived reference interval is simultaneously used by several studies as an operational threshold for erythrocytosis. ⚠ That does not make the interval wrong or the definition invalid. It shows that statistically typical for this population and crosses this study’s counting boundary are different propositions — which is the whole point of separating the six numerical objects above. 52% and 54% sit above the interval; 48% sits well below it.

And the harm was named by a laboratory-medicine group in 2019. Providers, they write, may become concerned for polycythaemia in patients whose haemoglobin or haematocrit exceeds the standardized female reference range, which may lead to a provider decreasing or discontinuing testosterone therapy… because of an unwarranted fear of side effects.[19]

⚠ That study derived its interval in a healthy cohort — current smokers, sleep apnoea, obesity, diabetes and prior cardiovascular events all excluded. A reference interval should be derived that way. But the thresholds are not applied only to that population.

9.3.3 Two conventional definitions, the same people, and a two-and-a-half-fold difference

One records study applied both standard criteria to the same 519 patients. Haemoglobin above 17·5 g/dL identified 7·8%. Haematocrit above 50% identified 20%.[29]

⚠⚠ These two conventional but non-equivalent operational criteria classify markedly different proportions of the same cohort: 7·8% by haemoglobin and 20% by haematocrit. Reported erythrocytosis prevalence is therefore highly definition-sensitive.

9.3.4 And the method that produced the number is almost never stated

⚠⚠ Of the thirteen sources audited under the Source Provenance Standard at the time that audit was run, ten do not state how the haematocrit was measured. ⚠ That audited set is named in the provenance pass and has not been re-run since the evidence base expanded; the ratio is not a claim about every source now in this review. Three do. This is not a gap in this review’s reading; in that audited set, the missing method information is a property of the source reporting rather than a gap in this review’s extraction.

Where the method is stated, it differs, and the difference is not negligible. One arterial cohort used the spun microhaematocrit, cited to the packed-cell-volume standard.[39] The other used a flow-cytometer analyser, with a coefficient of variation of 2·9%.[40] The prospective trans-masculine cohort used automated analysers at both sites, with coefficients of variation of 1·4% and 1·1%.[22]

They cannot be assumed interchangeable without method- and laboratory-specific verification, and the difference has been measured in one such comparison. In 150 duplicate samples from 84 polycythaemic patients, centrifuged microhaematocrit read 0·467 and the analyser-calculated value 0·448 — a difference of 1·9 percentage points (p = 6·8 × 10⁻²⁵), attributed to plasma trapped between packed cells.[41]

⚠ The difference is not a constant, and the figures are CENTRE-SPECIFIC MEAN DIFFERENCES. Across four centres those means ran from 0·5 to 3·8 percentage points, and the difference was not statistically significant at one of them; they are neither individual limits of agreement nor a conversion that can be applied to a patient (corrections 668, 671). The authors accordingly recommend running both methods in parallel locally rather than applying a fixed correction, and reported in 1999 that the international standardisation body had not then recommended correcting for trapped plasma; current standards have not been checked here.[41]

⚠ THE SAME PROBLEM HAS BEEN RECOGNISED ELSEWHERE. Treatment thresholds in polycythaemia vera were established using the microhaematocrit method, while current monitoring is mostly by automated counter. Those authors proposed a target below 0·43 on an analyser in place of the 0·45 established on spun blood, expressly conditional on local method bias matching their own, and recommended running both methods in parallel to establish it.[41] ⚠⚠ This review does not claim that any field adopted that correction. An earlier version did, and it was wrong: eight authors proposed it. ⚠ An earlier version read four, which is the number of hospitals in the study, not of authors — correction 217).

⚠ None of the guideline passages directly examined in this review specifies a haematocrit measurement method in the relevant recommendation. None of the eight historically verified Cannarella guideline rows specifies a haematocrit measurement method in the relevant recommendation. Current EAU-SRH-2026, read separately as a distinct current-guidance source, likewise does not specify measurement method in its threshold recommendations. ⛔ The historical Cannarella EAU row remains unverified for this feature. EAU-SRH-2026 was read separately but does not specify measurement method. ⚠ The separately-held EAU 2019 document DOES carry a pre-analytical instruction — minimal or no venous occlusion when sampling — but it is not one of the nine tabulated rows and does not alter this count. ⚠ An earlier version said the remaining primaries had not been read at all; see the provenance block at §5.

⚠⚠ This review states the consequence and no more: comparability between these sources is unestablished. ⚠ It does not claim that older values are inflated or that current thresholds are permissive by any particular amount. The direction is known from the method literature; the magnitude cannot be applied here, because ten of thirteen sources do not say which method they used.

9.3.5 And nobody measures duration

None of the cohorts reviewed here quantified cumulative time above threshold, though one set of authors suggests duration may matter more than the peak value reached.[29] ⚠ That is untested, and it names an exposure metric not quantified in the cohorts reviewed here.

Grading block — §9.3, two statements, and the first is graded against

1 — Haematocrit accurately quantifies absolute red-cell mass in secondary and idiopathic polycythaemia

Association — LIMITED. ⚠ The mark rests on the two sources that state in their own words that haematocrit cannot serve as an accurate surrogate. In the one cohort reviewed here that measured both, R² = 0·08 among the 20% who underwent nuclear measurement — a cohort assembled for raised haematocrit and further selected for testing, in which an attenuated R² is CONSISTENT WITH poor correspondence and cannot quantify it. An earlier form of this cell read evidence AGAINST close quantitative correspondence, which is a weak statistic read as a finding in the opposite direction (correction 666, Eden’s ruling). ⚠⚠ Corresponding performance in testosterone-treated trans-masculine people is unmeasured. Two further sources state in their own words that haematocrit cannot serve as an accurate surrogate.
Observed effect size — LIMITED. Eight per cent of variance explained, among roughly forty-four patients — the 20% of a 206-patient referral cohort who underwent nuclear measurement, doubly selected and range-restricted, so the figure bounds nothing (correction 666).
Population directness — indirect. Secondary and idiopathic polycythaemia, not trans-masculine.

⚠⚠ THE STRENGTH OF THIS STATEMENT RESTS ON THE TWO SOURCES THAT STATE IT IN THEIR OWN WORDS, NOT ON THE R². [35]’s cohort was assembled for investigation of a RAISED haematocrit — medians 0·51 and 0·54, interquartile range 0·52 to 0·57 — and red-cell mass was measured only in the subset whose clinicians ordered it. Selection for a raised haematocrit, and further clinical selection for red-cell-mass testing, limit what the observed relationship generalises to. Restriction of predictor range CAN reduce R² under common linear-model assumptions, but not invariably — a noiseless linear relationship retains R² = 1 under any restriction leaving variation in the predictor — and neither the direction nor the magnitude of the selection effect is established here. So 0·08 describes this selected sample and does not establish correspondence across the full haematocrit range (correction 533). ⚠ An earlier form of this passage asserted attenuation whatever the underlying relationship, which is mathematically false (correction 509, superseded). ⚠ Haematocrit is a red-cell volume fraction and does not uniquely determine absolute red-cell mass without information about blood or plasma volume. The randomised trial and the prospective cohort state that physiological limitation in their own words, and the attribution is faithful. ⚠⚠ THEIR STATEMENTS ARE NOT RANDOMISED VALIDATION EVIDENCE FOR IT. A trial that randomised haematocrit TARGETS did not thereby test whether haematocrit tracks red-cell mass, and a statement’s evidential class comes from the analysis behind it, not from the design of the paper it appears in (correction 534). ⚠ The selected-sample association is reported separately and carries its own limits.

2 — Measurements across this evidence base are commensurable

NOT ESTABLISHED. ⚠⚠ In the thirteen-source audit performed at an earlier stage of this review, ten of thirteen sources do not say which method they used; that audit has not been repeated against the expanded corpus. Where stated, the methods differ and the difference is quantified elsewhere at about 1·9 percentage points, varying 0·5 to 3·8 by site.
Population directness — not applicable.

9.4 The harm attributable to testosterone-associated haematocrit has not been established — and that is a different claim from safety

R61.3 executed (correction 330): §9.4 pruning ruling applied.

9.4.1 Arterial outcomes — the association survives two cohorts, and both measure it near 50% rather than near 54%

The two cohorts, and what each measured

§9.3 established what the thresholds are for and what they are not for. The question now is what happens to people, and the arterial evidence is where that question has actually been asked.

Two prospective cohorts have measured raised haematocrit against hard arterial endpoints in general populations. Toss and colleagues followed 417,099 Swedish male conscripts, haematocrit measured at 18–19 during national conscription between September 1969 and December 1977, for a median of 36 years [39]. Warny and colleagues followed 108,521 participants in the Copenhagen General Population Study, haematocrit measured at study entry, for a median of 8 years [40]. ⚠ Neither is a trans-masculine cohort; gender identity was not reported. Toss is entirely male; Warny is 55% women, mean age 58, and explicitly restricted to white individuals of Danish descent, a limitation its authors state. ⚠ Neither identifies or analyses a trans-masculine subgroup, and nothing below transfers to trans-masculine populations without the argument made in §9.5.

In Toss, 9322 first myocardial infarctions occurred, at a median age of 51. Men in the top band, ≥49%, had a fully adjusted relative risk of 1.38 (1.29–1.48) against a reference of ≤44%, with 2.8% of the top band and 1.8% of the reference band suffering an infarction. The attenuation across the adjustment sequence should be seen rather than asserted: unadjusted 1.58 (1.48–1.69), falling to 1.32 (1.23–1.41) once age, conscription year and place, education, income, body mass index, baseline diagnoses and systolic blood pressure were entered — and then rising to 1.38 when erythrocyte sedimentation rate was added, because ESR is inversely correlated with haematocrit in this cohort (r = −0.31) and had been masking part of the association. On a continuous scale, one standard deviation of haematocrit carried 1.18 (1.16–1.21) unadjusted and 1.13 (1.10–1.15) fully adjusted.

In Warny, the top sex-partitioned band — above 45% in women, above 48% in men, the 95th percentile in each — carried a hazard ratio for myocardial infarction of 1.46 (1.06–2.00) against the 25th–75th percentiles, with incidence rates of 43 against 25 per 10,000 person-years. Accounting for the competing risk of death, the sub-hazard ratio was 1.38 (1.00–1.90), p = .05. Results held after excluding the 320 participants with myeloproliferative neoplasia, after excluding the 22,859 with any Charlson comorbidity, and after further adjustment for C-reactive protein and for systolic blood pressure.

Two cohorts, two countries, two eras, two age structures, one direction, comparable magnitude. This is the most consistent general-population arterial association examined in this tranche, and the review should say so before it says anything else. ⚠ An earlier version called it the most robust finding in the erythrocytosis literature. That is too global for two observational cohorts, in a manuscript that also carries randomised evidence for testosterone raising haematocrit, and the superlative is withdrawn.

The strongest objection is one neither cohort can answer — a trait that predicts is not an exposure that causes

Toss measured haematocrit once, at 18–19, and never again. Its authors say so, note that intra-individual variability over time is considerable, and argue that regression dilution therefore probably underestimates the true association. That argument may well be right. It does not touch the objection that matters.

What Toss establishes is that a haematocrit measured at age 18–19 in adolescent males predicts myocardial infarction three decades later. ⚠ This is a single measurement at one point in adolescence, not an established constitutional trait. What it does not establish — what it was not designed to establish and does not claim — is that raising a given person’s haematocrit from 44% to 50% at the age of thirty alters that person’s risk. Those are different propositions. The first is about a baseline prognostic marker that may travel with other long-term determinants; the second is about a within-person pharmacological intervention. Testosterone poses the second question, and neither of these cohorts answers it.

This is not a technicality retrieved from a limitations section. It is the load-bearing joint of the entire argument that follows, and it is why §12 does not conclude what a reader arriving from the guidelines might expect it to conclude.

Warny partly escapes the measurement half of this objection and not the substance of it. Its main analyses were adjusted for regression dilution using a ratio of 0.63, derived from 5511 participants with repeat measurement after a median of 10.5 years — so its estimates are not raw single-measurement estimates. But its authors state plainly that the study is observational and cannot address causality, and they go further than that.

Warny’s authors cap their own finding, and the review should adopt their cap rather than their headline

Comparing unadjusted against adjusted estimates, Warny found that confounding barely mattered for platelet count and mattered a great deal for haematocrit. In the dichotomous whole-cohort analysis, the myocardial infarction estimate fell from 1.74 (1.27–2.40) unadjusted to 1.42 (1.05–1.92) adjusted; the stroke estimate from 1.50 (1.08–2.09) to 1.29 (0.92–1.77); the venous estimate from 1.46 (1.06–2.00) to 1.21 (0.88–1.62). The unadjusted venous figure coincides exactly — point estimate and both bounds — with the banded myocardial infarction estimate given above. The coincidence is Warny’s own, verified in both places with the source open; it is not a carried interval, and a review of this block suspected that it was. The authors’ inference from that pattern is that lifestyle and other factors influence haematocrit more than they influence platelet count, that further unadjusted confounding is therefore likely to be present, and that the risk estimates for high haematocrit should be considered a maximum estimate.

That sentence is the most useful thing either cohort says about its own arterial finding, and it is written by the people who produced it.

The two cohorts also disagree, and the disagreement is worth surfacing rather than smoothing. Toss concludes that haematocrit may aid future risk assessment in young individuals. Warny, having found a 1.5-fold risk, concludes that it seems unlikely adding haematocrit would improve risk prediction for this disease. A cohort of adolescents and a cohort with a mean age of 58 reached opposite views about the same measurement’s clinical utility. Neither is obviously wrong; the populations differ in exactly the way that would produce the difference. But no summarising document in this literature reports that the two exist and conflict.

Both cohorts measured the association between roughly 46% and 51% — and the threshold in use sits above both

This is where the arterial evidence stops, and the stopping point is precise.

Toss resolves the exposure in five groups: ≤44% open below, 45%, 46–47%, 48%, and ≥49% open above, with measurements retained between the protocol exclusion limits of 30% and 60%. Read row by row against the reference, the fully adjusted ladder is not a step-by-step ascent. At 45% the estimate is 1.05 (0.97–1.13) — on 1218 infarctions in 62,400 men, 2.0% against the reference band’s 1.8%. At 46–47% it reaches 1.18 (1.11–1.25) on 2785 in 124,198, 2.2%; at 48% it is 1.17 (1.09–1.26) on 1439 in 62,683, 2.3%, which does not exceed the band below it. Only the top group separates, at 1.38 (1.29–1.48). The gradient is carried by a significant test for trend — p < 0.001 in all six adjustment models — and not by each step exceeding the one beneath it. Above 48% there is one band. It is centred at a mean of 50% with a standard deviation of 1%, and it carries 1.38. ⚠⚠ The study’s exclusion limit was 60%, which is a protocol boundary and not an observed maximum; the paper reports no maximum, so 54% is roughly four standard deviations above the band’s own mean and the number of participants at or near it is unknown. Nothing in the paper distinguishes 49% from 52% from 55% within it.

Warny’s top band has a median of 49% with an interquartile range of 46 to 50.

Two cohorts, 525,620 people between them, and the arterial association is measured in individuals who sit at approximately 46–51% haematocrit. Both uppermost bands are open above a floor at or below 49% — Toss’s is 49% and above, with data retained to 60% by its own exclusion limits, and Warny’s is above 45% in women and above 48% in men with no reported maximum. 54% is therefore outside the resolution of either cohort, and whether anyone was measured there is not established by either — Warny reports no maximum, so its own range is not established in either direction. It is not contradicted by this evidence. It is not supported by it either. No reported estimate is specific to 54%.

The distinction matters because §9.3 established that arbitrariness is not the defect. A threshold not separately resolved by the evidence adduced for it is not an arbitrary threshold — it is an extrapolated one, and extrapolation is a claim that can be examined. This review examines where it came from in §9.6, and what it can support in §12.

Grading.

• Association between raised haematocrit and myocardial infarction in general-population cohorts not designed as transgender or gender-affirming-hormone cohorts: MODERATE. Two large prospective cohorts, consistent direction and magnitude, graded association in one (p < 0.001 for trend across all six adjustment models), validated endpoints in both, no declared conflicts in either. Downgraded from ROBUST because both are observational, both attenuate substantially on adjustment, and one caps itself as a maximum estimate.
• Effect of raising haematocrit on myocardial infarction risk: NOT ESTABLISHED. No interventional evidence exists in either cohort; the design cannot produce it.
• Any threshold effect at or near 54%: NOT ESTIMABLE. No cohort reports an estimate specific to that value. The uppermost category is open above its floor and yields one pooled estimate for everyone in it; ⚠ whether any participant was measured at 54% is not reported by either cohort.

⚠ In Tromsø, each five-percentage-point rise in haematocrit was associated with VTE HR 1.25 (95% CI 1.08–1.44) across the whole cohort, adjusted for age, smoking and body mass index, and 1.37 (1.10–1.71) for unprovoked events; 447 first lifetime events accrued over a median of 12.5 years in 26,108 people followed for 282,494 person-years.[71] Among men the crude incidence was 1.47 per 1,000 person-years below 43% and 1.90 at 46% or above; these crude rates carry the age structure of each band and are not an attributable-risk estimate.

⚠ Tromsø’s categorical cut-points were sex-specific by construction. No statistically significant interaction between haematological variables and sex was detected. Differences in which sex-specific estimates cross a significance threshold therefore do not establish sex-specific effect modification. The analysis was performed in an untreated general population; neither cut-point applies to trans-masculine people receiving testosterone without bridging evidence that does not exist in this review.

R61.2 executed (correction 334): upper-bound 1.71 for Tromsø unprovoked now carried. R61.5 executed (correction 334): four integration areas resolved.

9.4.2 Venous outcomes — the evidence is divided, and the division is the finding

⚠ Among the large population cohorts identified and analysed here, four have asked whether haematocrit in the upper tail predicts venous thromboembolism in the general population. ⚠ None is a trans-masculine cohort; gender identity was not reported in any. The four cohorts yield heterogeneous estimates: point estimates are not consistently near the null, but precision and statistical support differ substantially and the exposure bands, reference categories and adjustment structures are not commensurable. No single design feature orders the estimates. Adjustment depth, the floor of the upper band and reference-category placement do not provide a monotonic explanation. The heterogeneity itself is the finding.

Three of the four are set out below. The fourth, LITE [44], is held back from the table and carried in the argument that follows, because its most informative property is not its hazard ratio but its band construction, and a table row conceals that.

None of the proposed design orderings explains the spread in estimates. Adjustment depth does not order either point estimates or precision: Tromsø reports 1.54 (1.08–2.21), LITE 1.03 (0.69–1.53), Copenhagen 1.26 (0.92–1.72), and the most heavily adjusted ARIC analysis 1.72 (1.30–2.27). ⚠⚠ ARIC’s 1.72 is its TOTAL-VTE estimate, and ARIC’s own Results confine the association to the provoked stratum: provoked 2.05 (1.47–2.86), unprovoked 1.21 (0.73–2.02). That matters because unprovoked associations are conventionally read as more likely causal, and ARIC’s is null. Neither does the floor of the upper band; point estimates do not rise monotonically as the cut-point increases. Reference-category placement is a plausible source of non-comparability, but these between-cohort contrasts do not isolate how much of the difference it explains. ARIC and LITE additionally differ in band construction, follow-up, event accrual and cohort composition.

Band construction is the live explanatory candidate, and the evidence for it is drawn from overlapping analyses that do not hold follow-up, event accrual or cohort composition constant

The most instructive comparison in this literature is not between cohorts but inside one. ARIC is the long-follow-up component of LITE, and its Model 1 adjusts for age, race and sex — the same three covariates as the earlier LITE analysis, on a cohort supplying 14,594 of LITE’s 19,293 participants and 117,736 of its 148,054 person-years. The correspondence is weaker on the numerator than on the denominator, and the block says so rather than leaving it to be found: 85 of LITE’s 215 events come from the Cardiovascular Health Study, a separate cohort recruited at 65 and over, contributing events at more than twice the rate of the ARIC arm. LITE returned a hazard ratio of 1.03 (0.69–1.53) over a median 7.8 years of follow-up. ARIC Model 1 returns 1.95 (1.49–2.56).

The two analyses differ in what the top band contains. ARIC’s authors state their reason for partitioning: men and white participants were over-represented at high haematocrit, so a pooled cut point would not isolate the upper tail of anybody’s distribution. They therefore used race-sex–specific centiles, which places the male top band’s floor at 49.4% and the black female top band’s floor at 44.0%. LITE did not partition. It applied a single pooled-sex cut at 40% and 43.5% to a cohort 55% female, so its upper band is not ARIC’s upper band but a pooling of ARIC’s fourth and fifth: the range in which ARIC finds 1.17 (0.98–1.40) together with the range in which it finds 1.72 (1.30–2.27). Pooling lower-exposure participants with an upper-tail subgroup could dilute an association confined to the extreme tail toward the null. ⚠ It need not produce one, and follow-up, event accrual and cohort composition differ between these analyses too.

LITE reports no sex-specific haematocrit distribution and no sex-stratified haematocrit result, and its authors state that statistical power was insufficient for detailed analysis by race or sex. It therefore cannot exhibit the distortion ARIC’s rationale describes, and cannot exclude it either. The difference between the two analyses is not that one demonstrates a defect in the other; it is that only one of them was built to isolate the upper tail of anybody’s distribution.

This pairing is an informative quasi-within-dataset comparison rather than a controlled test of band construction, because it holds the covariate set and the outcome definition fixed, and the cohort substantially so, while varying the exposure construction. This pairing is particularly informative for interpreting the lower LITE estimate, because substantially overlapping data yield a materially different estimate under a different exposure construction.

Nothing here establishes that band construction explains the heterogeneity across all four cohorts. It establishes that in the one place the literature has varied band construction across substantially overlapping data, though not holding follow-up, event accrual or cohort composition constant, the estimate moved from 1.03 (0.69–1.53) to 1.95 (1.49–2.56).

Reference-category placement is a rival candidate, and this review does not quantify how much of the contrast it could account for

Warny raises the rival himself, about this exact literature: several earlier studies used the lowest third or fourth of the distribution as their reference, and those individuals may not be an optimal comparison, since some will have diseases making them more or less likely to suffer thrombotic events. The concern is well founded, and both cohorts able to see the low tail do find it elevated.

Aside — an exploratory reweighting INSIDE ARIC, and nothing is carried out of it (this review’s arithmetic). Elevation at the low end is confined to the bottom 5th centile in both cohorts that can resolve it. In ARIC the crude rates run 2.9, 2.4, 2.4 per 1,000 person-years across the bottom three bands, so the 5th-to-25th region is exactly flat against the reference; Copenhagen’s low band is likewise elevated only at its floor. A reference group drawn from the bottom 40% would therefore contain one stratum elevated by about a fifth and two that are not. Weighting ARIC’s own person-years — 13,828 at 2.9, 61,787 at 2.4, and three-tenths of the reference band’s 155,542 at 2.4 — gives a bottom-40% crude rate of 2.46 against a reference of 2.4, an inflation of 2.3%; weighting by centile width instead gives 2.6%. ⚠⚠ THAT IS THE WHOLE OF WHAT THE CALCULATION ESTABLISHES, AND IT ESTABLISHES IT INSIDE ONE COHORT.

⚠⚠ AN EARLIER VERSION CARRIED THAT FIGURE ACROSS TO TROMSØ AND LITE AND REPORTED THAT THEIR ESTIMATES WOULD BECOME ABOUT 1.58 AND ABOUT 1.06, THAT A SPINE FACTOR OF 1.89 WOULD BECOME 1.84, AND THAT A GENEROUS 5% INFLATION WOULD TAKE IT ONLY TO ABOUT 1.80. EVERY ONE OF THOSE DERIVED VALUES IS WITHDRAWN. The transformation was not defensible on four independent grounds: it borrowed a low-tail distribution from a different cohort; it applied a crude incidence-rate adjustment to ADJUSTED HAZARD RATIOS, which are non-collapsible; the covariate sets and exposure-band compositions differ between the cohorts involved; and the individual-level reanalysis that would license the transformation does not exist. ⚠ Labelling the exercise exploratory, and stating that it was neither a bound nor a correction, did not repair that. A printed number reads as authoritative whatever the sentence above it says, and this review’s own §9.6 catalogues exactly that mechanism operating between papers.

⚠ What survives is narrower and is the limit of the exercise: reference-category placement is a real rival candidate, and this review does not quantify how much of the between-cohort contrast it could account for. ⚠⚠ Direction: the withdrawn arithmetic ran WITH this review’s position, because it was constructed to show that a rival explanation was too small to reconcile the heterogeneity this review reports as its finding.

Tromsø cannot adjudicate the assumption behind that calculation, and it is worth saying why rather than leaving it implicit. Its crude rates rise monotonically across its three bands — 1.47, 1.60, 1.90 per 1,000 person-years in men and 1.11, 1.49, 2.57 in women — with no elevation visible at the bottom. But its lowest band is 40 percentile-points wide, and a 5-point stratum elevated by a fifth — ⚠ a magnitude imported from ARIC and not measured in Tromsø — would raise that band’s rate by around 2%, which is undetectable against a rate whose own interval runs 1.16 to 1.86. Tromsø is consistent with the low-tail pattern and consistent with its absence. It is not evidence either way.

Every structural non-comparability in this literature is recorded here as a limitation on what the evidence can support, and none is used to reconcile the heterogeneity. ⚠⚠ Where a difference can be approximated WITHIN a single cohort, the arithmetic is given and its scope is stated; it is not carried across cohorts to produce adjusted estimates. Where it cannot, it is stated and left open. The adjustment mismatch is the clearest case: removing it does not narrow the gap between ARIC and LITE but widens it, from 1.72-against-1.03 to 1.95-against-1.03. Copenhagen’s regression-dilution correction, ratio 0.63, applies to one row and not the other three, so the four estimates are not on a common measurement-error footing. And the provoking-factor definitions, below, are not the same partition.

The two cohorts that disagree most sharply on the hazard ratio report the same absolute rate

ARIC’s top band ran at 4.4 incident VTE per 1,000 person-years against 2.4 in the reference band — 61 events in 13,856 person-years, in a band where 54% were current smokers against 23% in the reference. Copenhagen’s top band also ran at 4.4 per 1,000 person-years, against a reference of 3.0.

The disagreement between them is not in the top band at all. It includes what each cohort compared the top band to, though populations, models and follow-up also differ. This is the strongest single reason to read the fold-changes only alongside the rates that generate them.

LITE cannot be brought onto that footing. Its incidence rates by haematocrit band are age-adjusted rather than crude, it reports no events by band, and no crude rate for its upper band is derivable from the paper — its only crude rates are for the cohort as a whole and by study, sex and race. The absolute rate underlying LITE’s 1.03 is not estimable, and that is a property of the evidence base rather than an omission in this review.

The provoked/unprovoked contradiction cannot be adjudicated, because the two cohorts are not partitioning the same events

ARIC estimates were 2.05 (1.47–2.86) for provoked VTE and 1.21 (0.73–2.02) for unprovoked VTE on 17 events. Tromsø’s largest estimate was for unprovoked VTE at 2.37 (1.36–4.15) in men. The subgroup patterns differ descriptively, but neither study supplies a formal test establishing that the haematocrit association truly differs by provoking status. Copenhagen does not report the split at all, and LITE classifies all 215 of its events as idiopathic or secondary — 93 and 122 — but never analyses haematocrit against that classification. Only two of four cohorts can speak to it, and those two are the two whose subgroup patterns differ descriptively; neither their within-study results nor this comparison establishes between-study heterogeneity of effect (correction 661).

Tromsø is broader by category and narrower by window, and the category difference dominates: acute medical conditions accounted for 15.4% of its 447 cases and the catch-all for a further 4.0%.

The direction of that difference runs against the convenient reading, and the block states it. Tromsø removes acute myocardial infarction, stroke and major infection from its unprovoked stratum — which makes that stratum cleaner of precisely the smoking-related cardiopulmonary pathway ARIC’s authors invoke — and that is where Tromsø’s association is strongest. ARIC’s unprovoked stratum retains those events; the estimate crosses unity but spans a wide interval. Confounding by smoking does not straightforwardly explain either result. Whichever reading is preferred, neither cohort resolves an estimate anywhere near 54%, and the absolute difference at ARIC’s top band is 4.4 against 2.4 per 1,000 person-years.

The authors of the largest cohort read their own pattern as confounding, and supply their own counterweight

ARIC’s authors state that because it is usually inferred that risk factors confined to provoked VTE are less likely to be direct causes, their findings may be most supportive of high haematocrit standing for smoking or smoking-related heart or lung disease arising during follow-up — and against that, that a Mendelian randomisation study found a polygenic risk score for haemoglobin concentration positively associated with VTE, suggesting a possible causal relationship. Both halves belong in the same sentence: the confounding reading without the counterweight misrepresents the paper, and the counterweight without naming the quantity loses the point, because the score is for haemoglobin and the paper is about haematocrit.

The mechanism sentence needs care in paraphrase. Heart and lung disease are not on ARIC’s provoking list, which is cancer within a year and major trauma, surgery or marked immobility within 90 days. The cancer limb maps directly onto that definition, and the authors’ worked example is a cancer example. The cardiopulmonary limb reaches “provoked” only through an unstated intermediate step of hospitalisation, surgery or immobility.

The smoking adjustment runs the same way, and against the reading. Tromsø’s categorical estimates are adjusted for age alone — its multivariable model, which adds smoking and body mass index, is applied only to the continuous estimates — and its smoking variable is in any case dichotomous, which its authors identify as leaving them unable to remove dose-dependent confounding by smoking. ARIC adjusts for smoking status and pack-years, in a top band where 54% were current smokers against 23% in the reference. The cohort with no smoking adjustment at all returns the smaller estimate, 1.54; the cohort that adjusts most thoroughly returns 1.72.

Tromsø’s own check on underlying illness is available and belongs here with its limits attached. Information on other medical conditions was collected only for those who developed VTE. Among those men, the presence of other conditions did not differ across levels of haematocrit — reported as data not shown — while among the women it was slightly higher in the upper 20th centile. The authors state that with no clinical information on the reference group they lacked sufficient information across the total population to consider underlying disease in detail, and they claim nothing from the check. It is case-only, the male null is unverifiable, the female result runs toward confounding, and writing any of it as reassurance would misrepresent the paper.

Against the confounding-by-illness reading at the top tail, Copenhagen’s own sensitivity analysis cuts the other way. Excluding every participant with any Charlson comorbidity (n = 22,859) strengthened all three of its associations rather than attenuating them: VTE from 1.21 (0.88–1.62) to 1.34 (0.92–1.92), myocardial infarction from 1.42 (1.05–1.92) to 1.58 (1.11–2.30), and stroke from 1.29 (0.92–1.77) to 1.42 (0.97–2.09). Excluding myeloproliferative neoplasms (n = 320) left all three essentially unchanged: the venous estimate moved from 1.21 (0.88–1.62) to 1.17 (0.87–1.60), and the myocardial infarction and stroke point estimates did not move at all.

And the counterweight to that, also Copenhagen’s own. Most candidate confounders in its Table 2 are U-shaped across the haematocrit bands, in the same shape as the outcome: any comorbidity 34, 22, 19, 20, 24%; antiplatelet therapy 20, 13, 10, 11, 14%; median age 63, 58, 57, 59, 62 — and male sex, ever-smoking, cumulative pack-years and C-reactive protein all follow the same shape. Two do not. Body mass index rises monotonically across the bands at 24.9, 25.0, 25.6, 26.3, 26.8 kg/m², and alcohol consumption above the Danish thresholds at 36.1, 39.1, 39.1, 39.7, 40.9%. The U-shape is the dominant pattern in that table and not a universal one, and an earlier form of this sentence claimed it was universal. Its authors report that adjustment barely moved the platelet analyses while moving the haematocrit analyses substantially, and conclude that the haematocrit estimates should be considered a maximum estimate. Both halves stand.

Every strand of this evidence base is on record from its own authors as not licensing inference to a therapeutic increase

The exposure throughout is a single baseline measurement of a trait, not an intervention on it. Three of the four cohorts have no repeat measurement and no regression-dilution correction, and two of the three say so: ARIC’s authors state that fluctuation over follow-up would tend to weaken the observed association, and Tromsø’s argue the same direction. Copenhagen alone escapes the objection, and quantifies at a ratio of 0.63 what Tromsø only asserts.

The genetic evidence carries its own version. The Mendelian randomisation authors state that small lifelong genetic differences are qualitatively and quantitatively different from large short-term changes, which may limit the applicability of their findings to intervening on haemoglobin.

Neither the cohort evidence nor the genetic evidence, on its own authors’ account, supports inference from an observed trait level to the effect of raising that level therapeutically.

Nothing in this evidence base separates haemoglobin from haematocrit

⚠ None of the cohorts analysed here separates the two. Tromsø finds haemoglobin and red cell count positive with mean corpuscular volume yielding an estimate crossing unity. LITE finds a haemoglobin estimate of 0.97 (95% CI 0.64–1.46). ARIC finds haemoglobin positive at 1.47 (1.10–1.95) and haematocrit stronger at 1.72 (1.30–2.27) — and ARIC did not measure red blood cell indices at all, so it could not have separated them. The Mendelian randomisation separates them only weakly: the genetic associations with the two traits correlate at r = 0.91.[43] Haematocrit’s marginal inclusion probability rises from 0.275 to 0.706 under a relaxed prior, and the authors cannot exclude that haemoglobin was selected for being the more accurately measured trait.

Tromsø is the only cohort that explicitly ranks haematocrit above haemoglobin, and it is worth being precise about why the ranking does not hold.

Aside — ranking predictors across non-standardised increments (this review’s arithmetic). Tromsø’s stated basis for preferring haematocrit is that it produced stronger risk estimates. The increments used — 5% haematocrit, 1 g/dL haemoglobin, 0.5 × 10¹²/L red cells — are conventional clinical units, and choosing them is entirely reasonable. The defect is in ranking three predictors by hazard ratios computed on three different increments and reading the largest as the strongest. Taking the male standard deviations from Tromsø’s Table 1 (haematocrit 3%, haemoglobin 0.9 g/dL, red cell count 0.4 × 10¹²/L) and assuming log-linearity, the multivariable male estimates of 1.33, 1.17 and 1.20 rescale to approximately 1.19, 1.15 and 1.16 per standard deviation. On a common scale the three predictors become similar as point estimates, which the published data do not permit a formal test of. Their numerical ordering remains; what the reported data do not establish is meaningful superiority of one predictor over the others, and an earlier form of this sentence said the ranking disappears (correction 661). ⚠ The rescaling is an approximation, it is confined to Tromsø’s own data, and it is not a substitute for the per-standard-deviation analysis Tromsø did not run.

The point earns its space because Tromsø’s conclusion that haematocrit is preferable to related haematological variables for assessing VTE risk is the closest thing the primary literature has to an endorsement of monitoring haematocrit specifically.

The substitution also occurs inside the primary literature and not only downstream of it. ARIC reports haematocrit at 1.72 as the stronger phenotypic association, then defends causality with a polygenic score for haemoglobin, and the sentence that performs the transfer states that the score suggests a possible causal relationship between haemoglobin or haematocrit and VTE. The disjunction is the substitution, made in a single clause.

The cohorts examined here in relation to the apparatus did not agree that the measure was clinically useful

Two kinds of position appear across these cohorts’ own discussions, unevenly weighted, and are reported here as the authors’ assessments rather than as recommendations. Two are mild endorsements: Tromsø concludes that haematocrit should be taken into consideration in risk assessment for VTE, and the young-men’s cohort in the arterial literature reaches a comparable view. One is conditional: ARIC’s authors write that further study could address whether haematocrit or haemoglobin in the upper tail might be helpful risk stratification for considering VTE prevention in high-risk settings for provoked VTE.

Copenhagen supplies no position on this outcome, and an earlier form of this passage gave it one. Its authors’ sentence — that it seems unlikely adding haematocrit would improve risk prediction for this disease — sits in a paragraph about arterial thrombosis in the heart, and this disease is myocardial infarction. Copenhagen makes no statement anywhere about risk prediction for venous thromboembolism. The sentence is quoted accurately in §9.4.1, where its referent is the outcome under discussion, and it was quoted accurately here too. What changed when it moved between blocks was not a word of it but what it was about.

The value of the record is provenance, not adjudication. Among the general-population cohorts examined here in relation to the apparatus, the authors did not agree that the measure was clinically useful, and two of the four had reservations.

Two sources held in the archive are excluded from this block

The 2010 Haematologica commentary accompanying Tromsø in the same issue is an editorial with no patients, no data and no analysis, and is secondary in its entirety; its reproduction of the LITE band characterisation is not independent attestation of it. The 2021 correspondence letter on thrombosis in secondary erythrocytosis is a retrospective chart review of 102 patients whose exposure is aetiological category rather than haematocrit level, with no banding, no adjusted model and no time-to-event analysis; its post-diagnosis thrombosis contrast of 0 of 36 against 9 of 66 rests on median follow-up of 10.2 months in the first group against 68.2 months in the second, and a zero over 36 patients and ten months is close to uninformative.

Graded statements

Association between upper-tail haematocrit and venous thromboembolism in general-population cohorts with no trans-masculine subgroup identified or analysed — LIMITED.

LIMITED here reflects divided evidence, not thin evidence. The four cohorts are large and well conducted, contributing 800, 447, 2,524 and 215 events — though 130 of LITE’s 215 are ARIC participants in an earlier window of the same follow-up, so the four counts are not four independent contributions — with registry-complete follow-up in three of the four. The grade records that they do not agree, and that no structural feature of their designs orders the disagreement — not a judgement on how they were done. The scope clause is part of the grade: the evidence at the two tails is sharply different, and a single unscoped grade would conceal that.

Causal contribution of haematocrit specifically, as distinct from haemoglobin or from what elevated haematocrit stands for — NOT ISOLATED.

No cohort separates haematocrit from haemoglobin, the genetic evidence separates them only weakly and selects haemoglobin, the largest cohort’s own authors read their pattern as consistent with confounding, and every strand is on record as not licensing inference from an observed level to a therapeutic increase.

Threshold effect at 54% — NOT ESTIMABLE. No cohort estimates risk at that value separately: every uppermost band begins at or below 49.4% and is nowhere subdivided above it, so any estimate formally covering 54% is the same estimate that covers its band’s floor. Not contradicted, not supported, not separately estimated.

ARIC’s band is not defined by a haematocrit value at all. Its top category is the 95th–100th race-sex centile, with cut points at 49.4% in white and black men, 44.9% in white women and 44.0% in black women — so the lowest subgroup-specific cut-point is 44.0% in Black women while both male cut-points are 49.4%, so the estimate combines subgroup-specific upper tails rather than risk above one common haematocrit value, and its highest spline knot sits at 48.1%, the 94th centile.

Dose-response shape belongs under this statement rather than beside it. ARIC’s spline is non-linear with knots at 35.6, 39.3, 41.6, 44.1 and 48.1; Copenhagen’s association is U-shaped and roughly symmetric; ARIC’s isolated DVT below the 5th centile runs at 1.69 (1.13–2.55), significant in both models. Tromsø’s banding is too coarse at the bottom to speak to the low tail at all, though within its unprovoked stratum in men the gradient is monotone across all three bands, at 1.00, 1.81 (1.08–3.03) and 2.37 (1.36–4.15). A monotonic upper-tail association would provide one rationale for an upper action threshold, but neither monotonicity nor a specific cutoff is established by these observational data.

9.4.3 Trans-masculine events — thirteen reports, thirteen descriptions of affected participants, and denominators that cannot carry the conclusion

⚠⚠ REBUILT AT v4.5, AND THE REBUILD IS CORRECTION 36. The v4.4 subsection carried a three-cohort, a four-cohort and a five-cohort account of its own evidence base, all live at once: a rebuilt opening and a five-row table stating five integrated cohorts, a synthesis beneath them saying one found five, one found three, and one found none and introducing a Chilean cohort as a fourth, and a grading block closing on the three cohorts. ⚠ The synthesis between the table and subsection (a) is DELETED AND REGENERATED FROM THE TABLE rather than amended, and every statement downstream now refers either to five cohorts or to a named subset.

⚠⚠ THE RULE ADMITTING A REPORT TO THIS SUBSECTION, STATED BEFORE THE REPORTS ARE LISTED. §9.4.3 IS A SUBCORPUS, AND THIS IS ITS SUBCORPUS RULE — not the review’s inclusion rule. ⚠ The Methods admit any trans-masculine testosterone study reporting a thrombotic-event count, and that rule is unchanged. A source held out of §9.4.3 is NOT excluded from the review. A report enters the §9.4.3 subcorpus only if it gives BOTH (i) thrombotic-event status AND (ii) a quantitative red-cell parameter — haematocrit or haemoglobin — in the same trans-masculine, testosterone-exposed population. ⚠ The ground is the subsection’s own question. This subsection asks whether the treatment-induced haematocrit rise causes thrombosis. A report giving events without a red-cell value, or a red-cell value without event ascertainment, cannot bear on that question in either direction, and admitting it would inflate the corpus without adding information. ⚠ Haemostatic markers do not satisfy limb (ii). Platelet counts, fibrinogen and coagulation times are haemostatic rather than erythropoietic; they are §9.5’s subject and are treated there. ⚠⚠ Every identified report that fails this rule is recorded below as an adjudicated exclusion with its ground stated. The rule is narrower than all published thrombotic events on testosterone, and that narrowness is deliberate and declared rather than concealed.

⚠⚠ THIS SUBSECTION IS NOT A REVIEW OF WHETHER TESTOSTERONE CAUSES THROMBOTIC EVENTS. That question is reserved for Tranche 2, and large treated-versus-untreated trans-masculine outcome studies belong there. It asks the narrower question whether trans-masculine studies examining erythrocytosis report clinical events in a form that connects those events to haematocrit.

⚠⚠ Within that scope, THIRTEEN REPORTS identified in this review give thrombotic-event status alongside quantitative haemoglobin or haematocrit in trans-masculine participants. Seven contain the thirteen DESCRIPTIONS of affected participants set out below, whose uniqueness across reports is not established; the rest report no thrombotic event. None estimates thrombotic risk as a function of haematocrit.**

⚠⚠ THE UNIT IS REPORTS AND NOT COHORTS, AND THE CHANGE IS CORRECTION 63. Several of these analyses draw on the same Amsterdam and Ghent clinical populations, and participant-level overlap between them has not been reconstructed — §9.4.6 carries the map. ⚠ Counting them as independent cohorts would inflate the evidence base, and a study count over an unreconstructed overlap does not mean what a study count normally means.

⚠⚠ THIS REVIEW STATES NO FIXED TOTAL FOR THAT CORPUS, AND THE REASON IS ITS OWN HISTORY. The number has been three, then four, then five across successive assemblies, each time because a further directly relevant cohort was identified after the count was written. ⚠ The corpus boundary is not known until the reproducible search is completed to protocol, and asserting a total implies a completeness this review has repeatedly failed to have. The thirteen set out below are those integrated here. ⚠ The count describes the table, not the field: it is a count of rows, not a claim that thirteen is what exists.

⚠⚠ THIRTEEN REPORTS ARE CURRENTLY INTEGRATED HERE, AND WHAT EACH CAN AND CANNOT ESTIMATE. THEY ARE NOT THIRTEEN INDEPENDENT COHORTS, AND THIS IS NOT A CLAIM OF FIELD COMPLETENESS.

⚠⚠ NONE OF THESE REPORTS ESTIMATES THE CAUSAL EFFECT OF HAEMATOCRIT LEVEL, THRESHOLD CROSSING, OR DURATION ABOVE THRESHOLD ON THROMBOSIS. That is the column that matters, and it is empty in every row.

⚠⚠ WHAT THE EVENT-TIME MEASUREMENTS SHOW ACROSS THE CORPUS, AND WHAT THEY DO NOT. Marked event-time erythrocytosis is documented once: Choy, haematocrit 62% with haemoglobin 19.3 g/dL at the retinal artery occlusion — the highest event-time value here. Connelly’s haematocrit was 47.2% at the NSTEMI. Oakes reports haematocrit at diagnosis for all five of its events — 38.8%, 43.7%, 45.4%, 47.7% and 35.2% — and states that none met its erythrocytosis definition at the time the event was diagnosed, although four of the five had met it earlier in their care. Chen reports haemoglobin 15.2 g/dL with no documented erythrocytosis. Iwamoto reports haematocrit 47% several months before the stroke, which is a pre-event rather than an event-time measurement. ⚠⚠ THAT HETEROGENEITY IS DESCRIPTIVE RESOLUTION AND NOTHING MORE. It does not show that raised haematocrit is harmless, and it does not identify haematocrit as a causal mediator. No report integrated here compares otherwise similar testosterone-treated people BY haematocrit exposure, models duration above a threshold, or estimates the effect of crossing 50%, 52% or 54%. ⚠ The Oakes observation in particular is the one most likely to be misread: that five events occurred without contemporaneous erythrocytosis constrains a simple contemporaneous-viscosity account, and says nothing about prior exposure, cumulative time above a threshold, or a lagged effect. The corpus increases what can be described and does not validate a numerical action threshold.

⚠⚠ HELD OUT OF THE SUBCORPUS, RECORDED SO THAT THE BOUNDARY IS AUDITABLE RATHER THAN ASSERTED. Each report below is identified, review-eligible event evidence, tested against the subcorpus rule above and held out of §9.4.3 on a stated ground. ⚠ None is excluded from the review; the broader thrombosis question is deferred to Tranche 2. None is held out for its direction.

⚠⚠ UNRESOLVED CANDIDATES — SIX, AND THE COUNT IS STATED HERE RATHER THAN BURIED. ⚠ Fan, Inoue, Mullins, Park-SOVT, Penney and Stanley. Retrieval was attempted for every one on 13 September 2026 and none was obtainable through this review’s available access routes. These are identified, review-eligible event reports whose SUBCORPUS eligibility could not be resolved because the full text was not obtained. None is excluded. ⚠ An abstract that does not report a haemoglobin or haematocrit is not evidence that the full report does not.

⚠⚠ WHAT THIS MEANS FOR THE COUNT. Thirteen is what the subcorpus currently holds, not what the field holds, and six identified candidates could still enter it. ⚠⚠ THIS REVIEW CLAIMS KNOWN-CANDIDATE DISPOSITION COMPLETE — every identified candidate has an auditable state. It does NOT claim ELIGIBILITY RESOLVED, which would require that no known candidate’s membership remains unknown, and which is false while Fan, Penney, Stanley, Mullins, Park-SOVT and Inoue are unretrieved. ⚠ The subcorpus is therefore NOT eligibility-resolved. Deposit gates on known-candidate disposition, which IS complete: every one of these six carries a formal unresolved-source exception. They are declared uncertainty, not unfinished deposit work (correction 444).

⚠⚠ THE SEVENTH ROW CARRIES A WARNING THAT THE OTHERS DO NOT. The 2021 Amsterdam analysis followed 100 trans men for twelve months and states that no thrombotic event was observed during follow-up, and that no participant received phlebotomy in that period.[27] ⚠⚠ It is added with the overlap warning attached rather than as an independent cohort: an Amsterdam analysis of 100 trans men sampled between August 2012 and September 2015 may substantially be the same people as appear in other rows of §9.4.6’s map, and the overlap has not been reconstructed. ⚠ An earlier version said it could not be added without that reconstruction. It can be added WITH the reconstruction declared missing; what cannot be done is counting it as independent.

⚠⚠ THE AFFECTED-PARTICIPANT TOTAL IS THIRTEEN DESCRIPTIONS, AND SUMMING THE ROWS CANNOT ESTABLISH THIRTEEN DISTINCT PEOPLE: uniqueness across reports is not established, as §12 records (correction 658). Seven reports with events contribute affected participants: five in one retrospective review, three in another, and one in each of five further reports; six report none. ⚠⚠ The number of thrombotic EPISODES cannot be recovered from the published reports, because one participant in the second cohort had recurrent transient ischaemic attacks and therefore contributed more than one.[30] Thirteen affected participants is a descriptive total across seven non-commensurable denominators. It is not a numerator over any shared population, and it is not an episode count.

⚠ A CORRECTION NOTICE, WRITTEN SO THAT A LATER SWEEP CANNOT REWRITE IT. An earlier version of this subsection reported the figure using the word events where it should have used the phrase naming people. That conflated episodes with people — the failure the Source Provenance Standard at limb B names. ⚠⚠ A subsequent global substitution corrected the claim and overwrote this notice at the same time, so the correction erased its own trace; and the version after that left the erroneous phrasing standing in one further sentence while the rest of the subsection carried the corrected form. A global replacement cannot distinguish a claim from the record of that claim having been wrong. Any such substitution must exclude the correction notices, and the notices must be re-read after the sweep rather than assumed intact.

⚠⚠ WHAT THE DENOMINATORS CAN AND CANNOT CARRY. Two of these reports have fixed observation windows and completer counts, so crude event frequencies are calculable — 57 people over ten years with none, and 66 over 24 months with none.[37][33] ⚠⚠ What is not available from any of them is a haematocrit-specific rate, risk ratio, or causal effect estimate. That is a sharper negative than the earlier form — the denominators cannot carry a rate — because it names the quantity actually missing rather than denying quantities that can be computed.

⚠ They cannot be pooled, and the reasons are structural rather than a matter of degree. One counts across a whole treated cohort; one counts only within the subgroup meeting its own erythrocytosis definition; one counts across a prospective cohort whose follow-up falls away sharply; and two count across fixed windows of very different length, one of ten years and one of two. Event ascertainment, denominators, observation windows and follow-up completeness differ materially across all of them, and several draw on overlapping populations.

⚠ AND THE AUTHORS’ OWN CONCLUSIONS DIVIDE. Of the three cohorts characterised in detail at (a) to (d) below, two draw a reassuring conclusion in their own words; the third describes polycythaemia as common and as a significant clinical concern, and this review does not attribute reassurance to it. ⚠⚠ None of these reports provides the person-time, comparator structure and event count needed for a precise incidence estimate or a causal comparison. That is the shape of the cohorts reviewed here, and a systematic search is required before treating it as the complete outcome literature.

(a) Where events were counted

A retrospective records review at a single academic centre identified 923 trans and non-binary adults prescribed testosterone between June 2017 and June 2020, of whom 519 had both pre- and post-treatment values documented. Mean age was 29; testosterone cypionate accounted for 837 of the 923 (90%). Mean haematocrit rose from 40·2% to a peak of 47·0% — a mean difference of 6·84 points — with mean time to peak of 31·2 months.[29]

⚠ The study’s threshold deserves stating before its results are used. Erythrocytosis was defined as haemoglobin above 17·5 g/dL or haematocrit above 50%, and the paper names this in its own Methods as the traditional reference range for men.[29] The two criteria then disagree in the same people by a factor of two and a half: 41 of 519 (7·8%) crossed the haemoglobin line, against 104 of 519 (20%) the haematocrit one. These two conventional but non-equivalent operational criteria classify markedly different proportions of the same cohort, demonstrating substantial definition sensitivity.

Thromboembolic events occurred in 5 of 519 (0·9%). Four of the five (80%) had met the erythrocytosis criterion at some point during their care. None met it at the time the event was diagnosed.[29]

⚠⚠ AND A SECOND COHORT REPORTS A RELATED PATTERN — BUT NOT THE ONE THIS REVIEW FIRST STATED, AND THE ERROR WAS THIS REVIEW’S. [18] cites a retrospective study of 1,180 hypogonadal men. The primary report distinguishes 694 receiving testosterone replacement from 486 receiving clomiphene: nine and one respectively developed deep vein thrombosis, and none was reported to be polycythaemic at diagnosis. The crude observed proportions are therefore 9 of 694 (1·30%) and 1 of 486 (0·21%); these are not person-time incidence rates and not adjusted treatment comparisons.[18]

⚠ What it does and does not establish. It does not establish the participants’ preceding haematocrit exposure, a threshold-specific risk relationship, or any effect of lowering haematocrit. It is indirect evidence for gender-affirming testosterone care, in a hypogonadal male population whose gender was not separately ascertained by the source.

⚠⚠ CORRECTION 483 RECORDS WHAT THIS PASSAGE SAID BEFORE AND WHY IT WAS WRONG. An earlier form described a cisgender cohort of 1,180 men on testosterone in which ten developed deep vein thrombosis — a mixed-treatment denominator read as a testosterone denominator, a combined event count read as a treatment-group count, and an ascertainment the source does not make. No claim of corroboration is made here.

What this does and does not support. It weakens a model in which contemporaneous threshold elevation is required for an event. It does not exclude a contribution from prior, intermittent or cumulative erythrocytosis — and that is the study’s own limitation, not this review’s gloss: the authors record that the duration of erythrocytosis was not identified, and that duration may matter more to thrombotic risk than the level reached.[29] Two of the five events followed reconstructive surgery, a setting in which haematocrit is depressed by blood loss and fluid replacement, so the event-date haematocrit in those two cases may have been biased downward by perioperative blood loss or fluid shifts; the report does not quantify that contribution. Four of the five occurred alongside recognised provoking or vascular factors; the fifth is recorded as having none. Co-existing causes complicate attribution; they do not remove it.

The second cohort with events, and its denominator is a subgroup

A single-centre retrospective chart review identified 511 trans-masculine patients on testosterone, of whom 113 met its polycythaemia definition of a haematocrit at or above 50%. Thromboembolic complications were a prespecified collected outcome — the study names deep-vein thrombosis, pulmonary embolism, myocardial infarction, cerebrovascular accident and transient ischaemic attack — and three of the 113 experienced one: two deep-vein thromboses, and one patient with recurrent transient ischaemic attacks.[30]

⚠⚠ The denominator is the polycythaemic subgroup, not the cohort. Three of 113 is not comparable with five of 519 or with zero of 192, because the first is conditioned on having crossed a haematocrit threshold and the others are not. Reporting three of 511 would be a different and misleading statistic, and this review does not report it.

⚠ The study does not relate the events to the haematocrit at which they occurred. It records that they happened in people who had at some point met its definition. That is the same limitation as the records review’s, arrived at from the opposite direction: one reports the haematocrit at the event and finds it low, the other reports the group and not the value.

⚠⚠ THIS COHORT WAS OMITTED FROM AN EARLIER VERSION OF THIS SUBSECTION AND THE OMISSION IS RECORDED RATHER THAN REPAIRED IN SILENCE. This review had opened and analysed the same paper in detail — for its oophorectomy finding, its multivariable model and its correction history — and did not carry its prespecified thromboembolic outcome into the subsection whose subject is thromboembolic outcomes. ⚠ Direction: the omission removed trans-masculine events from a review arguing that the relationship has not been estimated, and it was found by external review rather than by any internal check.

(b) Three limits on the counted events

The limits below concern the records review’s five events, which are the only ones reported with enough detail to characterise. ⚠ The chart review’s three are reported as counts and categories only, so nothing equivalent can be said about them — which is itself a limit.

The five are a heterogeneous composite. Superficial venous thrombosis, distal deep-vein thrombosis and arterial ischaemic stroke have different pathophysiology and severity, and superficial thrombosis is not interchangeable with venous thromboembolism as conventionally defined.

⚠ Both deep-vein events are recorded as non-occlusive, and both are confined to calf veins — anterior tibial in one, peroneal and posterior tibial in the other.[29] ⚠ Distal calf thrombosis generally carries a lower risk of pulmonary embolism than proximal deep-vein thrombosis, and superficial thrombosis lower still. Non-occlusive is a description of the clot, not a severity category, and this review does not use it as one. But the composite cannot be characterised as mild on that basis: the fifth event was an ischaemic stroke, of unreported severity, in a 35-year-old.

And the study has no comparator group and no person-time denominator. No incidence rate is derivable from it.

⚠ The five patients had been on testosterone for 49, 36, 207, 168 and 42 months at the time of their events. That range describes the five, and nothing else — the cohort’s own follow-up distribution is not reported, and the study window spans three years while individual exposure evidently ran much longer.

(c) What the authors conclude, and what this review does with it

The authors state that the thromboembolic risk in this population appears to be low.[29] This review does not adopt that conclusion, and the reason is structural rather than a disagreement about the count. Five events in 519 people with no comparator and no person-time is not an incidence, and 0·9% is not a rate. The authors’ own comparison runs the other way: they note that the risk in a predominantly young-adult sample is likely higher than in a similar-age demographic, while cautioning that cross-comparison is difficult.[29] A figure that cannot be converted into a rate cannot establish that a rate is low.

The inpatient limb does not repair this. A national hospitalisation sample identified 4,141 admissions involving transgender and non-binary patients, of which seven involved erythrocytosis and one a concurrent venous thromboembolic event.[29] The authors record that most care in this population is outpatient, so hospitalisation data may underestimate events — which makes the limb consistent with the records review without independently corroborating it.

(d) Where no events were counted at all

In the ENIGI prospective cohort, no trans man experienced a thromboembolic event during follow-up.[22]

⚠ The cohort is not event-free; the trans men in it are. Two trans women in the same study had events — a cerebrovascular accident at sixteen months in a 23-year-old, and thrombophlebitis at eighteen months in a 59-year-old.[22] The qualifier is load-bearing and is not a stylistic preference.

⚠⚠ The denominator is smaller than the headline suggests, and this matters more than the zero does. 192 trans men were analysed, and the Ghent protocol ran to 36 months — but the number actually under observation falls away across the schedule: 189 at baseline, 184 at three months, 96 at six, 88 at nine, 146 at twelve, 82 at eighteen, 67 at twenty-four, and 20 at thirty-six.[22] Amsterdam followed participants through the first year only, by the study’s own stated limitation, and beyond twelve months the cohort is Ghent — which prescribed testosterone undecanoate exclusively.

Twenty people reached three years. 192 participants over up to three years is true of the enrolment and of nobody’s follow-up. Person-time is not reported and is not constructed here. What can be said is that zero events against these counts is compatible with a non-trivial rate and excludes very little — and that it excludes still less than the enrolment figure implies.

The authors nevertheless draw the inference. Because erythrocytosis rates were low, the maximum measured haematocrit was 54·0%, and no trans man had an event, they state that they have no reason to assume the observed mild rise in haematocrit is associated with increased thrombotic risk in the short term.[22]

⚠ Set that beside the other two conclusions and the pattern is the section’s finding: three trans-masculine primaries characterised in detail here — one with five unclassifiable events, one with three counted only within its polycythaemic subgroup, and one with none — two of them reasoning from a small or absent numerator to reassurance, and the third describing the finding as a significant clinical concern. None is careless, and none is wrong about its own data. ⚠ The reassurance is an inference from sparse and structurally weak event evidence — five uncontrolled events without person-time or a comparator in one cohort, and zero events under limited and diminishing follow-up in another — and it is presented as a property of the treatment rather than of the study.

⚠ The two further zero-event cohorts add limited descriptive information and do not materially strengthen causal inference. A ten-year cohort of 57 and a 24-month cohort of 66 each report no thrombotic event; both are single-centre, uncontrolled and selected — 143 eligible reduced to 57 in one, and 398 treated reduced to 124 analysed across both groups in the other — and neither relates a thrombotic event to a measured haematocrit exposure.[37][33] ⚠⚠ An earlier version said they do not change that shape and are not treated as strengthening it, which was too categorical: a zero over 57 people across ten years is weak descriptive evidence, not an absence of evidence, and this review does not let weak evidence disappear into none. What remains true is that three zero-event cohorts of 192, 57 and 66, none of which relates an event to a haematocrit, do not become one larger negative by being counted together.

The same caution about secondary erythrocytosis is drawn independently within the trans literature, and drawn correctly. ENIGI’s discussion states that a causal effect has not been proven for secondary polycythaemia, and quotes a 2014 review concluding that there is no definite evidence that secondary polycythaemia increases thromboembolic risk.[22] [That conclusion is the cited review’s, transmitted by ENIGI; it is a secondary characterisation inside a primary paper and is used here as such.]

[Population flags. All of these reports are trans-masculine and all are directly on-population, which is unusual in this review and is the reason this subsection exists. But those that count events count them against thresholds derived from men — the records review says so explicitly, and ENIGI applies the 50% and 52% cut-offs from the hypogonadal-male literature. The populations differ from each other in ways that matter to baseline risk: mean age 29 against median 22·5, 90% intramuscular cypionate against a mixed undecanoate, ester and gel cohort against quarterly undecanoate in every participant, retrospective single-centre records against prospective multicentre follow-up.]

(e) Surveillance bias runs through all of this, and it cuts both ways

People on testosterone have full blood counts taken more often than the general population, which inflates detected prevalence. And detection prompts intervention: in the records review, 42% of those who developed erythrocytosis had their testosterone dose reduced and 4·8% underwent therapeutic phlebotomy.[29] The same monitoring that detects erythrocytosis can trigger dose reduction, venesection or other management, thereby altering subsequent haematocrit exposure. ⚠ Whether that intervention reduces thrombotic events is not established. Neither direction is measurable from these studies.

⚠ The records review adds a third direction that is easy to miss: it records that providers may have advised blood donation rather than scheduled phlebotomy, which the medical record would not capture.[29] Some proportion of the management is therefore invisible to the same records the outcomes were extracted from.

Only one cohort reviewed in this tranche addressed monitoring intensity formally, using inverse-probability weighting for the likelihood of having a blood test during follow-up. That is a methodological strength, and it is not reflected in how often that cohort’s hazard ratio is quoted relative to how often its method is described.

(f) Why these counts cannot be read across from anywhere else

The reports above give event status in trans-masculine people. None reports the person-time and exposure-specific comparator structure required for a useful incidence estimate, and the obvious remedy — borrowing a rate from a population where erythrocytosis has been studied properly — was addressed by only two sources in this review at the time this section was drafted. That characterisation was subsequently withdrawn at §9.5.4, where a non-trans secondary-erythrocytosis rate became available; that rate is also discussed there. ⚠ Both are excluded from §9.4.2 by R8.1 on grounds of study type, and both are retained here because §9.4.3’s subject is events rather than associations.

The first is a haematology-referral series that compared erythrocytosis by cause.[46] It examined 102 informative cases diagnosed between 1999 and 2019 — 36 with secondary erythrocytosis and 66 with WHO-defined polycythaemia vera. ⚠ [Population flag, mandatory: a Canadian multicentre referral population, not a trans cohort, and its exposure is aetiological category rather than haematocrit level — which is precisely why R8.1 excludes it from the venous table.] [Provenance flag: Correspondence, and the review does not treat it as a cohort study.]

Its two findings that bear here run in opposite directions and both are recorded.

Against reading secondary erythrocytosis as benign: thrombosis at or before diagnosis was comparable between the two groups — 9 of 36 (25%) with secondary erythrocytosis against 19 of 66 (29%) with polycythaemia vera, p = 0·68. The authors state that from the standpoint of thrombosis, secondary erythrocytosis may not be as benign as some reports have intimated.

Toward reading it as benign — and it does not survive its own follow-up: thrombosis after diagnosis occurred in 0 of 36 (0%) against 9 of 66 (14%), p = 0·004. ⚠⚠ But median follow-up was 10·2 months in the secondary group against 68·2 months in the polycythaemia vera group, and the authors say so themselves: limited follow-up in that cohort precludes accurate interpretation. A zero over ten months is not a lower rate.

⚠⚠ AND THE FINDING THAT MATTERS MOST HERE IS THE ONE THAT CONVERGES WITH §9.4.3(b). Among the nine polycythaemia vera patients who had a thrombotic event after diagnosis, median haematocrit at the time of the event was 42% (range 34–61·1), and only three of the nine (33·3%) exceeded 45%. ⚠⚠ SIX OF THE NINE OCCURRED AT OR BELOW 45%; THREE OCCURRED ABOVE IT, AND THE RANGE REACHES 61·1%. ⚠ This review does not say the events failed to occur at raised values — a third of them did. What the distribution shows is that the majority did not, which is a weaker and accurate statement. ⚠ It is the same shape as the records review’s five events, none of which met the erythrocytosis criterion on the date of the event. ⚠⚠ THAT OBSERVATION IS TREATMENT-CONFOUNDED AND CANNOT BEAR THE WEIGHT AN EARLIER VERSION PUT ON IT. In the polycythaemia vera cohort, 76% had undergone phlebotomy, 93% were on aspirin and 88% on cytoreduction. A low contemporaneous haematocrit in an intensively treated patient does not identify the antecedent exposure, and may instead show successful count control in a disease whose thrombotic risk persists through other pathways. ⚠ It shows only that thrombosis can occur despite a contemporaneous haematocrit below the treatment target. The earlier claim of convergence between two unrelated cohorts is withdrawn. This review does not offer that as a mechanism. It offers it as a reason why counting events against a threshold has not yet produced an answer.

The series adds one further observation that cuts across cause entirely: irrespective of aetiology, the variables clustering with thrombosis were age, hypertension, diabetes, obesity and hyperlipidaemia — the classic cardiovascular risk factors, not the haematocrit.

The second source is an editorial, and it appears for one reason: the boundary this subsection is about to draw is one the field drew itself, and what the field asserted is the single kind of claim an editorial can establish.[47] ⚠ R8.3’s exclusion of it as evidence about haematocrit and thrombosis is untouched. ⚠ [Provenance flag, mandatory: an editorial, secondary in its entirety under Amendment C as extended at R8.3 — the review’s own ruling that an editorial is secondary in its entirety, not only in its introduction (Appendix R). Nothing below is offered as a finding. It is offered as a record of how the field itself framed the question in 2010, and it is the framing rather than the authority that is being cited.]

Commenting on the Tromsø cohort’s report of a dose-response relation between haematocrit and venous thrombosis, its authors accepted the association as real and then declined to call it causal, naming as the live alternative that conditions such as smoking, lung disease or heart disease raise haematocrit and thrombotic risk in parallel but independently. ⚠⚠ And they stated, in terms, that studies in polycythaemia vera should not be used as a model to extrapolate the relationship between high haematocrit and venous thrombosis in the general population — the mechanisms being too many and too disease-specific. On the editorial’s own account, then, the three literatures this subsection might have borrowed from are not interchangeable: what is measured in polycythaemia vera does not transfer to the general population, and neither transfers to a young trans-masculine cohort on exogenous testosterone.

⚠ One figure in that editorial is NOT repeated here and the omission is deliberate. It attributes a smoking prevalence to the cohort it comments on which this review has been unable to locate in that cohort’s own tables, and §9 records the question as unestablished. The editorial’s argument does not depend on it and this passage does not carry it.

What follows for §9.4.3 is a boundary rather than a result. ⚠ The trans-masculine event counts are small, uncontrolled and without person-time; the nearest comparator literatures are, by the field’s own statement and by this review’s reading of them, not transferable; and where events were counted alongside the haematocrit at which they occurred, the majority did not sit at raised values — though a third of the polycythaemia vera events did, and this review does not claim otherwise.[46] ⚠⚠ None of this establishes that contemporaneous haematocrit is irrelevant to thrombosis. ⚠⚠ The threshold-subgroup cohort does not bear on that either way: it reports events within a threshold-defined subgroup and does not report the haematocrit at the event. That is not evidence of safety and it is not evidence of harm. It is the reason this subsection ends without a rate.

Grading block — §9.4.3, trans-masculine events

One statement. The outcome does not divide.

Thrombotic events in trans-masculine people on testosterone

Causal attribution — NOT ESTABLISHED. ⚠ None of these reports has a haematocrit-exposure-specific comparator. The affected participants number thirteen across seven sources, with the remainder recording none, and surveillance bias runs through the counts in both directions at once. ⚠ Only one of them reports the haematocrit at the event, and none of its five met that study’s own erythrocytosis criterion at diagnosis. The second reports its three events only as members of a threshold-defined subgroup, without event-time values; the remaining six recorded no event to attach one to. ⚠ Six reports record no event, but they draw on overlapping Amsterdam/Ghent populations and the observation periods differ substantially; six reports is not six independent negative cohorts.

Nothing here can attribute an event to a raised haematocrit, and nothing here excludes it.
>
> Observed effect size — NOT ESTIMABLE. ⚠ No haematocrit-specific thrombotic rate, risk ratio or causal effect estimate is available from any of them. ⚠⚠ The denominators are not commensurable: one counts across a treated cohort, one only within its own threshold-defined subgroup, one across a prospective cohort whose follow-up falls to twenty people by three years, and two across fixed windows of ten years and two years respectively. This is not a small estimate; it is the absence of one, and the subsection says so at length rather than reporting a proportion that would look like a finding.
>
> Clinical importance — NOT ESTIMABLE. No absolute risk can be attached to an exposure whose effect has not been estimated.
>
> Population directness — direct. ⚠ It is the smallest body of direct evidence in the review, and §10 now carries eight direct rows where it once carried one. An earlier version of this block called it the only such subsection; that is no longer true. ⚠ They are not described with one adjective: two are retrospective single-centre chart reviews; one is prospective and multicentre, with sparse and diminishing follow-up for clinical events; two are single-centre fixed-window cohorts, one over ten years and one over two, both heavily selected; two are prospective haemostasis reports, one over four months in fourteen participants and one over twelve months in 100 participants; one is a retrospective switch cohort (pellet formulation, n=30); two are single-participant case reports describing cerebral venous thrombosis; one is a single-participant case report describing a retinal ARTERY occlusion under unregulated supraphysiological exposure; one is a single-participant case report describing a myocardial infarction with angiographic multivessel plaque; and one is a single-patient case presented within a management review, describing an ischaemic stroke.

⚠ A residual the subsection carries and this block does not resolve. The referral series it holds reports thrombosis at or before diagnosis as comparable between secondary erythrocytosis and polycythaemia vera — 9 of 36 against 19 of 66, p = 0·68 — and thrombosis after diagnosis as 0 of 36 against 9 of 66. Those run in opposite directions and both are in the subsection. Whether either bears on grading is Eden’s; neither is used here.

9.4.4 What the expert bodies assert, and what has been measured

Everything above in §9.4 is evidence about outcomes. This subsection is not. It is an account of what the field’s own expert bodies say about the same question, and it is placed here because the gap between the two is the tranche’s subject.

⚠⚠ DIVISION OF LABOUR WITH §9.4.5, STATED AT v4.6 AND ENFORCED THROUGHOUT. This subsection carries the expert bodies’ DOCUMENTARY claims — what the panel asserts, recommends and cites. §9.4.5 carries the PRIMARY interventional and genetic evidence, and it is the single place where the panel’s central citation is verified against the trial and its nested sub-study. ⚠ An earlier assembly maintained two detailed accounts of the same anaemia-sub-study figures, the same trial exclusions and the same 54% protocol rule, in two subsections. They were concordant, and maintaining one scientific state in two places is precisely how this manuscript generated its earlier propagation failures. Where verification detail is needed here, this subsection points to §9.4.5 rather than restating it.

(a) What the document is

The European Expert Panel for Testosterone Research (PaTeR) convened in London in September 2024 and published a position statement on the cardiovascular safety of testosterone therapy in the light of the TRAVERSE trial.[54]

[Provenance flags, mandatory. The article’s own type line is REVIEW ARTICLE, and its data-availability statement records that no dataset was generated or analysed for it. It produces nothing; every quantity in it is a characterisation of another paper. The convention was funded by Androlabs, which also provided full funding for the publication and submission; the paper states that Androlabs had no involvement in developing its content. Sixteen authors across eight countries.]

[Population flag, mandatory. The document is about men with hypogonadism and ⚠ no trans-masculine subgroup is identified or analysed anywhere in it. The trial it is built on enrolled men aged 45–80 with a mean age of 63·3, a mean body mass index of 35, roughly 70% with type 2 diabetes and 90% with hypertension and dyslipidaemia — as the panel describes it. Nothing in this subsection is evidence about trans-masculine people, and it is not offered as such.]

(b) The panel’s position on haematocrit, and the four things it rests on

The panel accepts that testosterone raises haematocrit by stimulating erythropoiesis and that this raises concerns about hyperviscosity and thromboembolic risk. It then sets four sources against that concern.[54]

⚠⚠ THE FOURTH ROW CANNOT BE REPORTED IN THE PANEL’S OWN SUMMARY FORM. A point estimate near fivefold with an interval running from 0·83 to 28·91 is uninformative, not absent. Not found describes the significance test; it does not describe the evidence. Wherever this review touches that figure the interval travels with it, and this review does not repeat the phrase without it.

(c) One figure the panel supplies, now adjudicated against the primary

The panel reports that in the Australian T4DM trial, among more than 1000 men on intramuscular testosterone undecanoate, 22% on testosterone had a haematocrit above 0·52; that these samples were drawn under fasting conditions in a hot country; and that on repeat sampling in a non-fasting condition the incidence of an elevated haematocrit fell to 2·5%.[54]

⚠⚠ THE TRIAL IS NOW HELD AND HAS BEEN READ FIRST-HAND AND UNTRUNCATED, SO EACH COMPONENT OF THAT ATTRIBUTION CAN BE CHECKED DIRECTLY. An earlier version of this subsection declined the figure on the ground that the primary could not be obtained. That ground no longer holds, and the figure is declined for a stronger one. Correction 185.

⚠⚠ THE RELEVANT CLAIMS ARE DRAWN FROM A SINGLE PARAGRAPH OF THE TRIAL’S DISCUSSION, IN THE ORDER THE PANEL PRESENTS THEM. That paragraph reports the proportion reaching the safety trigger, then places it within a range observed in other studies, then describes what happened on repeat measurement.

The threshold. ⚠⚠ The 22% is correct and the value it is attached to is not. The trial reports 106 of 491 treated participants — 22% — reaching a haematocrit of 54% OR HIGHER, and states it four times: in the abstract, in its research-in-context panel, in its safety-trigger table, and in the discussion. ⚠ No figure at 0·52 appears anywhere in the trial report as held. The report’s only other haematocrit values are a baseline of 46% and 45% by arm, an exclusion criterion above 50%, and a mean change of 4 percentage points. ⚠ The appendices are not held, and the scope of that absence statement is the report as held.

⚠⚠ THIS IS §5’s DEFECT CLASS APPEARING OUTSIDE §5: THE CORRECT NUMBER CARRIED ACROSS AND ATTACHED TO THE WRONG THRESHOLD OBJECT.

The 2·5%. ⚠⚠ THE TRIAL REPORTS NO INCIDENCE AFTER REPEAT MEASUREMENT AT ALL. What it reports is that the trigger led to cessation in 26 participants, 25 of them in the treated arm, and that the remaining 86 either did not have the raised value confirmed on non-fasting repeat OR had already received their final injection when the trigger occurred — a disjunction the trial does not resolve and this review does not resolve for it. ⚠⚠ THE FIGURE 2·5% APPEARS IN THAT PARAGRAPH AS THE LOWER BOUND OF A RANGE — 2·5 to 40% — WHICH THE TRIAL GIVES FOR TREATMENT-LIMITING HAEMATOCRIT INCREASES IN OTHER STUDIES, CITED TO TWO REFERENCES THAT ARE NOT THIS TRIAL. It is not the trial’s own post-repeat incidence, and the trial does not present it as one.

The denominator. ⚠⚠ THE PANEL PLACES THE 22% AMONG MORE THAN 1000 MEN ON INTRAMUSCULAR TESTOSTERONE UNDECANOATE. THE TRIAL RANDOMISED 1007 MEN IN TOTAL AND GAVE TESTOSTERONE TO 504 OF THEM; THE OTHER 503 RECEIVED PLACEBO. The figure that describes the whole trial is presented as describing its treated arm, which is roughly twice the treated population. ⚠ And the 22% rests on neither number: its denominator is 491, which the trial defines as those in the treated arm with at least one haematocrit measurement during the study. ⚠⚠ THIS IS THE SAME SHAPE AS THE OTHER TWO COMPONENTS — A REAL FIGURE FROM THE TRIAL ATTACHED TO THE WRONG OBJECT — AND IT IS THE LEAST CONSEQUENTIAL OF THE THREE, BECAUSE IT CHANGES THE SIZE OF THE POPULATION DESCRIBED AND NOT THE DIRECTION OF ANYTHING. ⚠ It is adjudicated and recorded all the same, because a reader checking this review’s adjudication against the panel’s sentence would otherwise find a component this review passed over. Correction 188.

The fasting condition. Corroborated. The trial scheduled study visits in the morning after an overnight fast, with collection between 0700 and 1000 hours, and specified that a trigger be repeated in a non-fasting state. ⚠ The hot country gloss is the panel’s. The trial records six Australian centres and says nothing about climate.

⚠⚠ WHAT FOLLOWS FOR THIS REVIEW. The ninefold collapse on re-sampling does not exist in the primary. It is produced by pairing a proportion reported at one threshold with a bound from a range describing other studies. This review had already declined to carry the figure, and was right to, but for the wrong reason: it declined because the primary was unavailable, when the primary was in fact available and the attribution does not survive it.

⚠⚠ DIRECTION, STATED BECAUSE IT RUNS FOR THIS REVIEW. A further transmission defect in a secondary source is the most convenient class of finding this review can make, and it therefore carries the higher burden under §6.1. The burden is discharged textually rather than inferentially: the trial’s threshold is stated four times in its own words, and the range is explicitly attributed by the trial to other studies. Neither point requires this review to reconstruct what the panel did or why. ⚠ What the panel intended is not established and is not asserted.

(d) What the panel recommends, and what it says about its own thresholds

The panel recommends checking haematocrit every three to six months in the first year and annually thereafter, keeping levels below 54% to minimise thromboembolic risk; states a bar on initiating treatment where haematocrit is high, without attaching a number to it; and, where haematocrit exceeds safe limits, advises dose adjustment, a change of formulation, or temporary cessation, adding that phlebotomy may be warranted.[54]

⚠⚠ And in its own future-research section the panel states that refining the thresholds for haematocrit monitoring and intervention remains a priority, and that guidelines tailored to patients at higher thromboembolic risk are still needed.[54]

That is the hinge. A panel recommends a specific number, in a specific monitoring schedule, and separately records that the thresholds require refining and that the tailored guidance does not yet exist. Both statements are in the same document and neither is a slip. This review does not read the second as undermining the first. It reads them together as an accurate description of the position: the thresholds are operating values carried forward because clinicians need a number, not conclusions the evidence has established. ⚠ They are also, as §9.2 and §9.3 set out, derived from men treated for hypogonadism — and the panel applies them to that population, which is the population it is writing about. The extrapolation to trans-masculine care happens elsewhere, and it is not the panel’s act.

(e) One signal the body reports and the summary does not

⚠ In its account of the trial’s results the panel records a minor but significant increase in atrial fibrillation and non-fatal arrhythmias in the active arm, alongside modest lipid improvements and stable blood pressure and glycometabolic control.[54] It then lists among the trial’s limitations that endpoints such as arrhythmias and heart failure were underpowered for definitive conclusions, and elsewhere reports a 2024 meta-analysis in which the increased risk for arrhythmias and atrial fibrillation was not statistically significant.[54]

The abstract and the panel’s final position both state that testosterone therapy does not significantly increase the risk of major adverse cardiovascular events, and carry no arrhythmia caveat.

⚠ That is not a contradiction: atrial fibrillation is not a component of the major-adverse-event composite, so both statements can be true together. ⚠⚠ AND THE DIRECTION HERE RUNS WITH THIS REVIEW’S POSITION, SO IT IS STATED AT ITS WEAKEST DEFENSIBLE STRENGTH: the signal is reported in the body, and it is also qualified in the body. It is not concealed. What is observable is that it does not survive into the summary, and a reader who reads only the abstract does not meet it. ⚠⚠ An earlier version of this sentence read whether the underlying trial reports it as the panel does is not established here — TRAVERSE is unopened. That was stale from the moment §9.4.5 opened the trial: it is established, the panel’s arrhythmia figures verify exactly, and §9.4.5(d) records the verification. ⚠ A subsection stating that a source is unopened while a later subsection of the same document reports reading it is the state-propagation failure this manuscript spends its correction record on, and it is corrected here rather than quietly amended.

(f) What follows from the exhibit, in the world

⚠⚠ The panel records that the United States regulator is implementing label changes on the strength of the trial: the trial’s results added to all testosterone product labels, the limitation-of-use language for functional hypogonadism retained, and language relating to increased risk of adverse cardiovascular outcomes REMOVED FROM THE BOXED WARNING.[54]

The regulator describes a boxed warning as its most prominent safety warning. ⚠⚠ THE FDA ANNOUNCED TWO STAGES OF LABELLING ACTION, BOTH OF WHICH THIS REVIEW RECORDS.

In February 2025, following review of the TRAVERSE trial and required post-marketing ambulatory blood-pressure studies, the regulator informed sponsors of class-wide changes: adding the TRAVERSE results to all testosterone products, retaining the limitation-of-use language for age-related hypogonadism, and removing from the Boxed Warning the language on increased risk of adverse cardiovascular outcomes. The blood-pressure studies drove a separate requirement for product-specific increased-blood-pressure information and a new warning where none existed.[50]

In June 2026, following a further review of new and existing evidence including TRAVERSE, FDA requested updates to testosterone prescribing information: removing the limitation of use for age-related hypogonadism, and revising the information on prostate cancer and benign prostatic hyperplasia.[51]

⚠ The June 2026 action is a request for labelling updates, not a completed class-wide change, and this review does not describe it as one. ⚠ REGULATORY STATUS OF GAHT: FDA testosterone-product actions concern approved testosterone indications in males (hypogonadism as defined by the indication); they are not FDA determinations of safety or efficacy for masculinising gender-affirming hormone therapy, which is not an approved indication. ASRM explicitly notes this distinction.[5] ⚠⚠ Both actions concern testosterone replacement in men. Neither establishes the validity of any haematocrit action threshold in trans-masculine care, and neither mentions haematocrit thresholds at all. ⚠ The evidence that moved it was not a trans-masculine population, in men whose mean age was 63 with a mean body mass index of 35 and a heavy burden of diabetes and hypertension — as the panel itself describes the population, and as the panel itself lists among the trial’s limitations on generalisability.

This review takes no position on whether the labelling change is correct. What it records is the chain: a trial in older comorbid men, a position statement built on it, a regulatory warning withdrawn — with no step in the chain having examined a trans-masculine population. ⚠ Those developments may alter wider perceptions of testosterone safety, but this review has not traced their transmission into trans-masculine guidance or practice, and §9.6’s standard requires the transmitting and receiving sources to be opened and compared before such a claim is made. ⚠⚠ The haematocrit thresholds themselves predate that sequence entirely and cannot be downstream of it. That gap is not the panel’s failing. The panel is writing about the men it studied.

(g) A second panel, and the field’s own account of the disagreement

The document devotes a section to comparing itself with a position statement from the United States Androgen Society, and the comparison is the panel’s own.[54] It characterises the other statement as declarative in tone, framing the trial as a definitive resolution, critiquing the earlier studies that suggested harm, calling for regulatory reversal, and largely dismissing the need for further large-scale cardiovascular trials — a scientific and regulatory manifesto, in the panel’s phrase. It characterises its own as more cautious and clinically integrative — a clinical roadmap — acknowledging the trial’s older high-risk population and relatively short follow-up.

⚠ Both are secondary interpretations rather than independent estimates of a treatment effect. They provide documentary evidence about expert interpretation and recommendations, but do not add a new trial population or resolve the underlying causal uncertainty. What the comparison establishes is that the two most prominent expert statements in the field agree on the headline and differ on how far it settles the question — and that the more cautious of the two is the one carrying the arrhythmia signal, the claims-database analysis THIS REVIEW HAS NOT OBTAINED, and the statement that its own thresholds need refining.

Grading block — §9.4.4, what the expert bodies assert

Not a graded outcome. This subsection reports what funded expert panels assert and what their assertions rest on. That is a statement about published documents, not a finding about the world, and grading it for causal attribution would be a category error.

Documentary — VERIFIED. That the panel states the position described, recommends what is described, and cites what is described. Each quantity is attributed to its primary and no sentence in the subsection asserts an outcome of its own.

⚠ This follows the Presentation Standard §5.1 and ER-015’s own use of the same line. ⚠⚠ It is not a weaker grade than the axes. It is a different kind of statement, and treating it as a gap in the grading would misdescribe the subsection.

9.4.5 The interventional evidence, and the exposure it removed

Compiled 13 August 2026 UTC, ER-023 Part 30, the eleventh sitting.

Provenance: FIRST-HAND THROUGHOUT. Every figure below was verified against its source open in the step it was written. Gate run as one batch before any read (requested 9, held 8, one deliberate absence check returning missing — see the residuals block). Reads: TRAVERSE id 139, Hudson id 147, Zitzmann id 144 and Luo id 134, all four truncated: false, verified before any content claim and not after. ⚠ TRAVERSE was RE-OPENED rather than written from the 13 August 09:43 read of the previous sitting, because the governing rule is open in the STEP, not open in the session. ⚠ EXTENDED 15–17 August 2026 with the post-Pencina corrections: Pencina id 140, the TRAVERSE Anemia Study, was gated and opened first-hand and untruncated in the correcting step, together with Zitzmann id 144 and TRAVERSE id 139 re-opened (requested 3, held 3, gate_passes: true). ⚠ A first Zitzmann retrieval at a low character limit returned truncated: true and cut the funding block; it was re-run at full length and NO content claim was made from the truncated return.

(a) What this subsection carries, and what none of it was built to answer

§9.4.1 to §9.4.3 report associations and events observed in people who were not assigned to anything. This subsection carries the evidence that comes from intervening: the largest randomised trial of testosterone with cardiovascular safety as its primary endpoint, the largest individual-participant meta-analysis of randomised testosterone trials, and a genetic-instrument study that stands at the interface of the two.

⚠⚠ NONE OF THE POPULATIONS IN THIS SUBSECTION IS A TRANS-MASCULINE COHORT, AND THE FLAG IS NOT A FORMALITY HERE. [52] enrolled men aged 45 to 80 with hypogonadism and established or high cardiovascular risk. [53] pooled trials of men aged 18 and over with a screening testosterone at or below 12 nmol/L (350 ng/dL). [43] used 265 424 genetically verified white British participants of European ancestry, 123 809 men and 141 615 women, mean age 56·9 years. ⚠ No trans-masculine subgroup is identified or analysed anywhere in this evidence base, and ⚠ the randomised treatment evidence here concerns men with hypogonadism, predominantly older and, in the cardiovascular-safety trial, at high vascular risk. The genetic analysis is different in kind: a general-population study in a national biobank, not a treatment cohort. Their doses, indications, baseline haematocrits and comorbidity profiles are those of hypogonadal older men, not of masculinising hormone therapy.

⚠ And a second limit, which the sources state about themselves rather than this review stating it about them: none of these studies was designed to answer whether a raised haematocrit causes thrombosis. Two measured events in people who happened to have raised haematocrits; the third measured lifelong genetic differences, not therapeutic ones.

(b) The randomised evidence, pooled — and the dissociation it reports

[53] is a systematic review and individual-participant-data meta-analysis: 35 trials, 5601 participants, reported across 109 peer-reviewed publications. 17 trials (49%) supplied individual participant data covering 3431 participants from nine countries; 18 did not, two of them because their pharmaceutical sponsors declined to release the data. Mean participant age 65 (SD 11); mean body-mass index 30 (SD 5); mean duration of testosterone treatment 9·5 months, range 12 weeks to 3 years. Risk of bias was judged low in 12 of the 17 data-supplying trials and unclear in the remaining five. It was funded by a national health-technology assessment programme, and the funder is recorded as having had no role in design, collection, analysis, interpretation or writing.

What it establishes about the red cell, from its one-stage analysis:

⚠ FM1 BINDS. [53]’s Discussion describes this as a five-times higher risk of polycythaemia. That fold-change may not be written anywhere without 2·8% against 0·5% beside it. The ratio computes to 5·6, so the rounding is the source’s own and not a transcription defect.

And what it reports about events:

⚠ THE LABEL ON THAT LAST ROW IS NOT STABLE INSIDE THE SOURCE, AND THIS REVIEW STATES THE DISCREPANCY RATHER THAN PICKING A CATEGORY. [53]’s Table 4 heads the row venous thromboembolism; its Discussion renders the same five and seven as deep vein thromboses. The two categories are not co-extensive. Which was adjudicated cannot be established from the paper, and picking one would conceal a characteristic of the source.

⚠⚠ AND THE SENTENCE THAT MATTERS MOST TO THIS REVIEW RUNS AGAINST ITS POSITION. IT IS RECORDED HERE AT FULL STRENGTH AND IN THE SOURCE’S OWN TERMS. [53]’s Discussion states that testosterone treatment did not increase thrombotic events despite increased haematocrit. ⚠ That is a TREATMENT-LEVEL dissociation: randomised assignment raised the haematocrit and DID NOT DETECT an excess of venous events among sparse counts. ⚠⚠ It is NOT an individual-level analysis of whether participants with larger haematocrit rises had greater thrombotic risk — no such model is supplied — and an earlier version of this passage described it as one. ⚠ The change-in-haematocrit analysis discussed below is more directly on the exposure-response point, in a far more restricted subgroup. Little else in §9.4.1 to §9.4.4 is as close, and no observational finding in this tranche outranks it in design.

Its limits come from the same source and travel with it:

• The events are twelve in total across four trials. A comparison of 5 against 7 has essentially no resolution, and [53] does not claim otherwise.
• Mean follow-up 9·5 months, which [53] itself records may be too short for atherosclerotic plaque progression to accrue.
• [53] states that there was little available data evaluating the cardiovascular safety of testosterone beyond a twelve-month duration of administration. Its Interpretation records a paucity of data on long-term safety; its research-in-context box states that long-term safety is not yet established. ⚠ That limitation is the primary’s own. This review cites it from the primary and not from any commentary on it.
• [53] states that it cannot exclude that a high number of unreported cardiovascular events in the 18 trials that supplied no individual data would ultimately change its conclusions.

(c) The single large trial — and the exposure its protocol removed

[52] is a phase 4, multicentre, randomised, double-blind, placebo-controlled noninferiority trial at 316 sites. 5246 men underwent randomisation; the safety population was 5198, being 2596 assigned testosterone and 2602 placebo. Intervention: daily transdermal 1·62% testosterone gel, dose adjusted centrally to hold serum testosterone between 350 and 750 ng/dL. Mean duration of treatment 21·7 ± 14·1 months; mean follow-up 33·0 ± 12·1 months.

⚠ POPULATION AND FUNDING, BOTH STATED BEFORE ANY RESULT. Men aged 45 to 80, mean age 63·3 ± 7·9, with 47·7% and 46·5% of the two groups aged 65 or over; mean body-mass index 35·0 and 34·8; diabetes 68·7% and 70·8%; hypertension 93·2% and 92·3%; dyslipidaemia 90·1% and 89·6%; all with preexisting cardiovascular disease or elevated cardiovascular risk. The trial was funded by a consortium of testosterone manufacturers led by the lead sponsor; nine of its twenty-five authors are employees of that sponsor; and the sponsor reviewed the manuscript and provided suggested revisions, with the final decision on content reserved to the academic authors.

The endpoints, from Table 2 of the safety population:

[52] records in its own Discussion that the arrhythmia, atrial fibrillation and acute kidney injury excesses were not expected.

⚠⚠ THE MOST IMPORTANT THING THIS TRIAL GIVES THIS REVIEW IS NOT AN OUTCOME. IT IS A PROTOCOL.

Dose adjustments to respond to a haematocrit greater than 54% were managed centrally by an automated algorithm. Testosterone or placebo was discontinued in patients whose haematocrit exceeded 54% even after adjustment to the lowest dose.

The trial therefore suppressed, by design and centrally, the exposure it is most often cited to reassure about. It cannot say what happens to men who sustain a haematocrit above 54%, because on detecting one it first reduced the dose and then withdrew the drug. ⚠ This is §9.4.3(e)’s surveillance-and-intervention mechanism operating inside a randomised trial rather than inside a clinic, and it bears on §9.3’s threshold argument as much as it bears here.

⚠ The same point from the other side: [52] reports NO on-treatment haematocrit result at all. Haematocrit appears twice in the paper — as a baseline value, 42 ± 4% in both groups, and as the dose-adjustment and discontinuation trigger. There is no table, figure or sentence giving what haematocrit did on treatment. ⚠ The main cardiovascular-safety report provides no on-treatment haematocrit data and therefore cannot by itself support a claim about that variable’s consequences. The only analysis within the programme that models change in haematocrit against outcomes is the nested one discussed below.

And the trial states its own limits, which are recorded here because omitting them would leave the reader with a stronger conclusion than the source supports:

• Its modified intention-to-treat analysis may attenuate differences between trial groups and bias toward noninferiority where patients are not adherent — the trial’s own words for its own design.
• 61·4% of the testosterone group discontinued testosterone; the assigned regimen was received for 67·5% of potential treatment time; 1082 patients (20·8%) withdrew.
• Enrolment was truncated twice — the cardiovascular-risk-factor cohort closed in May 2019, enrolment terminated February 2022 — finishing at 5246 against estimated requirements of roughly 5400 for the primary analysis and roughly 6000 for the principal sensitivity analysis.
• ⚠⚠ The confidence intervals are unadjusted for multiplicity, and [52] states plainly that they should not be used to infer definitive treatment effects for secondary endpoints. That disclaimer governs the venous thromboembolism row above and this review applies it.

⚠ EXCLUSIONS, AND ONE OF THEM IS LOAD-BEARING FOR (d): the trial excluded severe hypogonadism below 100 ng/dL, prostate cancer or nodules, an elevated screening prostate-specific antigen, thrombophilia, and uncontrolled heart failure. [52]’s own Discussion states that most reported cases of thrombosis associated with testosterone therapy have been in men with underlying thrombophilia. The trial excluded a subgroup repeatedly implicated in case reports and clinical concern, and still reported a doubled count of pulmonary embolism.

✅ RECORDED WITH EQUAL WEIGHT: [52] cites [53] accurately. Its Discussion attributes to a meta-analysis of randomised trials the finding of no detected association between venous thromboembolic events and testosterone use in wider populations, and [53]’s figures are 5/1037 against 7/1034. That transmission is faithful.

(d) What the expert panel asserts, and what it rests on

⚠⚠ R1.6 GOVERNS THIS SECTION IN FULL: [54] IS AN EXHIBIT AND NEVER EVIDENCE. NO SENTENCE BELOW ASSERTS AN OUTCOME. EVERY QUANTITY IS ATTRIBUTED, AND PROVENANCE AND POPULATION FLAGS ARE INLINE.

[54] is a Review Article carrying a position statement of a European expert panel, Andrology 2026;14:294–302 — received December 2024, revised April 2025, accepted April 2025. Sixteen authors across eight countries. Its convening convention and the full funding of its publication and submission came from a commercial sponsor, which the paper states had no involvement in developing its content. Its data-availability statement records that no dataset was generated or analysed. Its population of concern is men with hypogonadism throughout.

The panel’s central reassurance: that findings from the trial confirmed that while testosterone therapy modestly raised haematocrit, this did not lead to a significant rise in thromboembolic events among participants.

⚠⚠ THE CITATION IS SOUND, AND THIS REVIEW STATES THAT FIRST BECAUSE IT HAD PLANNED TO SAY OTHERWISE. The sentence carries the panel’s reference 17 — [55], the TRAVERSE Anemia Study, JAMA Netw Open 2023;6(10) — and not its reference 7, the cardiovascular-safety report. Reference 17 is the correct source and reference 7 would have been the wrong one, because the main trial reports no on-treatment haematocrit result at all: haematocrit appears in it only as a baseline value, 42 ± 4% in both groups, and as the dose-adjustment and discontinuation trigger.

And the sub-study ran the test. Among its stated aims was whether changes in haematocrit were associated with the risk of major adverse cardiac events and venous thromboembolism. Time-dependent Cox models detected no association between change in haematocrit and MACE (HR 0·97, 95% CI 0·92–1·02) or VTE (HR 0·94, 0·84–1·05) in testosterone-treated men with anaemia.

⚠⚠ THIS REVIEW THEREFORE WITHDRAWS THE OBJECTION IT WOULD MOST HAVE LIKED TO MAKE, AND ONLY THAT OBJECTION. A test was specified, run and reported null in the paper the panel cites, so the charge that the panel called confirmed a signal nobody had tested does not stand. ⚠⚠ WHAT THE WITHDRAWAL DOES NOT CONCEDE: a non-significant result does not confirm the absence of risk, and this review does not endorse the word confirmed for what these estimates show. The reported intervals — 0·92 to 1·02 for major adverse cardiac events and 0·84 to 1·05 for venous thromboembolism — are compatible with a small reduction, with no effect, and with a small increase. ⚠ The haematocrit change was not randomised; this is an observational analysis of a post-randomisation variable, and its interpretation depends on the model’s exposure unit, analytic population and handling of repeated measurements (correction 490). The withdrawal runs against this review’s own position and it is stated here rather than in a footnote.

What does not survive the read is narrower, and it is two things.

(i) The null is a subgroup null, and the panel generalises it. The analysis was performed in testosterone-treated men with anaemia — 390 randomised to testosterone with anaemia at baseline, 349 in the primary analysis — whose mean baseline haematocrit was 36·4%, against 42·9% in the non-anaemic participants and 41·9% across the full analysis set. ⚠ The sub-study does not report how many venous thromboembolic events occurred in that subgroup. A null for change-in-haematocrit among anaemic men rising from 36% is not a null for men approaching 54% — and the panel’s sentence says among participants.

(ii) The exposure was removed at the top of the range, in both papers. The parent trial excluded thrombophilia and uncontrolled heart failure. The sub-study excluded men with a contraindication to testosterone treatment, naming erythrocytosis first among its examples. And haematocrit above 54% triggered central automated dose reduction and then discontinuation at the lowest dose — ⚠ six participants in the anaemia study had study medication discontinued on exactly that criterion. The population in which the question is live was excluded at entry, and those who reached it during the trial were withdrawn from the exposure.

⚠⚠ AND THE PANEL’S ACCOUNT OF THE EXCLUSIONS DOES NOT LET A READER SEE THIS. It gives recent cardiovascular events, uncontrolled hypertension, prostate cancer, or elevated PSA levels. The trial excluded uncontrolled HEART FAILURE, not uncontrolled hypertension — and thrombophilia is absent from the panel’s list entirely, the exclusion the trial’s own Discussion invokes when it records that most reported cases of testosterone-associated thrombosis have been in men with underlying thrombophilia. ⚠ This limb is untouched by the Pencina read and it is now the subsection’s principal charge.

The main trial’s venous rows are recorded here as context and NOT as a charge against the panel. Pulmonary embolism 24 (0·9%) against 12 (0·5%); venous thromboembolic events 44 (1·7%) against 30 (1·2%), HR 1·46 (0·92–2·32). ⚠⚠ THAT COMPOSITE INTERVAL CROSSES ONE, SO no significant rise IS LITERALLY DEFENSIBLE FOR THE COMPOSITE — stated first, and as a defence rather than as a concession. ⚠ No hazard ratio was prespecified for the individual venous components; the trial states its intervals are unadjusted for multiplicity and must not be used to infer definitive treatment effects for secondary endpoints; and FM1 binds those absolute counts to any fold-change language. The panel is not answerable for that row, because it is not the row it cited.

✅ WHAT THE PANEL GETS RIGHT, RECORDED WITH EQUAL WEIGHT:

• The arrhythmia limb is accurate — a minor but significant increase in atrial fibrillation and non-fatal arrhythmias. The figures verify: nonfatal arrhythmia warranting intervention 134 (5·2%) against 87 (3·3%), p = 0·001; atrial fibrillation 91 (3·5%) against 63 (2·4%), p = 0·02. The trial’s own Discussion records that these events were not expected.
• Design and population verify exactly — 5246 men, 45 to 80, mean age 63·3, 47% aged over 65, body-mass index 35, approximately 70% with type 2 diabetes, gel titrated to 350–750 ng/dL. ⚠ Hypertension and dyslipidaemia are rounded to 90% from 93·2/92·3% and 90·1/89·6%; the rounding runs against the panel’s own interest.
• Its transmission of [53] is faithful.
• It states its own limitations unprompted: an older, high-risk population; the exclusion of men at low cardiovascular risk; that arrhythmia and heart-failure endpoints were underpowered for definitive conclusions.
• It reports a separate pooled meta-analysis giving non-significant arrhythmia and atrial-fibrillation estimates alongside the trial’s significant ones, rather than substituting one for the other.

⚠ Three residual imprecisions, recorded and not built on. The panel gives approximately 22 months as the limit of the trial’s reach; 21·7 ± 14·1 months is the mean TREATMENT duration and the mean FOLLOW-UP was 33·0 ± 12·1 months. The panel omits acute kidney injury — 60 (2·3%) against 40 (1·5%), p = 0·04 — which the trial reports in the same breath as the arrhythmia excess in both abstract and discussion. And the panel supports its no-significant-thromboembolic-risk position with what it calls independent datasets, the named one being a registry whose first author is the statement’s own first author.

⚠ One further panel claim, recorded and not checkable here. The panel reports that COVID-19 occurrence was associated with substantially increased risk of major adverse cardiac events, venous thromboembolism, atrial fibrillation and acute kidney injury across the whole cohort, with the hazard ratios for venous thromboembolism and acute kidney injury particularly high in infected participants and the relation of such events to testosterone therapy thereby no longer visible. The paper it attributes this to is NOT HELD — verified missing against the archive in the step this sentence was written. What is established is that the panel so reports.

⚠ What the panel RECOMMENDS is documentary and is carried once, at §9.4.4(d). It is not restated here. This subsection’s business with the panel is confined to whether its central citation is sound and what the primaries it points at actually report.

(e) The genetic evidence, and the two things it cannot reach

[43] is a two-sample multivariable Mendelian randomisation study using Bayesian model averaging to choose among 12 red-cell traits. Instruments were drawn from a genome-wide association study of 173 480 participants of European ancestry and applied to 265 424 unrelated white British UK Biobank participants, mean age 56·9, with 9752 cases of venous thromboembolism (4490 men and 5262 women). ⚠ Prevalent and incident cases were combined, under the paper’s own stated assumption that all events occur incident to the genetic exposure.

Haemoglobin ranks first and haematocrit does not. After exclusion of seven outlying variants (n = 641): haemoglobin marginal inclusion probability 0·912, model-averaged odds ratio 1·22; haematocrit 0·275, model-averaged odds ratio 0·95. The best-fitting model contains haemoglobin alone, posterior probability 0·461; models containing both haemoglobin and haematocrit have low probability. [43] states that the genetic associations with the two traits correlate at r = 0·91.

In univariable Mendelian randomisation, 81 variants: inverse variance weighted OR 1·21 (1·05–1·41) per g/dL of haemoglobin, p = 0·01; MR-PRESSO 1·21 (1·06–1·38); MR-Egger 1·39 (1·04–1·87), intercept p = 0·30. After excluding variants with known pleiotropy, 72 variants: 1·20 (1·05–1·37). FM1’s base rate: 9752 cases in 265 424 participants.

⚠ ONE GLOSS IS STATED RATHER THAN SMOOTHED. [43]’s Abstract describes the sensitivity analyses as giving consistent estimates. The weighted median at 81 variants is 1·18 (0·99–1·41) and its interval crosses one. The Discussion’s own phrase — directionally consistent — is the accurate one, and if the shorter phrase is used the weighted median goes beside it.

⚠ AND THE MODEL-AVERAGED FIGURES ARE DISCLAIMED BY THEIR OWN AUTHORS. [43] states that MR-BMA is a variable-selection method and, as is common for such methods, does not provide unbiased effect estimates; that its estimates are shrunk toward the null; and that unbiased estimation is supplied instead by the univariable analyses. The marginal inclusion probabilities rank traits. They do not size effects, and this review does not use them as though they did.

⚠⚠ TWO OF [43]’s OWN LIMITATIONS CUT AGAINST THE READING THIS REVIEW WOULD MOST WANT TO TAKE FROM IT, AND BOTH ARE THE PAPER’S, NOT THIS REVIEW’S.

First: [43] states that a more accurately measured trait would inevitably be prioritised over a less accurately measured one, and that it cannot exclude the possibility that haemoglobin is easier to measure accurately than haematocrit. Haemoglobin’s win over haematocrit may therefore be a measurement artefact by the authors’ own admission, and no use of this paper to argue that haemoglobin rather than haematocrit is the trait that matters may travel without that sentence attached.

Second: [43] records that the use of summary statistics precluded examination of non-linear associations, and that examination of threshold effects would be clinically relevant where possible. ⚠⚠ Threshold effects are §9.3’s entire subject, and [43] is silent on them by design. This paper cannot be used to say anything about 50%, 52% or 54%.

⚠ [43]‘s testosterone sentence is NOT rendered here as a finding.** Its Discussion states that testosterone induces erythrocytosis and substantially increases haemoglobin, and that its findings may provide a potential mechanism by which testosterone could increase the risk of venous thromboembolism, citing the authors’ own earlier work. That is hypothesis-generating, sited in a Discussion, and secondary under Amendment C even inside a primary paper.** It is recorded as what the authors suggest and nothing more.

⚠ A cross-source observation this subsection is the first place to carry, and it runs in both directions at once. [43] restricted its analysis to genetically verified white British participants expressly to reduce confounding by a hereditary tendency to thrombophilia. [52] excluded men with thrombophilia outright. ⚠⚠ THOSE ARE NOT THE SAME LIMITATION AND THIS REVIEW DOES NOT MERGE THEM. The trial excluded people with clinically recognised thrombophilia, so it is silent about them by design. ⚠ The genetic analysis has a different limitation: it restricted ancestry and genetic-data quality, partly to reduce confounding from inherited thrombophilic tendency, which does not remove participants carrying thrombophilic variants; it does not reproduce a therapeutic testosterone exposure; and it supplies no clinically defined thrombophilia subgroup. ⚠⚠ An earlier version of this subsection treated the two limitations as the same. They are not the same exclusion and should not be summed. Neither design is a defect. The consequence is that the two strands are silent for DIFFERENT reasons — and the population at issue is the one [52]’s own Discussion identifies as carrying most reported testosterone-associated thrombosis.

[43]’s other declared limits, recorded so they are not re-derived: no sex-specific genome-wide association study, so no sex-specific associations were assessed; approximately 30% sample overlap with an estimated bias of 0·004; selection bias in both contributing cohorts biasing toward the null; 23% of the instrument sample were blood donors with no adjustment for donation frequency; and small lifelong genetic differences are not the same intervention as a large therapeutic change over months.

(f) What this subsection establishes, and what it does not

Established, on first-hand reads of the primaries:

1. Testosterone raises haematocrit and haemoglobin in randomised trials, by 3·15 percentage points (2·42–3·88) and 10·87 g/L (8·19–13·55) respectively at pooled analysis.
2. It raises the frequency of high haematocrit roughly five-fold, 30/1079 (2·8%) against 5/993 (0·5%).
3. ⚠⚠ The pooled randomised trials DID NOT DETECT AN EXCESS of venous thrombotic events despite that rise — 5/1037 on testosterone against 7/1034 on placebo, twelve events in total across four trials, at a mean 9·5 months, in hypogonadal men. ⚠ An earlier version headlined this as reports no increase, which reads as a negative finding. Twelve events have little power to settle the direction, and the formulation is corrected throughout under the same principle correction 27 applies elsewhere.
4. The largest single trial reports a higher incidence of pulmonary embolism, 24 (0·9%) against 12 (0·5%), and did not test it, having prespecified no hazard ratio for the venous components and disclaimed its own unadjusted intervals.
5. The strongest expert reassurance available is correctly sourced and rests on a real null — but that null was estimated in anaemic men rising from a mean baseline haematocrit of 36·4%, no event count is given for the subgroup, and both papers excluded erythrocytosis and thrombophilia at entry and withdrew those who exceeded 54%. ⚠ The panel’s account of the exclusions does not disclose this, and that omission is the surviving charge.
6. ⚠⚠ The interventional evidence base CANNOT address sustained haematocrit above 54%, because the trials that reach that value curtail the exposure by protocol. This is the subsection’s central finding and it is a finding about what the evidence can say, not about what is true. ⚠⚠ A SECOND RANDOMISED TRIAL NOW CARRIES THE SAME STRUCTURE, AND IT WAS DESIGNED AROUND THE THRESHOLD RATHER THAN MERELY ENCOUNTERING IT. In a two-year placebo-controlled trial of testosterone undecanoate in 1007 men aged 50 to 74, with a waist circumference of 95 cm or more and expressly WITHOUT pathological hypogonadism, haematocrit of 54% or higher was a prespecified centrally monitored safety trigger; it occurred in 106 of 491 treated participants against 6 of 484 on placebo, and treatment was permanently ceased in 26, of whom 25 were in the treated arm. ⚠ Sustained haematocrit above 54% was therefore curtailed in that trial too, by design. The limitation is not an accident of one trial’s monitoring rules. It is what happens when a value is treated as an action level: the exposure of interest is removed at the moment it arises. ⚠⚠ Direction: this runs FOR this review, and it is stated as a fact about what these designs can observe and NOT as evidence that sustained exposure is harmful, harmless, or anything at all.

NOT established, and recorded here so no later record fills it by inference:

• That the raised haematocrit causes the events. Nothing here demonstrates it.
• That it does not. [53]’s dissociation is the strongest evidence available on that side and rests on twelve events over nine and a half months.
• Anything about any threshold. [43] is silent by design; [52] intervened at 54% rather than observing it; [53] reports a dichotomised high haematocrit count without stating the operator.
• Anything at all about trans-masculine people. ⚠ NONE of these treatment trials was conducted in a trans-masculine gender-affirming-hormone cohort, and none identifies or analyses a trans-masculine subgroup. That limit is flagged wherever a figure from them is used. ⚠⚠ The trials record participants as MEN. They do not report ascertainment of gender identity, and this review does not supply one — correction 44, and now correction 195.

Provenance

Gate: one batch, 13 August 2026 UTC, gate_passes: true on all eight held DOIs; one deliberate absence check returning held: 0.
Reads, all truncated: false, all verified before any content claim: TRAVERSE id 139; Hudson id 147; Zitzmann id 144; Luo id 134, the last with Tables S1–S8 and Figures S1–S6 reached in the one call.
Added in the correcting step, 15–17 August 2026: Pencina id 140 (the TRAVERSE Anemia Study), Zitzmann id 144 re-opened, TRAVERSE id 139 re-opened — all truncated: false. ⚠ Pencina has NO live reference number in §3.6 and one was NOT assigned; it resolves at assembly.

Nothing in this subsection was written from the register, from a prior sitting’s read, or from an earlier draft.

Grading block — §9.4.5, the interventional and genetic evidence

⚠⚠ The outcome genuinely divides, so this is a per-statement block under §6 of the Grading Token Standard. Compressing five statements into one would lose the subsection’s central finding.

1 — Testosterone raises haematocrit and haemoglobin in randomised trials

Causal attribution — ROBUST. Randomised evidence, pooled, consistent in direction.
Observed effect size — MODERATE. 3·15 percentage points (2·42–3·88) and 10·87 g/L (8·19–13·55) at pooled analysis. Precisely estimated in the trials; the trials are not the treated population.
Population directness — indirect. Hypogonadal men.

2 — It raises the frequency of high haematocrit roughly five-fold

Causal attribution — ROBUST. Randomised.
Observed effect size — MODERATE. 30 of 1,079 against 5 of 993 — 2·8% against 0·5%, an absolute difference of about 2·3 percentage points. The fold-change is reported with its base rates and is not reported without them.
Clinical importance — NOT ESTIMABLE. ⚠⚠ Crossing a laboratory threshold is a surrogate, not a patient outcome, and what follows from crossing it is the subject of every other statement here — none of which establishes it. ⚠ The pooled operator for high haematocrit also differs between the contributing trials. The absolute difference is reported — about 2·3 percentage points, roughly one additional person in forty-three — and it is not graded for clinical importance. ⚠ An earlier draft graded it MODEST; that mark is withdrawn.
Population directness — indirect.

3 — The pooled randomised trials did not detect an excess of thrombotic events despite that rise

⚠ The statement is worded as a failure to detect and not as a negative finding, and the wording is load-bearing. 5 of 1,037 on testosterone against 7 of 1,034 on placebo.
Causal identification of the effect of randomised testosterone assignment — MODERATE. ⚠ Randomisation identifies the effect of assignment, and it does so however many events accrue. What limits the mark is scope rather than sparseness: only four of the thirty-five pooled trials contributed venous-event data, mean treatment duration was 9·5 months, and eighteen trials supplied no individual participant data at all.
Observed thrombotic effect size — LIMITED. ⚠ The aggregate difference is calculable and extremely imprecise — twelve events in total across four trials — rather than non-estimable. ⚠⚠ An earlier version graded this NOT ESTIMABLE, which under §3’s own definition means the quantity cannot be extracted. It can; it is simply almost uninformative.
Population directness — indirect.

⚠⚠ THE CAUSAL MARK MOVED FROM LIMITED TO MODERATE AT v4.6, AND THE MOVE IS A CORRECTION RATHER THAN A CONCESSION. v4.5 downgraded causal attribution because there were twelve events. That is a precision fact and it belongs on the effect-size axis, which §3 defines as a separate question. ⚠ Sparse events do not make random allocation cease to identify treatment assignment causally. The direction of this correction runs AGAINST this review’s position, because it strengthens the causal standing of the strongest piece of evidence on the reassuring side, and it is recorded for that reason. ⚠ ROBUST was considered and refused: the venous-endpoint corpus is too sparse and too incompletely reported across the randomised programme for the design alone to carry the top mark.

4 — That the raised haematocrit causes the events

Causal attribution — NOT ESTABLISHED. Nothing in the interventional evidence demonstrates it. ⚠⚠ Nor does anything demonstrate the reverse, and the strongest evidence on that side rests on the same twelve events.

5 — Any threshold effect in the interventional and genetic evidence, at 54% or elsewhere

Effect estimate specific to 54% — NOT ESTIMABLE. Causal effect of reaching or sustaining 54% — NOT ESTABLISHED. ⚠⚠ This is the subsection’s central finding. The largest cardiovascular-outcome trial curtailed the exposure by protocol, intervening at 54% rather than observing sustained values above it. ⚠⚠ A SECOND RANDOMISED TRIAL, READ IN PRIMARY, DID THE SAME — in men without pathological hypogonadism, and it was NOT designed as a transgender or gender-affirming-hormone cohort, identifies no trans-masculine subgroup and analyses none, so it is INDIRECT for this review’s population: 54% was a prespecified safety trigger, 106 of 491 treated participants reached it, and treatment ceased permanently in 25 of them. ⚠ Two independent randomised designs now curtail the same exposure at the same value, which strengthens the grade’s REASON without moving the grade: neither trial can supply an effect estimate for an exposure its protocol removes. The genetic evidence is silent on thresholds by design. The remaining trial reports a dichotomised high haematocrit count without stating the operator. ⚠ An earlier draft graded this NOT REACHED; that mark is withdrawn, because the value was contained in the observed range and not resolved rather than never encountered.
Population directness — not applicable; no study resolves the exposure.

⚠ A charge the subsection makes and this block carries forward rather than grades. The strongest expert reassurance available rests on a real null, but that null was estimated in anaemic men rising from a mean baseline of 36·4%, no event count is given for the subgroup, and both papers excluded erythrocytosis and thrombophilia at entry and withdrew anyone exceeding 54%. The panel’s account of the exclusions does not disclose this. That is a documentary charge and it belongs at §9.4.4’s Documentary line, not in an axis here.

9.4.6 Three statements relocated from §10 into Part V

⚠⚠ AN EARLIER ASSEMBLY GRADED THESE IN THE GATHERED TABLE WITH NO GRADED SUBSECTION ANYWHERE ABOVE IT. §10 states that every mark is the mark its section assigned. For these that was not true, and the grades are relocated here so that it is.

1 — Erythrocytosis on testosterone and thrombotic events, records outside gender-affirming care

Association — LIMITED. ⚠ One large propensity-matched records study: 5·15% against 3·87%, odds ratio 1·35 (1·13–1·61) — an absolute difference of about 1·3 percentage points over one year.[20] ⚠ No reciprocal is reported: in a confounded observational contrast it invites a causal reading the design cannot support. It is the largest study bearing on this review’s question and it points against this review’s scepticism.
Causal attribution — NOT ESTABLISHED. The matched groups were balanced on the reported covariates, ⚠⚠ but baseline haematocrit was not among them and differed materially — 47·4 against 42·5. The exposure is a single unconfirmed record value, and classification requires a test to have been performed. ⚠ Its authors concede they cannot distinguish risk from reaching the threshold from risk in men who started higher.[20]
Observed absolute difference — 1·28 percentage points over one year for the composite outcome. ⚠ No qualitative magnitude mark is assigned: §3 lists what clinical importance may be assessed by, but publishes no anchor separating limited from moderate on an absolute risk difference, and this review does not invent one here. ⚠⚠ The reciprocal is not reported, because in a confounded observational contrast it invites a causal reading the design cannot support.
Clinical importance attributable to haematocrit — NOT ESTABLISHED. ⚠⚠ Residual confounding and the unmatched baseline haematocrit prevent attributing that difference to crossing the threshold. An earlier version graded the axis NOT ESTIMABLE after calculating the very quantities §3 defines it by — which quietly made clinical importance mean causally attributable clinical importance, a meaning §3 does not give it.
Population directness — indirect. Men, mean age 53.

2 — Testosterone and direct haemostatic measures in trans-masculine people

⚠⚠ THIS STATEMENT IS RENAMED AT v4.7. IT WAS PLASMA-BASED HAEMOSTATIC MEASURES, AND THAT LABEL IS FALSE OF ITS OWN CORPUS, which contains platelet-activation markers measured in serum, conventional coagulation factors and inhibitors in platelet-poor plasma, global thrombin generation in platelet-poor plasma, fibrin-clot properties, and inflammatory markers. Specimen matrix and biological pathway differ across it, and the corpus label must not assert otherwise.

⚠⚠ WHAT THIS EVIDENCE BASE IS, STATED BEFORE ANY GRADE: AN OVERLAPPING AMSTERDAM/GHENT ASSAY PROGRAMME, HETEROGENEOUS BY ASSAY FAMILY, AND NOT A SET OF INDEPENDENT REPLICATIONS. Almost all of the direct trans-masculine haemostasis data identified here come from one clinical lineage — the Vrije Universiteit / Amsterdam UMC gender clinic and its ENIGI successor, with Ghent — analysed repeatedly by different assay families across more than twenty years. Counting the participant totals of those analyses as though they were separate studies would overstate the evidence base severalfold, and this review does not do it.

⚠ OVERLAP MAP, BUILT FROM WHAT EACH PAPER STATES ABOUT ITS OWN RECRUITMENT. Blank cells are not established and are not guessed.

Analysis	Centre(s)	Sampling window	Trans men	Treatment	Assay family	Overlap
Toorians 2003[48]	VU Amsterdam	before 2003	14	T esters 250 mg/2 wk IM	APC resistance, protein C/S, prothrombin (platelet-poor plasma)	predates ENIGI; same clinic lineage
Defreyne / ENIGI[22]	Amsterdam + Ghent	ENIGI era	192	undecanoate, esters, gel	haematocrit only	source population for the rows below
Scheres 2021[27]	ENIGI Amsterdam only	August 2012 – September 2015	100	59% intramuscular (Sustanon 250 mg/2 wk or Nebido 1000 mg), 41% gel 50 mg/d	coagulation factors and inhibitors, APC ratio, haematocrit (platelet-poor plasma)	⚠⚠ window sits inside Bøgehave’s and overlaps Schutte’s; Amsterdam-only, so overlap with the Amsterdam arm of every row here is LIKELY and is not quantified
van Zijverden 2025[56]	Amsterdam UMC VU, nested in the SHAMVA study	baseline, 6, 12 and 52 weeks	18	gel 40·5 mg/d PLUS triptorelin for 12 weeks, then 7 switched to injections	7 platelet-activation markers (whole blood flow cytometry) and 12 inflammation markers	⚠⚠ shares four authors with the rows above and draws on the same centre; overlap not quantified. Its first 12 weeks are a DIFFERENT EXPOSURE from every other row: testosterone with gonadotrophin-releasing-hormone-agonist co-suppression
Schutte 2022[57]	ENIGI Amsterdam	June 2012 – July 2019	47	gel 50 mg/d only	platelet activation, coagulation, inflammation	same network and overlapping window
Bøgehave 2025[26]	Amsterdam 223 + Ghent 125	April 2012 – March 2020	348	7 modalities	global thrombin generation (platelet-poor plasma)	window overlaps Schutte’s entirely
Bøgehave 2026[58]	as above	as above	320	as above	fibrin formation and clot lysis	nested inside the 348

⚠⚠ THE MAP IS THE POINT AND THE BLANKS ARE PART OF IT. Two of these analyses draw on windows that overlap entirely, one is nested inside another, and one cannot be placed at all because it has not been obtained. No participant-level overlap has been reconstructed, and until it is, no study count over this corpus means what a study count normally means.

⚠⚠ THE 2021 ANALYSIS IS NOW FILED AND READ FIRST-HAND, AND ITS CONTENT IS INTEGRATED BELOW. Scheres, Selier, Nota, van Diemen, Cannegieter and den Heijer, J Thromb Haemost 2021;19(4):1029–37, 10.1111/jth.15256, gate-checked held: true and opened untruncated in the step this passage was written. ⚠ It entered this review because an external reviewer named it, not because this review’s retrieval found it, and that remains a fact about the retrieval.

⚠⚠ THE WHOLE-BLOOD PLATELET ANALYSIS IS NOW FILED AND READ FIRST-HAND, AND ITS RESULTS ARE INTEGRATED AT FAMILY (iii) BELOW. van Zijverden, Schutte, Tebbens, Madsen, van Diemen, Wiepjes, den Heijer and Thijs, Platelets 2025;36(1):2567294, 10.1080/09537104.2025.2567294, gate-checked held: true and opened untruncated in the step this passage was written.

⚠⚠ THE PROVENANCE STATE, DEFERRED TO THE GENERATED TABLE RATHER THAN RESTATED HERE. All direct haemostasis analyses identified BEFORE the repaired search were obtained and read first-hand. The repaired search subsequently identified van Kesteren 1998, which has NOT been obtained by this review. ⚠⚠ The search has been PARTIALLY EXECUTED and not completed to protocol, so neither the identified corpus nor its provenance should be read as field-complete.

⚠ Source-entry provenance for every analysis in this corpus is reported in the generated disposition table at Evidence identification, and is not summarised again here. ⚠⚠ An earlier version of this paragraph asserted that no identified analysis remained unobtained, spoke prospectively of a search that had already been partially run, and summarised provenance as every source except the two oldest. All three were falsified by the search this review itself executed. Correction 131.

⚠⚠ AND THE PROVENANCE CLASS IS CORRECTED HERE, BECAUSE AN EARLIER VERSION CONFLATED TWO THINGS. That a paper is absent from this review’s archive is a limitation of THIS REVIEW’S RETRIEVAL SYSTEM. It is not a statement that the paper is publicly inaccessible, and v4.6’s verified missing wording invited exactly that reading. ⚠ Both papers above are ordinary journal articles that any reader can obtain. The obstacle is this review’s filing, not the literature’s availability, and the distinction is now stated wherever an unobtained source appears.

⚠ The secondary characterisations this review previously carried of the 2021 analysis are now SUPERSEDED BY THE PRIMARY and are recorded rather than deleted: one paper described it as observing no apparent changes[57]; another attributed to it, jointly with the 2003 study, increased Factor IX, decreased Factor II and Factor XI, increased protein S and a non-significant decrease in protein C.[26] ⚠⚠ Both characterisations are ACCURATE against the primary, which is worth recording in a review whose §9.6 catalogues the cases where they are not. ⚠⚠ Nothing is established about the 2025 whole-blood analysis, and its mere existence is why no global directional claim is made below.

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THE EVIDENCE, BY ASSAY FAMILY. IT IS HETEROGENEOUS AND IS NO LONGER SUMMARISED IN ONE DIRECTION.

(i) Global coagulation potential — plasma. In 348 trans men, endogenous thrombin potential or peak thrombin generation fell after six of seven testosterone modalities and the lag time lengthened for six; the authors conclude that masculinising therapy was anticoagulant.[26] The nested 320-participant analysis found fibrin formation reduced and clot lysis increased on every modality.[58] ⚠ These are one dataset analysed twice.

(ii) Conventional coagulation factors and inhibitors — plasma. In 14 trans men given testosterone esters for four months, the normalised activated-protein-C sensitivity ratio fell from 2.0 ± 0.8 to 1.3 ± 0.7 (p < 0.001); total protein S rose from 105 ± 22 to 118 ± 19% (p = 0.037) and free protein S from 83 ± 20 to 97 ± 15% (p = 0.012); protein C did not change significantly (110 ± 18 to 100 ± 15%, p = 0.070) and prothrombin did not move (106 ± 18 to 107 ± 16%, p = 0.826). Its authors conclude that testosterone administration had a mild antithrombotic effect, and record that no participant developed clinical signs of venous thrombosis during the four-month study period.[48] ⚠ Fourteen participants, four months, one centre, no control group, one factor V Leiden carrier among them, and it is the oldest source in this corpus.

And the same assay family, twenty years later and seven times larger, in 100 trans men before and 12 months after starting testosterone. ⚠⚠ The changes run in BOTH directions and the paper says so. Anticoagulant: the activated-protein-C ratio rose 2.84 → 3.32, +0.48 (95% CI 0.34–0.61); Factor II fell −5.7 IU/dL (−9.9 to −1.5); Factor XI fell −7.7 IU/dL (−11.8 to −3.7); free protein S rose +7.6 IU/dL (1.6–13.6). Procoagulant: Factor IX rose +7.9 IU/dL (2.2–13.6), and haematocrit rose +0.06 L/L (0.05–0.07), a 14.6% relative increase. Fibrinogen and protein C did not apparently change.[27]

⚠⚠ ITS AUTHORS’ OWN READING IS THE ONE THIS REVIEW ADOPTS, AND IT CUTS BOTH WAYS. They conclude that testosterone use in trans men did not result in apparent procoagulant changes, which they say provides some reassurance regarding venous thromboembolism risk — and in the same discussion they state that the only potential procoagulant change was the rise in haematocrit, adding that the causal role of haematocrit in venous thromboembolism is currently debated, citing the same 2010 editorial §9.4.3 carries.[27] ⚠⚠ That is this review’s own question, named as the residual by the largest conventional-coagulation study in the population, and it is recorded at full weight.

⚠ Its limits are its authors’ own: a selection of parameters chosen for their known behaviour under oestrogen, so other changes in vivo were not measured; no statement of how the measured changes sum to a net procoagulant or anticoagulant state; and no adjustment for concurrent changes in body-mass index or smoking during transition.[27] ⚠ Each participant is their own control and there is no untreated comparator.

(iii) Platelet activation and inflammation — serum and plasma. In 47 trans men on transdermal gel, no statistically significant changes were detected in the measured platelet-activation and coagulation markers at twelve months — ⚠ the estimates and intervals below do not establish equivalence or exclude potentially relevant changes — platelet factor 4 +0% (95% CI −18; 23), β-thromboglobulin +2% (−16; 23), fibrinogen −8% (−23; 9), with no statistically significant changes detected in P-selectin or PAI-1.[57] ⚠⚠ But its inflammatory limb runs the other way and is reported at full weight: high-sensitivity C-reactive protein, which its authors describe as directly associated with cardiovascular risk, ROSE by 71% (95% CI 19; 145) over twelve months, while no statistically significant changes were detected in the other systemic inflammatory markers.[57] ⚠ That analysis is expressly exploratory and hypothesis-generating, tested many markers with no adjustment for multiple comparisons, has no untreated control because each participant serves as their own, and followed young healthy people for one year.[57]

⚠⚠ AND THE WHOLE-BLOOD ANALYSIS, WHICH IS THE ONE MEASUREMENT IN THIS CORPUS MADE UNDER CONDITIONS DESIGNED TO AVOID ARTEFACTUAL PLATELET ACTIVATION, POINTS THE OTHER WAY. In 18 trans men sampled at baseline and 6, 12 and 52 weeks, the authors conclude that testosterone may increase platelet activation and inflammation, and read the overall pattern as a stimulating effect on primary haemostasis.[56]

⚠⚠ THIS REVIEW REPORTS THE MARKER-LEVEL PATTERN AND NOT ONLY THE AUTHORS’ READING, BECAUSE THE TWO ARE NOT THE SAME STRENGTH. Of seven platelet-activation markers at 52 weeks, ONE reached significance: platelet–leukocyte complexes, +68% (95% CI 8; 162). CD62p rose 24% (−40; 153) and CD63 3% (−38; 71), both intervals spanning zero and both very wide; plasma thromboxane B2 fell 7% (−40; 45); and closure time by two methods and the immature platelet fraction were stable. Of twelve inflammation markers at 52 weeks, THREE reached significance: high-sensitivity C-reactive protein +68% (23; 128), interferon gamma +109% (18; 270) and tumour necrosis factor alpha +26% (9; 45).[56]

⚠⚠ THE STUDY’S OWN LIMITS ARE SEVERE AND ARE ITS AUTHORS’. It is expressly exploratory, was an add-on to a larger study, performed NO sample-size calculation, and applied NO correction for multiple testing across nineteen markers. Eighteen participants, aged 22 at median, healthy by design — cardiovascular, inflammatory and metabolic comorbidity, antiplatelet drugs, NSAIDs, anticoagulants and SSRIs all excluded — and the authors state they cannot extrapolate beyond young healthy people and had insufficient power for any sensitivity analysis.[56] ⚠⚠ AND ITS EXPOSURE IS NOT THE ONE THE REST OF THIS CORPUS STUDIES: for the first twelve weeks every participant received testosterone gel TOGETHER WITH a gonadotrophin-releasing-hormone agonist, so the early timepoints are testosterone with ovarian suppression rather than testosterone alone.[56]

⚠ It measures no haematocrit, so it bears on this review’s exposure only through the platelet pathway and not directly.

⚠⚠ AND THE TWO PLATELET ANALYSES IN THIS CORPUS DISAGREE, WITH THE LATER ONE CHALLENGING THE EARLIER ON METHOD. The whole-blood authors note that the earlier analysis found no clear change, and observe that it assessed only three platelet-activation markers and used neither whole blood nor indomethacin-treated plasma, so subtle in vivo changes could have been missed.[56][57] ⚠ This review does not adjudicate that. It records that the disagreement is between two papers from the same centre sharing authors, and that the one reporting a signal is the smaller and the more exploratory of the two.

(iv) Smaller separate work and pooled analysis. Factor VII down 15% with the international normalised ratio slightly prolonged,[24] and a pooled meta-analysis detecting no statistically significant change in fibrinogen, aPTT or PAI-1.[23]

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Association, global coagulation and fibrinolysis — MODERATE. Consistent neutral-to-anticoagulant direction across the largest analyses, which are one overlapping dataset. ⚠⚠ THAT DIRECTION IS A PROPERTY OF THE MEASURED ASSAYS AND NOT A DEMONSTRATED CLINICAL EFFECT. These are plasma-based measurements. Thrombin generation, fibrin structure and lysis are laboratory quantities whose mapping onto clinical thrombosis is not established here, and plasma assays do not directly assess red-cell-dependent rheology, cellular margination or intact whole-blood interactions — they may still reflect plasma changes accompanying treatment, and their sensitivity to a haematocrit-mediated effect has not been analysed here (correction 546) — which is the whole subject of this review. ⚠ Nor do the fibrin features all move in one direction, so anticoagulant is a summary of a mixed profile rather than a reading of every measure. Where a ratio is reported, its definition and polarity govern its sign: numerically opposite changes in differently defined activated-protein-C measures can express the same biological direction (correction 530).
Association, platelet activation and inflammation — LIMITED, and the evidence is DIVIDED rather than thin. ⚠⚠ Two analyses from the same centre point different ways, and that is not the same as a demonstrated disagreement. ⚠ A significant result in one small study and a non-significant result in another is not a tested difference between them; the studies use different specimens, assays and processing, and no between-study comparison has been performed (correction 525). One, in 47 trans men on gel, finds no statistically clear change detected in the measured platelet-activation and coagulation markers at twelve months — which does not establish equivalence, and does not exclude changes compatible with the reported intervals (correction 538) while high-sensitivity C-reactive protein rises 71%.[57] The other, in 18 trans men measured in whole blood, concludes that testosterone may increase platelet activation and inflammation — on one of seven platelet markers reaching significance and three of twelve inflammation markers, with no correction for multiple testing and no power calculation.[56] ⚠⚠ What the two agree on is the inflammatory limb: high-sensitivity C-reactive protein rises in both, by 71% and by 68%. ⚠ No direction is asserted for platelet ACTIVATION. A rise in an inflammatory marker is the one directionally consistent finding in this family, and it is not a haemostatic one.
Within-person treatment-associated change — MODERATE for the global coagulation family; LIMITED for the platelet and inflammatory family.
Causal attribution specifically to testosterone — LIMITED. ⚠⚠ No analysis in this corpus has an untreated contemporaneous control, so none removes secular change, repeated-testing effects, other treatment-associated changes such as the weight and lipid changes the same programme reports, or time-varying confounding. ⚠ An earlier version graded this MODERATE on prospective design and directional consistency across assays; consistency across assays run on overlapping participants is not independent replication and does not substitute for a comparator.
Clinical thrombotic consequence — NOT ESTABLISHED. ⚠⚠ NONE OF THESE ASSAYS IDENTIFIES THE EFFECT OF THE TESTOSTERONE-ASSOCIATED HAEMATOCRIT RISE, AND THE REASON IS NOT THE ONE THIS REVIEW GAVE AT v4.6. The thrombin-generation, fibrin and conventional-factor analyses are performed on platelet-poor plasma and therefore exclude red cells from the assay itself;[26][48] others measure platelet-related markers in serum or plasma, and the van Zijverden 2025 whole-blood flow-cytometry analysis is now read first-hand and integrated above. ⚠⚠ The objection that survives across all of them is different and stronger: none randomises or otherwise isolates haematocrit from the concurrent hormonal, metabolic and haemostatic changes testosterone produces, and none links any assay change to an adjudicated thrombotic event. ⚠ v4.6 said the red cells are centrifuged out of these assays. That was true of some and false of others, and it is corrected here.
Population directness — direct. ⚠ Participant-level overlap within this corpus, and with the 192-person ENIGI haematocrit sample, has not been reconstructed.

3 — Change in haematocrit after starting testosterone, and subsequent major adverse cardiovascular events, records outside gender-affirming care — ⚠ abstract-level; full text not appraised

Association — LIMITED. ⚠ One large propensity-matched retrospective analysis reports greater subsequent major adverse cardiovascular events among men whose haematocrit rose after starting testosterone than among matched men whose haematocrit remained stable — 10,511 in each arm.[49]
Causal attribution — NOT ESTABLISHED. Retrospective records, no randomisation of the haematocrit change, and no mediation analysis.
Observed effect size — not graded here. ⚠⚠ The full article was not obtained and the published abstract does not supply the complete effect estimates this review requires. This is an abstract-level read, and the axis is left ungraded rather than filled from a summary.
Population directness — indirect. Men.

⚠⚠ This is the closest published study to the exposure this review argues for — change from an individual’s own baseline rather than crossing a fixed line. ⚠ It is NOT independent of statement 1 above: the same records network over overlapping periods, and three shared authors including that paper’s first author. It is not a replication, and the two must not be counted as two signals.

4 — Drug-induced erythrocytosis and thrombotic events, non-trans populations

⚠⚠ THIS STATEMENT IS NEW AT v4.18 AND IT IS A PARTICULARLY CLOSE INDIRECT ANALOGUE IDENTIFIED BY THE REPAIRED SEARCH: A DRUG IS GIVEN, IT RAISES HAEMATOCRIT, AND THROMBOTIC OUTCOMES ARE COUNTED. ⚠ Pharmacologically induced erythrocytosis with counted thrombotic outcomes is a closer analogue than spontaneous haematocrit variation. An earlier version called it the nearest analogue THAT ANY LITERATURE PROVIDES, which an incomplete search cannot establish. Correction 118. The sodium-glucose cotransporter 2 inhibitors produce erythrocytosis by a different mechanism from testosterone and in a different population, so nothing transfers directly. ⚠ What transfers is the SHAPE of the question: that literature has EXAMINED the structural question in ways the trans-masculine corpus has not. ⚠⚠ It has not answered it — the grades below record that marker and mediator are not distinguished — and an earlier version said answered, whose referent was ambiguous between the structural question and the causal one. Correction 126.

The randomised evidence. A post hoc pooled analysis of patient-level data from two randomised, double-blind, placebo-controlled trials, with baseline haematocrit available for 14,321 of 14,543 participants, median follow-up 2.4 years.[59] Canagliflozin raised haematocrit regardless of baseline category in both sexes, including in those already erythrocytotic, and raised the proportion erythrocytotic at one year from 5.5% to 16.9% in males (OR 3.51, 2.97–4.15) and 1.0% to 5.2% in females (OR 5.48, 3.36–8.95). The effect on the composite of myocardial infarction, stroke and any thromboembolism showed no statistically significant overall association — males 0.97 (0.86–1.10), females 0.95 (0.78–1.15). In males, baseline haematocrit modified the treatment effect, P interaction <0.01 categorical and 0.01 continuous, with a three-way treatment-by-haematocrit-by-sex interaction at P = 0.02. No modification in females.

⚠⚠ AND THE STRATUM ESTIMATES THEMSELVES ARE NULL, WHICH THE PAPER’S TITLE DOES NOT CONVEY. In erythrocytotic males the composite estimate is 1.08 (0.68–1.69) and the myocardial-infarction estimate is 1.77 (0.76–4.13); both cross unity. The significant interaction is carried substantially by BENEFIT at the anaemic end — 0.64 (0.49–0.85) composite and 0.53 (0.34–0.83) for infarction. The authors state that no adjustment was made for multiplicity, that the analysis is exploratory in trials not designed to detect treatment heterogeneity, and that the limited number of patients and events among those with erythrocytosis raises concerns about statistical power.[59] ⚠ §9.6 carries the transmission consequence.

The observational evidence, and it converges. In 6,787 patients prescribed a sodium-glucose cotransporter 2 inhibitor at a single centre over ten years, erythrocytosis occurred in 1,145 (16.9%) — 13.1% excluding those erythrocytotic at baseline — with a median haematocrit rise of 3.5 points overall and 4.7 in those affected.[60] Thrombosis occurred in 33 (0.5%). Multivariable predictors were antiplatelet use 3.57 (1.60–7.97), anticoagulant use 5.93 (2.60–13.57), and baseline erythrocytosis 3.75 (1.41–9.96).

⚠⚠ BUT WITHIN THE ERYTHROCYTOTIC GROUP THERE IS NO GRADIENT WITH THE DEGREE OF ELEVATION. Stratified by peak haemoglobin tertile, thrombosis occurred in 1.4%, 2.2% and 2.0%, P = 0.649.[60] ⚠⚠ And of the 33 thrombotic events, only FIVE occurred in someone erythrocytotic at the time; all five were arterial, and all five had atrial fibrillation, severe coronary calcification or large-artery atherosclerosis. The authors conclude that thrombotic risk appears primarily driven by pre-existing conditions, and name confounding by indication for the antiplatelet and anticoagulant associations.[60]

⚠ A third line, at conference-abstract provenance only, and it is a RECURRENCE analysis in an event-selected population rather than a study of incident thrombosis. Among 3,931 people presenting with a thromboembolic event in one Swedish county, 12.7% had elevated haemoglobin or haematocrit at some point but only 4% had it at the time of the initial event; and in the recurrence analysis, those with UNEXPLAINED erythrocytosis had better recurrence-free survival than those in the same cohort with no elevation — 3.72 years (3.55–3.89) against 3.27 (3.20–3.30) ⚠ which is a within-cohort comparison among people who had all already had an event, not a comparison with a general population — while secondary erythrocytosis, meaning a cause was found, had the worst at 2.30 years (1.95–2.64).[61] ⚠⚠ This is an EHA congress abstract, not a full paper: no multivariable adjustment is reported, and the groups differ by construction, since being investigable enough to yield a cause is itself a marker of comorbidity. It is carried at that weight and no further.

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Association between drug-induced erythrocytosis and thrombotic events — LIMITED. ⚠⚠ Across these designs the data are compatible with the erythrocytotic phenotype acting partly as a marker of underlying thrombotic risk, and they DO NOT DISTINGUISH that interpretation from a causal contribution of the haematocrit rise. None isolates mediation by the rise. Drug-induced erythrocytosis occurs in populations with measurable thrombotic risk; baseline erythrocytosis and underlying vascular disease may identify higher-risk patients; and the available analyses establish neither mediation by haematocrit nor its absence.

⚠ CORRECTION NOTICE, IN ITS OWN PARAGRAPH. An earlier version concluded that three designs CONVERGE on the phenotype marking a high-risk patient RATHER THAN the haematocrit acting as the operative exposure. Correction 114.

⚠ CORRECTION NOTICE, SECOND LIMB, IN ITS OWN PARAGRAPH. The same version described the randomised analysis through its NULL STRATUM ESTIMATES, which understates it. Correction 120.
Causal attribution to the haematocrit rise — NOT ESTABLISHED. The randomised design identifies the effect of the DRUG, not of the haematocrit it produces, and the drug has many other actions. No analysis in this class isolates the erythrocytosis.
Observed effect size — NOT ESTIMABLE for the haematocrit contribution. The drug-level and phenotype-level estimates are estimable and are given above.
Population directness — indirect, and the mechanism differs. ⚠ Sodium-glucose cotransporter 2 inhibition raises haematocrit through a different route from testosterone, in older people with type 2 diabetes and high cardiovascular risk. One of these cohorts states that testosterone therapy was not evaluated at all, so testosterone users sit unmeasured within it.[60]

⚠⚠ WHAT THIS CLASS DOES FOR THIS REVIEW, STATED PRECISELY. It does not license any claim about trans-masculine people. What it does is supply randomised and observational evidence bearing on the question §9.4.1 and §9.4.2 leave open — whether a raised haematocrit is the thing that carries the risk or a marker of who is carrying it.

⚠⚠ AND IT DOES NOT ANSWER THAT QUESTION. NONE OF THE THREE DESIGNS IDENTIFIES MARKER VERSUS MEDIATOR, AND THE REASONS ARE DIFFERENT IN EACH. The randomised analysis randomises the DRUG and not the haematocrit, so baseline haematocrit enters as an effect modifier and null estimates inside the erythrocytotic stratum do not show the haematocrit to be non-causal. The absence of a gradient across peak-haemoglobin tertiles does not distinguish a marker from a causal intermediate: a causal relation could be non-linear, thresholded, obscured by treatment, time-varying, or misclassified by a peak value. And the Swedish cohort begins with people who have ALREADY had a thromboembolic event and then compares aetiologically different erythrocytosis groups — groups that differ by construction — so it cannot identify whether haematocrit mediates INCIDENT thrombosis.

⚠⚠ WHAT THE CLASS ESTABLISHES IS THEREFORE A COMPETING EXPLANATION, NOT A RESOLUTION. Underlying phenotype and comorbidity are a plausible competing explanation for the association between erythrocytosis and thrombotic events, and the haematocrit increment’s own contribution is not isolated in any of them. ⚠ That is compatible with this review’s own trans-masculine observation at §9.4.3 — that none of the five events reported with an event-time haematocrit met that study’s erythrocytosis criterion — and it does not confirm it.

⚠⚠ AND THE MOST INFORMATIVE FEATURE OF THE RANDOMISED ANALYSIS IS THE INTERACTION, NOT THE NULL STRATA. A statistically significant treatment-by-baseline-haematocrit interaction, holding under both categorical and continuous representations and with a three-way sex interaction at P = 0.02, is evidence that baseline haematocrit MAY MODIFY the drug’s cardiovascular effect.[59] The subgroup intervals crossing unity correctly prevent a claim of harm within the erythrocytotic stratum; they do not erase the interaction. ⚠ The defensible synthesis is that there is exploratory evidence of treatment-effect heterogeneity by baseline haematocrit in males, and that sparse subgroup events, absent multiplicity adjustment, and the absence of haematocrit randomisation or any mediation analysis prevent attributing it to the haematocrit itself.

⚠⚠ AND THE PAPER’S OWN AUTHORS REACH A DIFFERENT READING FROM THIS REVIEW’S, WHICH IS STATED RATHER THAN RESOLVED IN THIS REVIEW’S FAVOUR. They conclude that canagliflozin may pose a safety concern for thromboembolism in males with erythrocytosis at baseline.[59] ⚠ This review’s §9.6 entry 31 criticises the paper’s TITLE for asserting harm in a stratum whose own estimates are null. Criticising a headline and then recruiting the paper to the opposite conclusion are two different acts, and an earlier version of this passage ran them together. Correction 120.

9.5 Formulation — randomised and observational evidence remains inconclusive

9.5.1 Three questions, kept apart

Does formulation change the haematocrit? Does management after a raised haematocrit lower it? Does intervention reduce thrombotic or cardiovascular events? The third question has no established comparative benefit in this population in the evidence identified here. Symptomatic response is a separate clinical outcome, addressed by uncontrolled reports. The first two questions are no longer supported only by observational evidence.

9.5.2 Formulation and haematocrit: two randomised formulation trials identified and three non-randomised comparisons

⚠ A second randomised formulation trial has been identified via citation in the Okano systematic review [36]. Pelusi et al. (2014) randomised 45 female-to-male transsexual persons to testosterone enanthate or esters, transdermal gel, or testosterone undecanoate, following haematological outcomes for 54 weeks. Haematocrit increased in all three arms.[82] This review has not independently verified a formal between-formulation haematocrit contrast from the primary report; characterisation is based on [36]’s secondary description. This trial was not retrieved by this review’s direct search; it is registered as a secondary-citation source read through [36].

The first directly retrieved randomised trial also detected no statistically significant between-formulation difference. ⚠ It was not an equivalence trial. Fifty-eight testosterone-naive trans men were randomised to enanthate or undecanoate; at one year, haematocrit 38·0 against 38·5% (p = 0·627) and haemoglobin 12·5 against 13·0 g/dL (p = 0·478), despite six injections against eighteen.[25] ⚠⚠ AND ITS DISCUSSION STATES THAT NO PARTICIPANT EXCEEDED A HAEMATOCRIT OF 48%.[25] The only directly retrieved randomised formulation comparison in this review therefore contains NO EVENT AT ANY THRESHOLD IT EXAMINES — not 50%, not 52%, not 54%. It compares trajectories within the normal range; it does not compare erythrocytosis risk, because no erythrocytosis occurred in it (correction 596). ⚠ Its baseline of 35–36% is also far below the 42% of the cohorts §9.2 draws on, so neither the starting point nor the ceiling transfers without argument.

The observational studies report different outcomes and contrasts. One compares haematocrit CHANGE between formulations; another reports threshold-crossing ODDS relative to gel. ⚠⚠ THEIR RESULTS DO NOT PROVIDE A DIRECT TEST OF OPPOSING UNDECANOATE-VERSUS-ESTER EFFECTS. Together with the small randomised comparisons they do not establish a reproducible causal ordering of formulations for erythrocytosis risk, and a non-significant ester-versus-gel estimate does not establish equivalence (correction 557). The prospective multicentre cohort found undecanoate gave a smaller rise than shorter-acting esters — a difference of 0·8 percentage points, with 9·2% against 15·9% reaching ≥50% and none against 6·8% reaching ≥52%.[22] The Amsterdam cohort reports no undecanoate-versus-ester contrast at all. What it does report is a different comparison: relative to gel, undecanoate carried the highest adjusted odds of exceeding 0·50 L/L (OR 3·1, 1·7–5·6), the ester-versus-gel adjusted estimate was 1.1 (95% CI 0.7–1.7).[28]

⚠⚠ AND A THIRD DATASET, PAIRED AND PROSPECTIVE, ESTIMATED LITTLE BETWEEN-ROUTE DIFFERENCE. In 100 trans men — 59 on intramuscular testosterone and 41 on transdermal gel — the twelve-month change in haematocrit differed between routes by −0·008 L/L (95% CI −0·028 to 0·012), transdermal as reference, and by −0·005 (−0·026 to 0·016) after adjustment for age. Its authors describe the differences between routes as small.[27] ⚠ It is not a randomised comparison, and allocation was by clinical practice.

⚠ This review does not adjudicate between them. These studies differ in design, outcome and comparison group. Their findings do not establish opposing undecanoate-versus-ester effects or a reproducible causal ordering of formulations for erythrocytosis risk (correction 570). ⚠⚠ What has changed at v4.13 is the balance rather than the grade: the randomised comparison and one non-randomised comparison estimated little route difference; the two observational comparisons differ in outcome and contrast and are not summarised as opposite effects.

⚠ Two additional studies have been identified via external review: a retrospective pellet cohort (Cheng et al.) and a prospective testosterone-undecanoate cohort (Lethin et al.), Cheng et al. (2023; PMC10732158) reported a retrospective pellet/implant switch cohort of 30 trans men in which 14 of 30 reached haematocrit ≥50%; some participants had elevated haematocrit before the switch, limiting interpretation as incident pellet-specific risk. Lethin et al. (2024; PMID 38085088) reported a prospective testosterone undecanoate study with a treatment-naïve trans-male subgroup followed for one year; this has since been integrated into §9.2 as a secondary-citation source (correction 348; Hct 42→46%, n=10, 2/10 ≥50%).

What all four share is that none measured the mechanism. Short-acting esters are said to drive erythropoiesis through supraphysiological peaks. No study reviewed here measured individual pharmacokinetic profiles. Serum testosterone did not differ significantly between formulations in the prospective cohort (p = 0·387), and its authors state they could not link the formulation difference to testosterone levels because peak concentrations were never measured.[22] The randomised trial reports one-year testosterone of 865 and 866 ng/dL — nearly identical point estimates — while its authors note mass spectrometry was not used.[25]

Three further limits on the observational comparisons. Allocation was not randomised in either cohort, and one site used undecanoate exclusively, partly confounding centre with formulation.[22] Route was unavailable for roughly 38% of Amsterdam measurements, and the unknown-route category itself carried an elevated adjusted odds ratio of 1·9 (1·3–2·8) — indicating missingness associated with outcome risk.[28] And the categorical differences rest on few events in small subgroups.

⚠ The formulation landscape also includes pellets/implants and oral testosterone undecanoate. Current trans-specific professional guidance names implant pellets and oral testosterone undecanoate as recognised formulations. Neither has prospective direct trans-masculine haematocrit evidence of the quality reviewed in this section; evidence is sparse and the pellet cohort data identified (Cheng et al. 2023) is from a selected switch population. Their erythrocytosis risk should be treated as unknown rather than equivalent to the formulations with direct comparison data.

⚠ A discrepancy in [36]: the systematic review characterizes comparative formulation evidence as arising from ‘small observational studies’, but PubMed classifies Pelusi et al. (2014) as a randomized controlled trial. This review does not transmit [36]’s characterization unqualified; Pelusi is registered as an RCT per PubMed [82].

⚠ An independent systematic review reaches the same place, describing the comparative data as conflicting and drawn from small observational studies, and its own discussion is internally inconsistent — endorsing the peak account in one passage and the contrary finding in another.[36]

9.5.3 Management after a raised haematocrit: it lowers the number

A randomised pilot compared continued reduced-dose testosterone against complete withdrawal in trans men with haematocrit at or above 50%. Forty-three completed. Withdrawal reduced haematocrit more (−3·5 against −0·8 percentage points), while dose reduction preserved therapeutic testosterone concentrations. No serious adverse events were reported over the three-month study period; with 43 completers, this is descriptive and is not treated as a safety estimate.[81]

⚠⚠ READ IN PUBLISHED ABSTRACT AND REGISTERED PROTOCOL; FULL TEXT NOT OBTAINED. The registered protocol names its primary outcome as reduction of haematocrit below 50% and four secondary outcomes — bleeding pattern, anxiety and depression scores, hormone maintenance, biochemical alterations. ⚠⚠ No thrombotic outcome was prespecified.[62]

⚠⚠ AND THE TWO REGISTRATIONS DISAGREE ON WHETHER AN UPPER HAEMATOCRIT BOUND WAS PRESPECIFIED. The ClinicalTrials.gov record excludes patients with Ht > 54;[65] the REBEC record registers no haematocrit ceiling.[62] ⚠ An upper bound at this review’s own threshold is therefore prespecified in one of the two primary registrations (correction 457). ⚠⚠ WHAT THE PUBLISHED REPORT STATES AS ITS EXCLUSION CRITERION IS NOT ESTABLISHED BY THIS REVIEW. The full text is not held; [81]’s extraction is bounded to realised sample size, between-arm haematocrit change and adverse-event observation, and an eligibility criterion is a prespecification field routed to [62] and [65]. Neither held record states 55. Any figure differing from the registered > 54 would be a further divergence between registration and publication, and is not asserted here (correction 498). The trial cannot speak to the upper end of the range, and on the registered record that is by design rather than by reporting — the same structural feature as the large cardiovascular-outcome trial, which curtailed exposure above 54% by protocol. Two trials, independently, excluded the top of the range.

A retrospective cohort now separates the two management strategies, and only one of them showed a detectable effect. Among 181 trans men, 49 developed erythrocytosis over 104 tests. Discontinuation for three months lowered haematocrit in every group — by 7·1 points on monthly cypionate, 5·2 on fortnightly cypionate, and 2·7 on undecanoate — while extending the cypionate interval from 14 to 28 days detected no statistically significant change in haematocrit (p = 0·158), haemoglobin (p = 0·171) or testosterone (p = 0·227).[63]

⚠⚠ And undecanoate discontinuation lowered haematocrit and haemoglobin WITHOUT a significant fall in total testosterone (p = 0·147). ⚠⚠ That is an absence of a detected difference, not evidence of preserved concentrations, and an earlier version of this section read it as the latter. What the study establishes is that haemoglobin and haematocrit fell while no significant fall in total testosterone was detected over the study interval. ⚠ The randomised pilot does report preservation of therapeutic testosterone with dose reduction; that claim belongs there and not here.[63]

⚠ That cohort states both its measurement method and sampling time — an automated counter, with samples drawn between 07:00 and 09:00. Pre-analytical conditions are otherwise rarely reported in the sources examined here; no corpus-wide count is asserted.

⚠ Dose reduction is the most commonly reported management strategy in the cohorts reviewed here. Among those developing erythrocytosis in one records study, 42% had a dose reduction and 4·8% underwent phlebotomy; the rest were observed.[29] A second found no standardised approach at any haematocrit band, with observation the commonest response even at ≥54%.[30]

9.5.4 Whether lowering haematocrit reduces thrombotic or cardiovascular events: benefit not established; indirect evidence is population-dependent

No comparative study identified here establishes that lowering haematocrit reduces thrombotic or cardiovascular events in transmasculine people. Galamgam and colleagues report symptom relief following phlebotomy in two testosterone-treated patients. Table 2 gives peak haematocrits of 51% and 48%; Table 1 lists topical corticosteroid use only for the first patient.[102]

Both reported peaks were below 54%, the commonly cited on-treatment action level discussed in this review. This is not evidence that symptoms occurred below every possible clinical indication for intervention. Peak values do not specify the haematocrit immediately before each procedure. The authors’ erythrocytosis attribution does not establish that either value is abnormal under every applicable reference interval, or that haematocrit mediated the itching or its improvement.

The report is uncontrolled and does not establish treatment efficacy, a numerical intervention threshold or thrombosis prevention. It provides clinical information about symptoms, not a validated threshold for management. Source [102] is now in the numbered reference list. The reviewer’s focused appraisal of it is HELD AND NOT DEPOSITED with this manuscript, being the reviewer’s own document; the author’s register was reconciled with it at correction 587, after each of its eight claims was checked against the full text (correction 673).

⚠⚠ AND A HAEMATOLOGY SOCIETY SAYS THE SAME THING ABOUT SECONDARY AND IDIOPATHIC ERYTHROCYTOSIS, IN ITS OWN VOICE, IN THE GUIDELINE DEVOTED TO MANAGING THEM.[17] Of idiopathic erythrocytosis it states that there are NO CLINICAL STUDIES evaluating the use of aspirin or venesection, and that EVIDENCE-BASED RECOMMENDATIONS CANNOT THEREFORE BE MADE; that THERE IS NO EVIDENCE ON WHICH TO DETERMINE THE TARGET HAEMATOCRIT, because the relevance of polycythaemia vera data is unclear; and that its own suggested target of below 0·55 is ARBITRARY — its word, at GRADE 2C. ⚠ It recommends against cytoreductive therapy in that condition at GRADE 1B, and notes that contemporary data suggest the thrombotic and bleeding risk is LOW.

⚠⚠ THAT IS THE STRONGEST EXTERNAL CORROBORATION IDENTIFIED HERE OF THE THRESHOLD-TRANSFER BRANCH OF THIS REVIEW’S ARGUMENT — a close external analogue to the finding that an intervention threshold can remain clinically specified while its outcome basis stays unresolved. ⚠⚠ IT DOES NOT CORROBORATE THE REVIEW’S CENTRAL CAUSAL PROPOSITION, which concerns the thrombotic effect attributable specifically to testosterone-associated haematocrit elevation in trans-masculine people, and on which a haematology guideline about a different condition is silent. ⚠ Correction 182, and correction 111’s class for the second time: an external source that strongly supports ONE branch was described in terms that reached the whole argument. ⚠ It is not about trans-masculine people, it does not license a number, and it is carried for what it is: a haematology guideline declining to set an evidence-based target in a secondary erythrocytosis because the evidence to set one does not exist.

⚠⚠ AND THE SAME GUIDELINE SUPPLIES THE DIRECT COUNTEREXAMPLE TO THRESHOLD TRANSFER, WHICH IS MORE VALUABLE TO THIS REVIEW THAN THE CORROBORATION. In hypoxic pulmonary disease, venesection reducing the haematocrit to 0·50–0·52 improved exercise tolerance, a further staged reduction to 0·45 gave NO ADDITIONAL BENEFIT, and a haematocrit BELOW 0·45 IS, in the BSH guideline’s characterisation, detrimental to those patients — on the basis of a retrospective comparison in severe chronic obstructive pulmonary disease patients on long-term oxygen, in which a haematocrit at or above 0·55 carried a better prognosis than ANAEMIA.[17] ⚠⚠ THAT IS AN ASSOCIATION BETWEEN OBSERVED LEVELS IN PATIENTS WHO WERE NOT VENESECTED TO ANY TARGET, and it may reflect disease severity and other causes of anaemia, which is plausible and is not quantified here. The guideline’s own INTERVENTIONAL finding is the neighbouring one: staged reduction to 0·45 gave no additional benefit over 0·50–0·52 (correction 516). ⚠⚠ The polycythaemia vera target is therefore not transferable to another erythrocytosis: the same guideline sets 0·50–·52 there and records that a staged reduction below it did not demonstrate additional benefit for the endpoint studied. Same society, same guideline series, same authors, different disease and different outcome. The cited exercise-tolerance evidence did not demonstrate additional benefit from further reduction under the conditions studied; that is not an opposite effect on thrombosis, and it is not equivalence (correction 542).

⚠ The condition-specific targets that guideline sets, listed so the spread is visible: polycythaemia vera below 0·45 at GRADE 1A; idiopathic erythrocytosis an arbitrary 0·55 at 2C, or 0·45 only where a prior thrombosis is thought related to the erythrocytosis; hypoxic pulmonary disease 0·50–0·52, and only where symptomatic from hyperviscosity or above 0·56; congenital erythrocytosis 0·52, with the explicit instruction NOT to reduce to the normal range and a note that venesection may be DETRIMENTAL in Chuvash polycythaemia; post-transplant erythrocytosis 0·50, where venesection reduces the haematocrit but DOES NOT REDUCE THROMBOSIS.[17]

⚠⚠ ONE SOCIETY, ONE GUIDELINE, FIVE ERYTHROCYTOSES AND FIVE DIFFERENT TARGETS, SET BY ONE SOCIETY IN ONE GUIDELINE. The conditions and the outcomes differ, and the targets should not be read as opposite effects on thrombosis (correction 564). This review’s population-transfer argument has until now been made from first principles. It is now made by the field.

In polycythaemia vera, a randomised trial found it does. The primary endpoint occurred in 2·7% of the below-45% group and 9·8% of the 45–50% group; ⚠ the HIGHER-target group carried a hazard ratio of 3·91 (1·45–10·53) RELATIVE TO the lower-target group — 4·4 against 1·1 events per 100 person-years.[38] ⚠ An earlier version stated the ratio without naming which arm it attaches to, which reads as though it attaches to the first arm named. ⚠ The trial closed at 365 of a planned 1000 for slow recruitment, its authors call the benefit largely unanticipated, and they concede in their own discussion that persistent leukocytosis in the less-treated arm could have contributed, since hydroxyurea use differed between arms.

In secondary and idiopathic polycythaemia, an observational study did not detect an association. Venesection modelled as a time-dependent exposure gave a hazard ratio of 0·88 (0·36–2·17), and persistence of the erythrocytosis was likewise unassociated with thrombosis. Haematocrit fell by 0·04 in the first year irrespective of venesection, which the authors attribute to reversion to the mean or lifestyle advice.[35]

⚠⚠ Polycythaemia vera is a clonal myeloproliferative disease with disease-specific thrombotic biology. Testosterone-associated erythrocytosis is a secondary erythrocytosis and does not share that clonal pathophysiology. The randomised result is not transferable, and this review does not transfer it.

⚠ And a 2026 secondary-erythrocytosis cohort adds a larger real-world person-time dataset — ⚠⚠ the randomised trial above already reports events per 100 person-years, so this is additional person-time rather than the first. A 2026 single-centre cohort of 581 patients with secondary erythrocytosis accrued 2,121 patient-years of follow-up overall. ⚠ The combined arterial and venous thrombosis analysis included 567 of them, contributing 2,029 patient-years, with 55 events — 2·71 per 100 patient-years — significantly higher than the 1·40 observed in 103 patients with treated polycythaemia vera, hazard ratio 3·16 (1·25–7·95).[64]

⚠⚠ That comparison is confounded by treatment and its authors say so. Antiplatelet agents were used in 37% of the secondary group against 83% of the polycythaemia vera group, and hydroxyurea in 2% against 87%. It is substantially untreated against treated. ⚠ And in its multivariable model only diagnosis group and age predicted thrombosis — smoking, sex, prior thrombosis, diabetes, hypertension, body mass index and sleep apnoea were all null, in a cohort selected for carrying those very factors.

⚠ Its androgen-exposed subgroup is the lowest of five named aetiologies: 3 events in 55 people over 229·23 patient-years, 1·31 per 100 patient-years, against 3·00 for smoking, 2·45 for hypoxic respiratory disease and idiopathic erythrocytosis, and 1·68 for SGLT2 inhibitor use.[64] ⚠⚠ Three events give an exact Poisson interval of 0·27 to 3·82, which contains every other subgroup’s rate. The ordering is not a separation, and this review does not report it as one. The cohort is not trans-specific: 54 men and one woman, with no gender identity recorded.

⚠⚠ An earlier version of this section said the obvious remedy — borrowing a rate from a population where erythrocytosis has been studied properly — was barred by the two sources this review held. That is no longer true, and the claim is withdrawn. A rate now exists. It is confounded, it is not trans-specific, and it does not isolate haematocrit; but it exists.

⚠ And lowering the haematocrit is not free. ⚠⚠ The harms below are taken from the DISCUSSION sections of two cohort papers, which under Amendment C are secondary sources for the primaries they cite. Those primaries have not been opened, and the statements are therefore attributed as characterisations rather than as established findings.

As characterised by those authors: venesection can induce iron deficiency, syncope and nerve injury, and in cyanotic congenital heart disease is said to increase stroke risk and to be reserved for severe hyperviscosity;[35] and where erythrocytosis is a compensatory response to hypoxia, lowering it is said to risk reducing oxygen delivery.[64] ⚠ This review does not assert either as verified.

⚠⚠ AND THE FOUR THINGS A MANAGEMENT DECISION TRADES BETWEEN ARE NOT ONE THING. A haematological response, a symptom response, a thrombotic outcome and the burden of the treatment itself are separate, and the evidence differs across them: haematological responses have been studied, symptomatic improvement after phlebotomy has been described in an uncontrolled two-patient report, and no comparative evidence establishing thrombotic benefit from intervention has been identified here (correction 540). ⚠ Dose reduction, interruption, switching formulation and venesection are not interchangeable: they differ in what they do to hormone concentrations, to bleeding and menstruation, to iron status, to wellbeing, and to a person’s access to and continuity of gender-affirming care.** ⚠⚠ THE ASYMMETRY IS THE POINT AND IT RUNS BOTH WAYS. Mechanistic concern about repeated phlebotomy does not establish clinical harm, exactly as lowering a haematocrit does not establish thrombosis prevention. Neither absence is a finding (correction 529). ⚠ No antiplatelet or other pharmacological recommendation follows from anything in this review, and none is made.

⚠ Switching to transdermal administration is a commonly proposed management strategy, supported by pharmacokinetic rationale, by some observational data and by the Amsterdam authors’ own recommendation.[28] ⚠⚠ The small randomised comparisons did not establish a between-formulation difference, and the non-randomised studies report different outcomes and contrasts. They do not establish a reproducible causal ordering of formulations or demonstrate a clinical-outcome benefit from switching (correction 570). ⚠ No comparative evidence establishing a clinical-outcome benefit from switching has been identified in this review.

Grading block — §9.5, three statements, deliberately separated

1 — Formulation changes the haematocrit response

Association — LIMITED. ⚠ Heterogeneous evidence. One small randomised superiority trial found no statistically significant between-formulation difference (it was not powered as an equivalence trial). The non-randomised comparisons differ in outcome and contrast; they should not be summarised as demonstrating opposite formulation effects. The evidence is insufficient to establish a reproducible formulation effect.
Causal effect of formulation on haematocrit — LIMITED. ⚠ Two small randomised formulation trials have been identified; neither established a clinically important between-formulation difference, and neither was powered as an equivalence trial. The non-randomised comparisons differ in outcome and contrast; they should not be summarised as demonstrating opposite formulation effects, and no reproducible causal ordering of formulations is established. The evidence corpus is insufficient to establish a reproducible causal formulation effect. Randomisation can estimate a formulation effect without measuring pharmacokinetics, and an earlier version of this block conflated the two.
Peak-mediated pharmacokinetic mechanism — NOT ESTABLISHED. No study reviewed here measured individual profiles.
Population directness — direct.

2 — Management after a raised haematocrit lowers it

Association — MODERATE. Randomised: withdrawal −3·5 against dose reduction −0·8 percentage points.
⚠⚠ THE TWO TRIAL REGISTRATIONS HAVE NOW BEEN READ FIRST-HAND, AND THEY CLOSE SOME APPRAISAL ITEMS AND OPEN ONE.[65][62] Closed: random allocation by computer-generated number; parallel groups; an actual enrolment of 46 against a planned 40, so the trial OVER-recruited; a prespecified PRIMARY OUTCOME that is haematological — whether halving the cypionate dose normalises the haematocrit, measured by automated counting at recruitment and three months; prespecified secondaries confined to lipids and glucose, an anxiety and depression scale, bleeding pattern and hormone maintenance; and study completion on 17 December 2024 with primary-outcome data collection ending 18 November 2024.

⚠⚠ THIS REVIEW’S EARLIER STATEMENT THAT NO CLINICAL OUTCOME WAS PRESPECIFIED IS WITHDRAWN, AND ITS OWN SENTENCE ABOVE IS WHY. An anxiety and depression scale and a bleeding pattern ARE clinical outcomes. What the registrations establish is narrower and is what this review now says: no THROMBOTIC OR CARDIOVASCULAR endpoint appears in either registration.[65][62] ⚠ The registered primary outcome is haematological. Prespecified secondaries include bleeding pattern and an anxiety and depression scale alongside hormonal, lipid and glucose measures. The trial therefore informs haematological management and may inform symptomatic or patient-reported outcomes; it does not establish thrombosis prevention (correction 487).

⚠⚠ AND ONE APPRAISAL ITEM IS OPENED RATHER THAN CLOSED: THE TWO REGISTRATIONS DISAGREE ON MASKING. One records TRIPLE masking — care provider, investigator and outcome assessor. The other records the trial as SINGLE-BLIND, described as blinded to the researcher and statistician.[65][62] ⚠ Two primary registry records of one trial, disagreeing on a risk-of-bias item. §9.6 carries it.

⚠⚠ AND THE REGISTRATIONS DIVERGE ON THREE FURTHER FIELDS, ONE OF WHICH IS A STALENESS RATHER THAN A DISAGREEMENT. Recruitment status — Completed[65] against Recruiting[62]; enrolment — 46 actual[65] against a target of forty[62]; phase — Not Applicable[65] against Phase 3[62]. ⚠ [62] also posts no results and carries a declared completion date that has passed. It is read as a record of what was registered, not of what the trial did (correction 457).

⚠ Still unappraisable without the full text: allocation concealment mechanism, attrition, analytic population, and which masking description is correct.

⚠⚠ THE FULL TEXT IS IDENTIFIED AND HAS NOT BEEN OBTAINED BY THIS REVIEW, AND THE REASON IS RECORDED RATHER THAN GLOSSED. Okano SHP, Franceschini SA, Cantelli DAL, Rosa e Silva ACJS, Ferriani RA, Brito LGO, Lara LAS. Effect of 100 mg of testosterone cypionate in trans men with erythrocytosis: a randomized controlled pilot study. J Sex Med 2025;22(11):2149–53, 10.1093/jsxmed/qdaf263. ⚠ No route available to this review as of 4 September 2026 yields a copy that can be deposited in its source archive. The routes examined would at best permit on-screen reading without a retainable file.

⚠⚠ THAT DISTINCTION IS NOT PEDANTRY IN THIS REVIEW, BECAUSE THE ARCHIVE IS THE MECHANISM BY WHICH THE SAME-STEP RULE IS ENFORCED. CORRECTION 510 SEPARATED READING FROM HOLDING AND THIS PARAGRAPH DID NOT FOLLOW IT. Lawful on-screen access can support appraisal where the version, access date and page or table locations are recorded; lack of a retainable copy is recorded separately from lack of access, and an author’s reading record and independently repeatable access are distinct audit features, neither of which is an archive holding (correction 539). The read state of this source is unchanged and remains unread; a policy change is not a read. An earlier form held that a source which cannot be deposited cannot be re-opened in the step a claim is written from it, cannot be gate-checked, and cannot be verified by anyone auditing this review afterwards. ⚠⚠ THAT PROHIBITION IS SUPERSEDED AND IS RETAINED ONLY AS AN EARLIER FORM. Lawful on-screen access CAN support appraisal where the source version, access date and relevant page or table locations are recorded. Read status, retention status and permission to redistribute a copy are three separate fields. ⚠ This source remains unread in full: changing the access policy does not change that status (correction 561).

⚠ This is a statement about THIS REVIEW’S retrieval, not about the paper’s availability in the world — corrections 64, 119 and 130. No characterisation of the publisher or its access process is offered, because none is established.
>
> Causal attribution — LIMITED — full-text appraisal not performed. ⚠⚠ The qualifier is changed at v4.30, because the previous wording implied a retrieval in progress and there is none: the full text was not obtained through this review’s available access routes in a depositable form. ⚠ Random allocation materially strengthens causal identification, but only the published abstract, the two trial registrations and the registered protocol were appraised; allocation implementation, allocation concealment, departures from assigned treatment, attrition, analytic population, outcome completeness and selective reporting could not therefore be assessed.
>
> ⚠⚠ THIS MARK WAS MODERATE AT v4.5 AND IS DOWNGRADED HERE. The objection external review made, and this review accepts, is that MODERATE treated randomised as sufficient rather than as one input: the realised causal estimate cannot be graded above LIMITED when the risk-of-bias features that determine whether randomisation was preserved in execution are unappraisable from an abstract and a registry entry. ⚠ The qualifier is prose beside the existing mark and NOT a new grading token; §3 records the rule rather than the scheme gaining a provenance-limited category. ⚠ The mark may revert to MODERATE if the full text is obtained and appraised in primary, if the risk-of-bias features it contains support the causal grade — obtaining the text might equally confirm the LIMITED mark if the methods are inadequately reported. This is Eden’s to decide.
>
> Observed effect size — LIMITED. 43 completers, one centre, three months.
> Population directness — direct.

3 — Lowering testosterone-associated haematocrit reduces thrombotic or cardiovascular events in trans-masculine people

NONE IDENTIFIED — no eligible study identified in this population. ⚠⚠ The nearest evidence comes from two other populations and does NOT point in opposite directions: randomised benefit in a clonal myeloproliferative disease, and an imprecise observational estimate in secondary erythrocytosis, 0·88 (0·36–2·17), whose interval permits appreciable benefit and appreciable harm. ⚠⚠ CORRECTION 517 SAID THAT INTERVAL CONTAINED THE RANDOMISED RESULT. IT DOES NOT, ON EITHER ORIENTATION: the randomised hazard ratio is 3·91, and inverted it is 0·26 with an interval of 0·10 to 0·69 (correction 531). ⚠ The conclusion is unchanged and rests where it should have rested: these are different diseases, different interventions and different biases, and an imprecise observational estimate is not evidence of an effect opposite to a randomised one. ⚠ A detected effect and an undetected effect are not opposite directions, and treating them as contradictory is the move this review indicts elsewhere (correction 517).
Population directness — indirect, and the transfer is refused.

⚠⚠ §9.6 IS FINISHED. NUMBERED ONCE, ORDER FIXED.

⚠⚠ ENTRY 9 IS NOW PERMANENTLY WITHDRAWN, AGAINST PRIMARIES, AND THE WITHDRAWAL IS VINDICATED. Entry 9 charged the Discussion with dropping the three lowest thresholds in its own table. Its withdrawal at v0.3 was provisional, on the suspicion that the Discussion might be describing a different clinical decision. ⚠⚠ Eight of the nine Cannarella rows have been verified in primary, and the Discussion’s claim — that societies suggest monitoring to reduce or discontinue above 54%, with 50% for AACE and ICSM — is CORRECT for seven of the eight as to the 54% rule and for AACE at 50%. ⚠⚠ It is WRONG about ICSM, whose on-treatment level is 54% and not 50%; the 50% is ICSM’s initiation contraindication, and the Discussion has made the same substitution the table makes elsewhere.[7][8][9][11][12][13][15] The Discussion was describing the ON-TREATMENT object accurately while the table beside it lists the INITIATION object, and this review mistook the difference for a compression. ⚠ The charge is withdrawn permanently rather than left provisional. A charge against named authors that dissolves on reading the primaries is recorded at full weight, and it is the second in this ledger to do so.

⚠⚠ AN EARLIER WITHDRAWAL NOTE NAMED ENTRY 6 AND WAS WRONG. Entry 6 concerns a different paper and a different compression; it was never in question and it stands. The error withdrew a sound charge and left the unsafe one live — the exact inversion of its purpose, in the section whose fixed numbering is offered as a reliability mechanism. It is recorded rather than quietly repaired.

✅ Entry 8, the inverted inequality, is unaffected and stands. ⚠ Live entries are therefore one fewer than the numbering implies; the count is not renumbered, because the order is fixed.

9.6 How qualification is lost — a ledger of what happened to these findings in transmission

⚠⚠ AND ONE INSTANCE HAS BEEN FORMALLY CORRECTED BY ITS OWN AUTHORS, WHICH MAKES IT VERIFIABLE IN BOTH DIRECTIONS. A 2025 literature review of venous thromboembolism risk in trans-masculine people stated that the only significant change induced by four months of testosterone in a 14-person cohort was a decrease in the level of activated protein C. ⚠⚠ A corrigendum published on 21 August 2026 replaces that with a decrease in the normalised activated protein C sensitivity ratio, reflecting REDUCED acquired resistance to activated protein C — and records that the original authors concluded testosterone had a mild ANTITHROMBOTIC effect.[66]

⚠⚠ THE ERROR INVERTED THE DIRECTION OF THE FINDING. A fall in the level of an anticoagulant protein reads as pro-thrombotic; a fall in the resistance-to-anticoagulant ratio reads as its opposite. It appeared in a review whose subject is thrombotic risk in this population, and it stood for more than a year before correction.

⚠ This review has read the corrigendum and not the original article, and states only what the corrigendum states.

Front matter — before the entries, excluded from the count

What this ledger is, and what it is not

Every finding in this section reached its clinical readership through other people’s sentences. Guidelines cite reviews; reviews cite cohorts; a clinician reads the guideline. At each step a finding can lose a qualifier — the population it was measured in, the adjustment it did or did not survive, the width of the interval around it — without losing its number. The number is what travels. The qualification is what is dropped.

This ledger records what happened to the findings reviewed in this tranche as they moved between papers. It is not a survey of the literature’s reliability, and it should not be read as one. Its claim is narrower and, if it holds, more useful: that qualification is lost at recurring points — and that being aware of them shapes what a careful reader looks for in advance. ⚠ The ledger demonstrates recurrence across a purposive sample, not systematic predictability across the literature.

That claim is only worth anything if the clean transmissions are counted alongside the failed ones. A ledger that records only errors demonstrates that its compiler looked for errors. Several of the transmissions examined here are accurate, one of them more faithful to a paper’s data than the paper’s own abstract, and they are recorded at the same weight as the failures.

Prior numbering is superseded. Four working records of this review carried entry numbers for these instances, and at least three numbers named different instances in different records. Those schemes are discarded rather than reconciled, and each instance below is rebuilt from the primary papers. Reconciling four secondary records against one another would have been this section’s own subject matter, enacted in this section’s own construction.
*
One source in this section is a secondary document, and the exception is deliberate. Schreijer 2010 is an editorial accompanying Braekkan in the same issue — no patients, no data, no analysis — and this review excludes it as evidence of anything about haematocrit and thrombosis. It appears here for one reason: it is the control that makes the entry against Folsom checkable.**

Folsom’s Discussion describes Tromsø’s categorical estimates as multivariable. Braekkan reports no multivariable categorical estimate anywhere — the categorical table gives crude rates and age-adjusted hazard ratios, and the multivariable model is applied to the continuous analysis only. The obvious defence is that the source was ambiguous. Schreijer closes it: writing first, in the same issue, from identical material, the editorial carries the categorical result without upgrading it — an upper-quintile comparison, roughly 1.5-fold, no adjustment claimed — and gives the continuous adjusted estimate separately and correctly. Two readers had the same paper at the same time. One of them did not make the error.

The editorial is not thereby endorsed, and this section enters an instance against it as well — a cohort-wide baseline smoking prevalence rendered as a prevalence among cases at the time of their thrombosis. It is examined on the same terms as every other source here, and it fails on one of them.
***
## The ledger
## 1. The anchor — an adjustment the source never performed

Braekkan and colleagues report the Tromsø Study’s categorical haematocrit analysis in Table 4, under a header that states what the table contains: crude incidence rates and age-adjusted hazard ratios [71]. Women with a haematocrit at or above 42% carried a hazard ratio for total venous thromboembolism of 1.51 (95% CI 1.08–2.12) against women below 39%; men at or above 46% carried 1.54 (1.08–2.21) against men below 43%. The paper’s multivariable model — age, smoking and body mass index — is applied to the continuous analysis in Table 3, where a 5% increment in haematocrit gives 1.25 (1.08–1.44) in the pooled population against an age-adjusted 1.35 (1.17–1.55).

The two estimate types therefore sit on facing pages, each under its own header, and the attenuation between them is visible: adjustment moves the continuous pooled estimate from 1.35 to 1.25, and removes significance from the women’s continuous estimate entirely.

Folsom and colleagues, reporting ARIC’s parallel analysis, characterise the Tromsø categorical figures in their Discussion as multivariable hazard ratios, and reproduce both numbers and both intervals exactly [45].

The figures are correct. The description is not, and it is not a mislabelling of an existing quantity. Braekkan reports no multivariable categorical estimate anywhere — not for women, not for men, not pooled. The object Folsom names does not exist in the source. A reader who took the characterisation at face value would understand Tromsø’s threshold-level finding to have survived adjustment for smoking and body mass index, when what survived adjustment was the continuous estimate, and when the paper’s own continuous analysis shows that adjustment costs the association a fifth of its excess and, in women, its significance.

The absolute rates behind these ratios are small and are given in the same table. Total venous thromboembolism ran from 1.11 to 2.57 per 1,000 person-years across the three haematocrit categories in women, and from 1.47 to 1.90 per 1,000 in men, against an overall crude rate of 1.58 per 1,000 person-years over 447 events and 282,494 person-years of follow-up [71].

This is the one instance in this ledger where the loss of qualification was demonstrated rather than inferred, in the sense that both the sentence and the table it describes were read against each other directly. That is the whole of its force, and it is the reason it is placed first.
***
## 2. The same claim, transmitted intact — and the transmission is the finding

Schreijer, Reitsma and Cannegieter’s editorial accompanied Braekkan in the same issue [47]. It carries the same categorical result and does not upgrade it: the upper 20th percentile is reported as associated with a 1.5-fold increased risk, with no adjustment claimed for it, and the continuous multivariable figure of 1.25 per 5% rise is given separately and correctly.

The editorial then argues at length that the association may not be causal at all — that conditions raising haematocrit and conditions raising thrombotic risk may rise together, that Braekkan’s data required adjustment the authors could not fully perform, and that the study accordingly does not exclude confounding by other conditions.

The editorial’s characterisation of the adjustment runs opposite to the error in §9.6.1, and it was published first. Two papers took the same table from the same issue of the same journal. One described an adjustment the table had not undergone; the other correctly identified the absence of that adjustment as the paper’s central limitation. The material available to both was identical, which is what makes this pair worth recording: the failure in §9.6.1 is not attributable to the source being unclear.

An editorial is secondary in its entirety, not merely in its introduction (Amendment C, extended at R8.3). It is cited here for what it says about Braekkan, verified against Braekkan, and for nothing else.
***
## 3. Where the same editorial does lose a qualification — a cohort denominator read as a case denominator

The editorial supports its confounding argument with two figures about Braekkan’s cases. One is accurate. The other describes a quantity Braekkan does not report.

Accurate: some medical condition was present in 22% of the venous thromboembolic cases. Braekkan’s Table 2, which tabulates the characteristics of the 447 patients at the time of their event, gives other medical condition as 21.7% overall — 24.6% in women, 18.6% in men [71].

Not supported: that 37% of the cases were smokers at the time of the thrombosis. Braekkan’s Table 2 contains no smoking row. The figures near 37% in the paper are in Table 1, which reports the baseline characteristics of the whole enrolled cohort in 1994–95: current smoking in 36.5% of the 13,714 women and 37.4% of the 12,394 men. The paper states in its limitations that smoking was recorded only as a baseline dichotomy, and that information on other medical conditions was available only for those who developed venous thromboembolism.

A cohort-wide baseline prevalence has been rendered as a case-level prevalence at the time of the event. The two are placed in a single sentence with the comorbidity figure, which is a genuine case-level figure, and the construction invites the reader to take both as properties of the cases.

The direction is worth stating plainly, because it does not run the way the errors in this review have run. The mistaken figure strengthens the argument the editorial is making — that smoking may account for the association — by appearing to show that smoking was prevalent among those who had events, when what it shows is that smoking was prevalent in Tromsø in 1994.
***
## 4. A significance boundary crossed in summary — Tromsø’s unprovoked estimates

Folsom’s Discussion contrasts ARIC’s larger provoked-VTE estimate with Tromsø’s, and reports that the Tromsø Study also found positive associations of haematocrit with unprovoked venous thromboembolism, at 1.62 (0.98–2.69) in women and 2.37 (1.36–4.15) in men [45].

Both figures and both intervals match Braekkan’s Table 4 exactly. But Braekkan’s own Results state the finding with a partition that the summary does not carry: haematocrit and haemoglobin were significantly associated with unprovoked venous thromboembolism in men and in the total population, but not in women [71]. The women’s interval crosses unity, and Braekkan says so.

This is a weaker instance than the anchor, and it is recorded as weaker. Folsom prints the interval alongside the claim, so an attentive reader can see what the plural is doing; nothing is concealed. What is lost is the source’s own explicit refusal to read the women’s estimate as positive — and the loss runs in the direction of a cleaner contrast between the two cohorts than the data support, which is the direction the passage’s argument benefits from.

Both sides of every instance below were open in the session this was written, per R13.3a — the review’s own rule that a propagation charge requires both sides read first-hand in the same step, and that an entry with only one side open is not an entry but a suspicion (Appendix R): Toss [39] 10.1016/j.ijcard.2013.05.065 and Warny [40] 10.1111/jth.14574, both untruncated. Warny was re-opened for this step rather than carried from the earlier step of the same session, because a characterisation cannot be verified against a summary of the characterising paper — that is the failure this section exists to catalogue.

## 5. The search was run in the other direction, and the erythrocytosis literature’s transmission of Toss is where it succeeded

Every instance recorded to this point moves an estimate toward the more alarming reading. A ledger that only ever finds error in one direction is describing its own method, not its material, so the same procedure was applied to a transmission selected for the opposite possibility.

Warny’s Discussion places Toss among the studies of haematocrit and coronary risk. Its sentence reports a study of 417,099 young Swedish men aged 18–19 which found a 1.3-fold increased risk of myocardial infarction at a haematocrit of 49% or above compared with 44% or below.

Toss’s fully adjusted estimate for that exact contrast — Table 2, model F, ≥49% against ≤44% — is 1.38 (1.29–1.48). Toss’s own abstract renders it as 1.4-fold, using the same contrast and very nearly the same words.

1.38 does not round to 1.3. Warny’s figure is below both the table value correctly rounded and the source’s own summary of it. The clause structure — N-fold increased risk of myocardial infarction with a haematocrit ≥49% compared with a haematocrit ≤44% — tracks Toss’s abstract closely enough that the abstract is the likely proximate source, in which case the number was altered rather than recalculated. Either route produces the same result: the receiving paper reports the association as weaker than the source does.

The magnitude is small and the direction is the point. Nothing in this instance inflates a risk, and it is recorded at the same weight as the instances that do.
***
## 6. Warny’s summary of Toss’s cohort compresses four decades into the age at measurement

The same sentence describes the cohort as young Swedish men aged 18–19. That is the age at which haematocrit was measured, and nothing in the sentence says so.

Toss measured haematocrit at Swedish conscription, which took place at 18–19 years of age, and then followed the cohort for a median of 36 years, to a range of 9–41 years. The 9322 first myocardial infarctions occurred at a median age of 51, across a range of 28–59. The paper’s own conclusion describes a follow-up period of more than three decades.

The sentence is not false. The study did include men aged 18–19; they were 18–19 when measured. What the compression removes is the interval between exposure and outcome, and with it the study’s actual claim — that a single adolescent measurement carries prognostic information across a working lifetime. Read as written, the sentence attributes myocardial infarctions to teenagers.

This is the trait-as-marker collapse recorded at finding 19, occurring inside a citation rather than inside an argument. The mechanism is the same: a measurement taken at one time is transposed onto the population as a standing characteristic, and the time structure that made the finding interesting is what gets dropped.

Direction is not straightforward here and is recorded as such. Compressing a 36-year follow-up to the age at baseline makes the association read as more immediate, which is the more striking reading; but it sits in the same sentence as an understated point estimate, which is the less striking one. A single sentence carrying errors in both directions is evidence against a directional account of this particular transmission, and is recorded that way rather than assigned to either column.
***
## Recorded and not entered

Toss did nothing wrong, and no entry is manufactured to balance the section. Its abstract gives 1.4-fold, its Table 2 model F gives 1.38 (1.29–1.48), and the two agree within rounding. The comparison wording is identical in both places. There is no Toss entry.

The clinical-utility disagreement between the two papers is real but narrower than a flat contradiction. Toss concludes that haematocrit may aid future risk assessment in young individuals and may complement other variables in identifying young people with substantial lifetime cardiovascular risk. Warny concludes that adding haematocrit is unlikely to improve risk prediction in myocardial infarction. These are different prediction problems — lifetime risk from an adolescent measurement, against incidental findings in a cohort of mean age 58 — and neither paper engages the other’s. Recorded so that §9.4.1’s aside is not written as a sharper conflict than the sources support.

Warny’s classification of the null studies is in tension with its own placement of Toss, and the tension is Warny’s, not a transmission error. Warny attributes non-detection partly to cohorts composed of younger individuals aged 20–65. Toss’s cohort was measured younger than any of them and detected a strong dose-dependent association. No entry: this is Warny reasoning about its own field, not Warny characterising a source.
***
## Population flag — mandatory, and it applies to this whole section

Toss is 417,099 male conscripts ⚠ (not designed as a transgender or GAHT cohort; no trans-masculine subgroup identified), haematocrit measured once in late adolescence, varying naturally. Warny is 108,521 white individuals of Danish descent, mean age 58. ⚠ Neither cohort was designed to include people on exogenous testosterone, and neither reports any, and neither speaks to erythrocytosis that is induced, sustained and monitored. Nothing in this section transfers to trans-masculine populations without the argument made in §9.5.

Both sides of every instance and every non-instance below were open in the step it was written, per R13.3a: Tsai [44] 10.1001/archinte.162.10.1182, Schreijer [47] 10.3324/haematol.2009.017285, and Braekkan [71] 10.3324/haematol.2009.008417, all three untruncated. Braekkan was read last, because it is the characterising side of both candidates this session was staged to test.

## Braekkan’s account of LITE is accurate in every particular that bears on a conclusion — including the one it is least obliged to get right

Braekkan’s Discussion describes LITE as a prospective cohort of subjects aged 45 and above, primarily investigating traditional cardiovascular risk factors, which found no increased risk of venous thromboembolism across increasing tertiles of either haematocrit or haemoglobin; and qualifies that finding by noting that the cut-off for the highest tertile was 43.5% for haematocrit and 14.5 g/dL for haemoglobin, and that further levels were not explored.

Against LITE: the age floor is correct, the study aim is LITE’s own title, the null is LITE’s own Results prose, and both cut-offs are Table 2’s top-band floors to the digit.

The last limb is the one worth dwelling on, because it is stronger than Braekkan claims. LITE did not merely decline to explore higher levels. Its Statistical Analysis section describes computing hazard ratios for the upper decile against the lowest tertile for five lipid variables — low- and high-density lipoprotein cholesterol, total cholesterol, triglycerides and lipoprotein(a). The technique for interrogating an extreme tail was in the paper and was not applied to haematocrit. A characterisation that could have understated this instead understated its own case.

No entry. Recorded because a ledger that reports only failed transmissions is reporting on its selection procedure rather than on its material.
***
## The sample size Braekkan gives for LITE is wrong, and it is recorded rather than entered because nothing rests on it

Braekkan gives LITE’s cohort as 19,923. LITE’s own figure is 19,293, appearing in the abstract, at the opening of the Results, and in Table 1’s overall row.

LITE’s figure reconciles from every direction: 14,594 in the Atherosclerosis Risk in Communities arm plus 4,699 in the Cardiovascular Health Study arm; 10,633 women plus 8,660 men; 14,356 white plus 4,937 black participants. The exclusion cascade closes exactly — 21,680 recruited across both parent cohorts, less 2,387 excluded, with the stated exclusion categories summing to 2,529 and the 142-person difference accounted for by participants meeting more than one criterion.

A digit transposition, on Braekkan’s side, bearing on no estimate and no inference. It is recorded and not entered because the ledger’s entries are material by construction and a sample size that changes nothing would pad it. It is recorded rather than dropped because five successive records carried 19,923 without resolving it, and the point at which a figure stops being checked is the point this section exists to identify.
***
## Two accounts of the same study are not one account at two removes — and the error is what demonstrates it

An earlier ruling held that the editorial’s characterisation of LITE reproduced Braekkan’s Discussion almost verbatim, and therefore that two apparent attestations were a single source transmitted twice.

Set side by side, the two texts share three tokens: the study’s name, the word cut-off, and the figure 43.5%. The editorial adds an evaluative judgement Braekkan does not make — that the cut-off was relatively low. Braekkan adds the sample size, the study’s stated aim, the haemoglobin cut-off, and the observation about unexplored levels.

The decisive test is the error. Braekkan’s sample size was available to be carried and Braekkan’s transposition was available to be inherited. The editorial carries neither — it gives no sample size at all, and cites Tsai directly, with authors, journal, year, volume, issue and page range complete and correct.

The non-independence claim is withdrawn. The route it proposed remains plausible — the editorial reviews Braekkan in the same issue, and its LITE citation sits in a paragraph whose function is to contextualise Braekkan — but plausible is not demonstrated, and this section does not enter what it cannot show.

⚠ Recorded against this review’s own record. The withdrawn ruling had already been corrected once, in the same direction: its first issue asserted that LITE’s 43.5% rested on no first-hand read, which was false. A claim corrected twice toward the same overstatement is a claim that was wanted.
***
## 7. An editorial converts a disclosed null interaction into a phenomenon requiring explanation

Braekkan’s Statistical analyses section states that there were no statistical interactions between the haematologic variables and gender, and that sex-specific analyses were nonetheless carried out for the sake of interest, because the normal ranges of haematocrit, haemoglobin and red blood cell count are generally higher in men.

The paper then reports throughout that risk estimates were higher for men than for women, and concludes that haemorheological factors might contribute more to venous thromboembolism risk in men than in women.

Against its own Table 4, for total venous thromboembolism in the uppermost band, the estimates are 1.51 (1.08–2.12) in women and 1.54 (1.08–2.21) in men. The directional claim is carried by the continuous multivariable estimates, where the association retains significance in men at 1.33 (1.05–1.70) and loses it in women at 1.11 (0.88–1.40), and by the unprovoked categorical analysis. It is not carried by the total categorical analysis, where the difference is three hundredths across near-identical intervals.

The editorial reviewing that paper closes with a list of open questions, the first of which asks why the association is more pronounced in men than in women.

The qualification is lost at the point of transfer. Braekkan discloses the null interaction, states the sex-stratified analyses were exploratory, and gives the reason. The editorial asks why the phenomenon exists — a question that presupposes it, and that arrives with none of the material a reader would need to doubt it.

The underlying move in the source is the familiar one: the difference between a significant estimate and a non-significant estimate read as a significant difference, in a paper whose own interaction test found none. The entry is not against that. Braekkan reports the null interaction plainly, in the section where it belongs, and an entry aimed at a paper for disclosing its own limitation would punish the candour this section elsewhere relies on. What is entered is the transmission, in which the disclosure does not survive.

⚠ Direction is not assigned. The editorial’s argument runs toward confounding as an explanation for the haematocrit association; a sex difference does nothing for that thesis, and the sentence sits in a closing list of questions rather than in the argument. A dropped qualification that serves no rhetorical purpose in the receiving text is evidence against a directional account of this transmission, and is recorded that way, as finding 19’s double duty was.
***
## Population flag — mandatory, and it governs everything above

Braekkan is 26,108 adults from a single Norwegian municipality, haematocrit measured once in 1994–95, varying naturally. LITE is 19,293 adults aged 45 and above pooled from two United States cohorts which themselves recruited from two of the same six communities. ⚠ Neither cohort was designed to include people on exogenous testosterone, and neither reports any. ⚠ Neither ascertained gender identity, so neither is describable as a cisgender cohort; what each recorded is sex. Sex-stratified findings in these populations describe naturally occurring sex differences in haematocrit distribution and cannot be read across to a population in which haematocrit is pharmacologically raised. Nothing in this section transfers without the argument made in §9.5.
***
## Cannarella — a threshold table and the paragraph that summarises it, disagreeing in opposite directions

Two losses sit in this paper, and they do not run the same way. One would bar treatment from almost everyone; the other makes the guideline literature look more permissive and more unanimous than it is. They are recorded together because a directional account of either one alone would be wrong.

⚠ This is the source this review leans on most heavily for the threshold range. R3.1 already operationalises the consequence: the range is cited to Supplementary Table 7 row by row, never to the Discussion.
***
### 8. A contraindication printed with its inequality reversed — as set, it bars testosterone from nearly everyone

Cannarella’s Supplementary Table 7 tabulates ten societies across nine rows, under a column headed Recommendation against TRT [6]. Eight rows carry a numeric haematocrit threshold. Seven of the eight are written with a greater-than sign. The Endocrine Society row is written Hematocrit<48%.

Read as printed, the Endocrine Society recommends against testosterone in men whose haematocrit is below 48% — which is most men with hypogonadism, and the near-inverse of the guideline’s actual position that the bar sits above that figure.

The number is right and the operator is wrong, which is the worst combination available. A reader checking the figure against any secondary account finds 48% and moves on; the digit is what gets verified, and the digit is not what is broken. Nothing else in the column is set this way, so there is no house convention to absorb it — the row is inconsistent with its own table.

⚠ Direction, stated with a caveat. Taken at face value the error is maximally restrictive, which is the treatment-sceptical direction. It is not plausibly a rhetorical move. It is a typesetting inversion in a supplementary table, and recording it as if it carried an argument would be reading intent into a printing defect. The direction of its face-value reading is recorded; no direction is assigned to the act.
***
### 9. The Discussion names two of the five thresholds in its own table, and cites the table while doing it

The Wider implications paragraph names ten societies and states that they recommend against starting testosterone in patients with elevated haematocrit and suggest monitoring to reduce or discontinue treatment if it exceeds 54%, with 50% according to AACE and ICSM — citing Supplementary Table 7 explicitly [6].

Against the table it cites, the eight numeric thresholds are: AUA >54 · BSSM >54 · EAU >54 · AACE >50 · ICSM >50 · ISAAM >52 · EAA >48–50 · ES <48. SIAMS/SIE states no number, directing instead an evaluation of global cardiovascular risk including haematocrit.

The sentence accounts for five of the eight rows and two of the five distinct values. The three rows it drops — the Endocrine Society’s 48, the EAA’s 48–50 and ISAAM’s 52 — are all below the 54% it names, and two of them are the lowest in the table. ⚠ An earlier form read the three lowest, which is arithmetically wrong: ISAAM at 52 is the fifth lowest, and AACE and ICSM at 50 are both lower and both named (correction 695).

The qualification lost is the spread. A reader of the prose takes away a consensus bar at 54%, softened to 50% by two bodies. A reader of the table finds a range running from 48% to 54%, with three societies setting a threshold below the figure the sentence leads with. The compression does not misreport any single society; it removes the fact that the societies disagree, and removes the disagreement asymmetrically, at the low end only.

⚠ Direction: AGAINST this review’s standing failure direction, and it is recorded here for exactly that reason. The compression makes the guideline literature look more permissive — a clinician citing the prose tolerates a haematocrit that three of these bodies would act on. That is the less alarming reading, not the more striking one. §6.1’s diagnosis does not fit this instance, and the entry says so rather than being quietly filed with the ones that do.

Amendment C in its purest form, and the register already says so: qualification lost inside a single paper, between its own data and its own prose. No second author is involved. The failure needs no transmission chain to occur.

⚠ TWO LIMBS ADDED AT PART 28, R29.8 AND R29.9, WITH BOTH SIDES OPEN.

R29.8 — the Discussion cites the table for something the table does not tabulate. The discontinuation claim is referred to Supplementary Table 7 explicitly, and Supplementary Table 7 has no discontinuation column at all. Its only action column is Recommendation against TRT. So the citation covers a proposition the table does not carry, independently of whether the numbers match.

R29.9 — the Endocrine Society row omits the 54% threshold its own guideline sets, and the altitude qualifier. Against Bhasin 2018 10.1210/jc.2018-00229 id 163 opened untruncated in the same step: the guideline gives >48%, or >50% for men living at higher altitudes, as a relative contraindication to T therapy, and Table 9 separately prescribes stopping therapy at >54% [7]. Supplementary Table 7 gives the Endocrine Society the 48% figure alone. ⚠ The consequence cuts toward the paper rather than against it: the Discussion’s discontinue above 54% claim is TRUE of the Endocrine Society, and cannot be verified from the table it points at.

⚠ A FOURTH LIMB WAS CHARGED AGAINST THIS PARAGRAPH AND HAS BEEN WITHDRAWN — see the closure of decision 3 below.
***
### Population flag — mandatory

Cannarella is a meta-analysis of hypogonadal men with total testosterone below 12 nmol l⁻¹, drawing on 4027 patients from randomised trials and 310 288 from observational studies. The societies tabulated write for that population. Every threshold in Supplementary Table 7 is a threshold for testosterone replacement in men with hypogonadism, and not one of them was set with masculinising hormone therapy in a trans population in view. Nothing in this block transfers without the argument made in §9.5.
***
## 10. Tatarian — a formal correction that moved an error instead of removing it, and a determinant list quietly rebuilt

Tatarian is a retrospective chart review of 511 transmasculine patients on testosterone for at least twelve months at a single United States centre; 113 of 511, 22%, experienced an episode of polycythaemia defined as haematocrit ≥50.0 [30]. That is the absolute base rate, and every ratio below sits on it (FM1).

The paper was corrected on 30 December 2022, four months after online publication on 3 August 2022. Six correction statements are printed in the article, and they are not of one kind.

The correction changed a number that was right, and left the number that was wrong

Tatarian subdivides the polycythaemia group by haematocrit band. Table 4, Figure 3’s legend, and the Results paragraph reporting the subdivision all agree: 65 patients at 50.0–51.9, 36 at 52.0–53.9, and 12 at ≥54.0.

The management-strategies paragraph disagrees with all three. It opens Of the 65 patients with a hematocrit ≥54.0 and, two sentences later, Of the 12 patients with a hematocrit of 50.0–51.9. The two outer denominators are swapped.

The paragraph’s own sub-counts settle it, and they settle it in the direction of the tables. The ≥54.0 sentence distributes 2 observed, 7 dose-adjusted, 1 route-switched and 2 referred — summing to 12. The 50.0–51.9 sentence distributes 46 observed, 16 dose-adjusted and 3 route-switched — summing to 65. The middle band’s 21, 12, 1 and 2 sum to 36 and its denominator reads 36. Only the two headline figures are wrong; the data behind them are right.

The correction is what put them there. The article’s own correction note attaches a marker to the 65 in the ≥54.0 sentence and states that it formerly read 12.

So the pre-publication text had one error and the corrected text has two. Before correction, the ≥54.0 denominator read 12 and was correct; the 50.0–51.9 denominator read 12 and was wrong. A corrector looking for a 12 that should have been a 65 found the wrong one. The error was not survived. It was relocated, and duplicated on the way.

⚠ Residual uncertainty, declared. The correction notes use a series of distinct dagger and section marks to bind each superseded value to its position in the text. The archive’s extraction flattens several of these glyphs into indistinguishable marks, so the binding above rests on the order of the correction list matching the order of the marked values in the body — which it does, across all five values in that note, including three that verify independently against Tables 2, 4 and 5. It does not rest on a distinct symbol. Settling it beyond order-matching needs the publisher’s typeset PDF. Recorded rather than smoothed over, because this limb runs toward the more striking reading and §6.1 makes that the direction requiring more evidence, not less.

The other five corrections were sound, and saying so is not a courtesy

Four values in the same note verify against the paper’s own tables once corrected: the BMI p-value moves to 0.0022 and Table 2 reads 0.0022; the OSA p-value moves to 0.034 and Table 2 reads 0.0336; a mean dose of 84.42 mg/week — the mean duration in months, slipped into the dose slot — moves to 100 and Table 4 reads 100; a serum testosterone of 873 moves to 860 and Table 5 reads 860. Each of those was a real defect and each was correctly repaired.

The ledger entry is against one correction, not against the correcting. A paper that issues six corrections and gets five right is not a careless paper. Recorded at the same weight as the failure, because a section that reports only the failed limb of a correction is reporting on its own selection procedure.

The Discussion’s determinant list was rebuilt, and R2.7 records only the part of the rebuild that makes the sharper story

R2.7 states that the December 2022 correction removed intramuscular testosterone injections from the determinant list. That is true and it is a quarter of what happened. ⚠ An earlier version of this sentence read a third, against this paragraph’s own count of four. Correction 216. The correction note quotes the superseded sentence in full: it formerly ended nicotine dependence, OSA, oophorectomy, and intramuscular testosterone injections all appeared to be important determinants for the development of polycythemia. The sentence now reads Age, BMI, testosterone dose, OSA, and oophorectomy all appeared to be important determinants for the development of polycythemia in univariate analyses.

Set against each other, the correction did four things: it removed intramuscular injections, it also removed nicotine dependence, it added age, BMI and dose, and it added the qualifier in univariate analyses.

The qualifier is the substantial change, and it is the one the record dropped. Without it the sentence asserts a set of determinants flatly; with it the sentence is scoped to the analysis that found them, which matters because Table 6’s multivariable regression retains only age (OR 1.038, 1.010–1.067, p = 0.008), BMI (1.030, 1.001–1.060, p = 0.045) and dose (1.011, 1.002–1.020, p = 0.012), and drops OSA to 0.971 (0.426–2.214, p = 0.945) and oophorectomy to 1.582 (0.974–2.568, p = 0.064). The correction brought the Discussion into line with the paper’s own Table 6.

⚠ The withdrawn intramuscular claim contradicted Table 2 on the day it was printed. Table 2’s route-of-administration row gives 30.97% of the polycythaemia group on intramuscular testosterone against 35.93% of the non-polycythaemia group, p = 0.2289 — non-significant, and pointing the opposite way. Tatarian’s Introduction states the received position that intramuscular routes carry higher risk; Tatarian’s own data did not reproduce it, and the corrected Discussion stopped saying they did.

⚠ Direction, against the review’s own record. R2.7 preserved the intramuscular limb — the limb that best fits an argument about route and erythrocytosis — and dropped the nicotine removal and the univariate qualifier. That is §6.1’s mechanism operating on this review’s record of another paper’s correction: the striking fragment retained, the scope-narrowing fragment lost. Recommended for §6.2 as a failure-log row, not smuggled into the source’s entry.

The propagation limb is NOT entered, because no citer has been opened

R2.7’s pre-correction citers propagate a withdrawn claim is a claim about transmission, and transmission needs a receiving paper open. None has been. Under R13.3a this is a suspicion, not an entry.

It is testable within this sitting. Krishnamurthy id 137 and Okano id 118 are both held; Krishnamurthy is opened later in this sitting for its own limbs, and its reference list and any Tatarian sentence can be read then. Routed forward, deliberately, rather than entered on plausibility.

Population flag — and this is the one source in the block that does not need the §9.5 bridge

⚠ Tatarian is 511 transmasculine patients on gender-affirming testosterone. Unlike Cannarella, Warny, Toss, Braekkan, LITE and ARIC, this cohort is the review’s own population. The flag is recorded in the opposite direction from every other flag in this section: nothing here needs to be carried across from a cohort outside gender-affirming care. What limits it instead is design — single-centre, retrospective, chart-derived, with missing data on 12 serum testosterones in the polycythaemia group and 139 in the comparison group, and no adjustment for the ascertainment differences a chart review introduces.

11. Antun on Defreyne — the numbers transmit exactly, the cohort and the follow-up do not

R3.4 recorded this as a clean transmission, held for symmetry on the ground that a claim about qualification being lost at predictable points is only credible if the clean cases are counted too. That ground is sound and the entry stays. What changes is that the transmission is clean in one place and not in two others, and only the clean place had been checked.

The threshold counts transmit exactly — the non-instance is real

Antun’s Discussion reports that in the ENIGI study 22 of 192 transmasculine participants receiving testosterone developed erythrocytosis on a haematocrit cutoff of 50%, and four met the criteria at 52% [31].

Defreyne’s Results: 22 of 192 trans men (11.5%) developed serum haematocrit ≥50.0% across the investigation period — 10 of 109 on testosterone undecanoate (9.2%), seven of 44 on esters (15.9%), five of 39 on gel (12.8%) — and four developed ≥52.0%, three on esters (6.8%) and one on gel (2.6%) [22].

Both numerators and the denominator carry across without distortion. Recorded as verified, and it weighs equally with an instance.

One sub-limb examined and declined. Antun writes developed erythrocytosis, where Defreyne devotes a paragraph to distinguishing absolute erythrocytosis — a red cell mass above 125% of predicted — from a merely raised haematocrit, warning that haematocrit does not discriminate absolute from apparent erythrocytosis. The tempting entry is that Antun collapsed that distinction. ⚠ It does not hold: Antun’s sentence attaches the operational definition to the label in the same breath — erythrocytosis based on the Hct cutoff of 50% — so the definition stays visible to the reader. Declined as an instance. Recorded because it ran toward the more striking reading and did not survive the check.

The inequality shifts — and the shift does not originate with Antun

Antun gives the cutoffs as >50% and >52%. Defreyne’s Results give ≥50.0% and ≥52.0%.

The distinction is load-bearing in this paper, not pedantic. Defreyne records that at three months eight trans men (4.3%) had haematocrit ≥50.0% with a range of 50.0–51.0% — the range opens exactly at the boundary, so at least one participant sits at 50.0 and is counted by ≥ and excluded by >.

⚠ But Antun did not introduce it. Defreyne is internally inconsistent on the operator. Its Results prose uses ≥50.0% and ≥52.0%; its Table 4 title uses >50.0% and >52.0%; its Table 4 sub-headings revert to ≥; and its Discussion and Conclusions use >50 Hct % and >52 Hct % throughout. Antun’s operator matches Defreyne’s Discussion and table title and mismatches Defreyne’s Results.

Under Amendment C the Discussion is a secondary source even inside Defreyne — so a reader taking the summarised rates from where a reader would naturally take them finds >50% and transmits it faithfully. Not entered against Antun. Entered instead as a new intra-Defreyne candidate, below.

The cohort size does not transmit — 285 arrives as 265

Antun’s Discussion introduces the ENIGI study as pertaining to 265 transmasculine and 340 transfeminine individuals [31].

Defreyne: the data lock analysed 625 participants: 285 trans men and 340 trans women [22].

The pairing identifies which line Antun was reading, and the arithmetic settles it. 340 is Defreyne’s data-lock figure for trans women, and it pairs with 285, not 265. 285 + 340 = 625, Defreyne’s stated total. 265 + 340 = 605, which is no figure in Defreyne at all. The transfeminine number transmits and the transmasculine number loses twenty.

⚠ Declared residual. Defreyne’s Figure 1 is a flowchart and its internal numbers are not present in the archive’s text extraction, so 265 cannot be excluded from appearing somewhere inside it. What is excluded is that 265 pairs with 340 in Defreyne’s prose, and that any pair beginning 265 sums to 625. Settling the flowchart needs the typeset figure. Recorded rather than smoothed, because this limb converts a recorded non-instance into an instance and therefore runs toward the more striking reading.

The follow-up structure does not transmit, and Defreyne’s own limitation goes with it

Antun states that the ENIGI participants were followed for 36 months and examined every 3 months [31].

Defreyne’s Methods: follow-up ran baseline, 3, 6, 9 and 12 months in Amsterdam and baseline, 3, 6, 9, 12, 18, 24 and 36 months in Ghent [22].

Two things are wrong and a third is missing. Thirty-six months describes one site, not the cohort. Every three months describes neither site past the first year, since even Ghent moves to six- and twelve-month intervals after month twelve. And Defreyne’s Limitations state plainly that Amsterdam follow-up covered only the first year, leading to a decrease in sample size and power in the analyses at 18, 24 and 36 months — the qualification that tells a reader how much weight the late timepoints can bear.

This is the ledger’s central mechanism, and it is the plainest example of it in the section. Nothing is misreported as a number. A two-site design with unequal follow-up becomes a uniform one; a stated power limitation at the three longest timepoints disappears; and the paper that inherits the sentence has no way to know either happened. ⚠ Direction: toward a cleaner, stronger-looking source than Defreyne claimed to be — which is the standing direction, and it is the first instance in this sitting that fits §6.1 without qualification.

A minor limb, recorded at its real weight

Antun adds that after twelve months the transmasculine average haematocrit rose from 41% to 46% and remained relatively unchanged thereafter. The rise is exact against Defreyne’s Table 2: 41.0 [39.0–42.6] at baseline, 46.0 [44.0–47.0] at twelve months. The tail is not. Defreyne reports a 1 Hct % decrease at 18 months, 95% CI 0.51–2.30, p < 0.001 — statistically significant — before finding no significant differences between 12, 24 and 36 months. A significant decrease becomes relatively unchanged. Small, real, and recorded as minor rather than promoted.

New candidate opened against Defreyne, not against Antun

⚠ Defreyne’s operator is inconsistent across its own Results, table title, table sub-headings, Discussion and Conclusions, at a boundary where the inequality changes the count — its own three-month range opening at 50.0 proves it. Both sides of this are inside Defreyne and it is therefore rulable from the read just performed. ⚠ It is not entered here, because R3.4’s entry is about Antun and folding a Defreyne-internal defect into it would misattribute the failure. Recommended as a separate intra-source instance in the marked internal subsection, alongside R18.2 and R21.5.

Population and base-rate flags

Both papers are trans cohorts — the review’s own population. No §9.5 bridge is needed and none should be implied.

⚠ FM1, and it bites here because Antun is where the fold-changes in this literature come from. Antun’s headline ratios must never travel without their absolute rates: 7.4 (95% CI 4.1–13.4) against cisgender male referents rests on 26 events, 7.5 per 100 000 person-days, 26 of 398 participants (7%) at a 52% cutoff; 83.1 (95% CI 36.1–191.2) against cisgender female referents rests on 91 events, 39.3 per 100 000 person-days, 91 of 311 participants (29%) at a 48% cutoff. The 83-fold is an artefact of applying a female haematocrit threshold to people taking testosterone — which is Antun’s own conclusion, that parameters be read against affirmed gender rather than sex recorded at birth.

Krishnamurthy on Antun — the fold-changes transmit exactly, and the frame that makes them interpretable does not

12. The fold-changes are Antun’s own words, and Krishnamurthy is exonerated on them

Krishnamurthy’s Introduction reports that trans men have an erythrocytosis rate 7 times higher (95% CI 4.1–13.4) than cisgender men and 83 times higher (95% CI 36.1–191.2) than cisgender women, citing Antun [21].

Antun’s Abstract reports that transmasculine participants experienced a 7-fold higher rate (95% CI: 4.1–13.4) relative to matched cisgender males and an 83-fold higher rate (95% CI: 36.1–191.2) compared with cisgender females [31]. Antun’s Table 2 gives the underlying Model 2 estimates as 7.4 (4.1, 13.4) and 83.1 (36.1, 191.2) [31].

Both intervals carry across exactly. The whole numbers are Antun’s, not Krishnamurthy’s — the rounding of 7.4 to 7 and 83.1 to 83 is performed in Antun’s own summary of its own results, and Krishnamurthy matches the Abstract rather than the table.

Under Amendment C an abstract is a secondary characterisation even when it sits inside a primary paper, and R8.3’s treatment of the editorial rests on the same principle. A citing author reading Antun’s summary finds 7-fold and 83-fold with no cutoff attached and transmits them faithfully. Not Krishnamurthy’s loss.

⚠ Direction: AGAINST this review’s standing failure direction. This was the limb the ledger seed pointed at most firmly and the one R27.20 held over for exactly this session. Opening both papers dissolved it. Recorded as verified, and a verified non-instance weighs equally with an instance.

⚠ F26.5 observed. A non-instance is a positive claim about a passage, so the passage was checked and not only the two figures it carries — which is how the two limbs below were found.

Two different outcome definitions are presented as two referents, and neither paper discloses it

Antun’s Methods apply a haematocrit cutoff of above 52% in the analyses comparing transgender participants with cisgender male referents and above 48% in the analyses using cisgender female referents. Table 2’s footnote states the same [31].

The 7-fold and the 83-fold are therefore not one outcome measured against two referent groups. They are two different definitions of erythrocytosis. Antun’s Abstract states both ratios in a single sentence with neither cutoff named, and Krishnamurthy’s sentence inherits that structure.

This is what makes the 83-fold uninterpretable as a risk statement. A reader takes away that trans men develop erythrocytosis eighty-three times as often as cisgender women; what the figure records is that applying a female haematocrit threshold to people taking testosterone classifies most of them as abnormal. Antun says so directly — its Conclusions recommend that haematological parameters be interpreted against affirmed gender rather than sex documented at birth.

⚠ The loss originates inside Antun and is inherited, not created, by Krishnamurthy. It is entered against Antun as an intra-source instance below. It is not entered here, because folding it into a Krishnamurthy entry would misattribute the failure — the same reasoning R27.19 applied to Defreyne.

13. What Krishnamurthy does lose — the conclusion that tells a reader which reference range applies

Antun states the affirmed-gender recommendation twice in its own voice — in the Abstract’s Conclusions and in the paper’s closing paragraph — and endorses it a third time in quoting the reference-interval study it relies on [31].

Krishnamurthy carries the two ratios into a paragraph surveying variable incidence and carries none of it. The sentence stands as a bare magnitude, and the reader is given no indication that the larger of the two numbers is an artefact of the threshold rather than a measure of harm.

This is the ledger’s central mechanism, and it is the one limb in this block that is unambiguously Krishnamurthy’s. Nothing is misreported. A ratio travels and the sentence that made it interpretable does not.

⚠ Amendment C cuts against the citing author here, and the entry says so. On the fold-changes Amendment C exonerated Krishnamurthy, because the compression was Antun’s. On this limb it does not, because the qualification Krishnamurthy dropped was in Antun’s Abstract — the same passage the ratios were taken from, four sentences down. It was not buried in a Discussion a citing author might reasonably not reach.

⚠ Direction: toward the more striking finding. The standing direction.

14. A cohort becomes a population

Antun writes that transmasculine study participants experienced the elevated rates, and specifies that the referents were matched [31]. Krishnamurthy writes that transgender men have an erythrocytosis rate seven and eighty-three times higher, and drops the matching [21].

A rate observed in one cohort of 424 becomes a property of a population. Small against the limb above, and recorded as minor rather than promoted. ⚠ Direction: toward the more striking finding.

FM1 — and the referent side of the 83-fold has not previously been recorded in this review

FM1 is in force and it binds hardest here, because Antun is where the fold-changes in this literature come from. Neither Antun’s Abstract nor Krishnamurthy carries an absolute rate of any kind. Antun’s Table 2 carries all of them [31]:

• Against cisgender male referents, cutoff above 52%: transmasculine 26 events, 7.5 per 100 000 person-days; referent males 17 events, 1.4 per 100 000 person-days; HR 5.5 (3.3, 9.2) in Model 1 and 7.4 (4.1, 13.4) in Model 2.
• Against cisgender female referents, cutoff above 48%: transmasculine 91 events, 39.3 per 100 000 person-days; referent females 6 events, 0.39 per 100 000 person-days; HR 81.4 (41.1, 161.3) in Model 1 and 83.1 (36.1, 191.2) in Model 2.
• Antun’s Discussion gives the participant-level denominators: 26 of 398 (7%) crossed 52%, and 91 of 311 (29%) crossed 48%, each restricted to participants never above the cutoff before hormone therapy.

⚠ The 83-fold rests on six events in its comparison group. That is the fact the fold-change conceals and the reason the interval runs from 36.1 to 191.2. The referent-side counts and rates are recorded in this review for the first time here.

⚠ Direction: against the more striking reading. Establishing the denominator makes the review’s strongest-looking figure visibly fragile, and it is recorded for that reason.

Model reporting, examined and not entered. Antun’s Abstract reports the Model 2 pair, which is the higher of the two on both comparisons, without naming the model. Model 2 is the pre-specified fuller adjustment and reporting it is defensible. Noted, not entered.

Population and design flags — mandatory

Both papers are trans cohorts, and neither needs the §9.5 bridge. Krishnamurthy is 6670 trans masculine and gender-diverse users of testosterone through a United States telehealth service, mean age 27 (range 18–65). Antun is 424 transmasculine and 559 transfeminine members of the STRONG cohort, drawn from three integrated health systems and matched to cisgender referents on birth year, race or ethnicity, site and enrolment year.

What limits them is design, and the two designs fail in opposite directions. Krishnamurthy is cross-sectional, and only the most recent laboratory value per patient was included — a single-draw point prevalence. Antun is longitudinal but built on opportunistic electronic-health-record bloodwork, not collected at set intervals, which Antun addresses with inverse-probability-of-testing weights and names as a limitation.

⚠ The two figures most likely to be set against each other in this literature are not the same quantity. Krishnamurthy’s 8.4% is a prevalence on one blood draw; the longitudinal cohorts report cumulative incidence over years of follow-up. Neither number may appear in this review beside the other without that clause inline.

15. Item 57 — the Tatarian propagation limb does not exist in Krishnamurthy

Krishnamurthy’s reference list runs to 32 entries and contains no Tatarian. No sentence in the paper carries Tatarian’s withdrawn intramuscular-injections determinant. Krishnamurthy’s own finding that intramuscular users have higher mean haematocrit than transdermal users (44.96% against 43.41%) is derived from its own cohort, and its citations for the formulation-stratified literature are DeLoughery, Nolan, Perez-Luis and Strange.

⚠ A clean negative, and it runs against this review’s own case. R2.7’s propagation limb was routed to this sitting at R27.11 precisely so it would not be entered on plausibility. Checked, and absent.

⚠ Verified against the reference list itself, not against a search of it.

Limbs examined and declined, recorded at equal weight

Declined 1 — the design-versus-size attribution. Krishnamurthy’s Discussion attributes the gap between its own low prevalence and the higher figures in prior studies to the size of its cohort. The design difference is the better explanation, and the tempting entry is that Krishnamurthy substituted one for the other. ⚠ It does not hold: the same paper’s Limitations concede the cross-sectional design, name the loss of over-time observation, and raise regression to the mean explicitly. The exonerating material is in the source. Declined. It ran toward the more striking reading and did not survive the check.

Declined 2 — the 54% operator. Krishnamurthy’s Abstract sets the top band at ≥54%; its Results, its Discussion and its Madsen paragraph set it at >54%, with at or above 54% one sentence later. This is R27.19’s shape — an operator inconsistent across a paper’s own sections at a threshold. ⚠ But it is not R27.19’s strength: Krishnamurthy reports no boundary case, so it cannot be shown that the operator changes the count. Defreyne’s own 50.0–51.0 range is what made that entry rulable and Krishnamurthy has no equivalent. Declared and not promoted; no direction assigned.

Attempted, having become attemptable

⚠⚠ THE NOLAN LIMBS ARE NOW ATTEMPTED, BECAUSE NOLAN HAS BEEN OBTAINED AND READ. Nolan 10.1111/imj.14839 was filed to the archive on 4 September 2026 and read first-hand and untruncated in the same step this was written. ⚠ The statement this paragraph replaced — that Nolan was verified missing from the archive and abstract-only by design — was true when written and is now false. Correction 163.

Krishnamurthy’s reference 19 is the Nolan work and carries the formulation ranking as 25% on intramuscular enanthate against 16.7% on undecanoate and no cases on transdermal. ⚠⚠ NOLAN’S OWN RESULTS SECTION REPORTS 23.3%, 15% AND 0%. ⚠⚠ AND THE DISCREPANCY IS NOT A TABULATION ERROR. Nolan’s own Conclusions state the same result as one in four and one in six. One in four is 25.0%. One in six is 16.7%. Krishnamurthy read Nolan’s conclusions rather than Nolan’s results, and a rhetorical rounding acquired decimal precision in transmission. ⚠ The false precision is manufactured downstream; the rounding is the source’s own.

⚠⚠ THIS IS AN AMENDMENT C INSTANCE AND IT IS RECORDED AS ONE. The conclusions and abstract of a primary paper are secondary, and here a citing paper’s fidelity to the conclusions is precisely what introduced the error. It is not entered as a defect in Krishnamurthy, which transmitted its source accurately.

⚠ Denominators, because they are not the group sizes. Nolan’s groups are 125 undecanoate, 31 enantate and 24 transdermal, totalling 180. The haematocrit analysis runs on 122, 30 and 24 — 176 — which reconciles exactly against the 73 of 249 records declared to be missing a haematocrit. 23.3% is 7/30.

⚠ A restriction on what may be cited from this copy, declared rather than left implicit. The document held is the author accepted manuscript deposited in an institutional repository, and it states of itself that it has not been copyedited, typeset, paginated or proofread. Its cover-sheet citation block carries no volume, issue or page numbers, and at least two of its in-text citation numbers do not resolve to plausible targets. ⚠⚠ NO BIBLIOGRAPHIC DETAIL AND NO CITATION-ATTRIBUTION CLAIM ABOUT NOLAN IS WRITTEN FROM THIS COPY. Both require the version of record, which this review has not obtained.

⚠ One further retrieval fact, recorded and not resolved. Nolan states that this cohort’s clinical characteristics were previously published, and the citation given does not resolve. A prior publication of the cohort therefore exists and is unidentified from this source, which bears on the overlap map and is flagged there rather than guessed at here.

Recorded as facts about Nolan and about Krishnamurthy’s reference list — §9 #21, and R16.3’s guard against a suspicion hardening by repetition.

⚠ One further Krishnamurthy claim is unreachable and is declared rather than pursued. Krishnamurthy states that the haematocrit threshold at which mortality risk begins to rise is lower for those recorded female at birth, citing Boffetta and Paul. Neither is held. The claim sits in tension with Antun’s affirmed-gender conclusion, and that tension cannot be adjudicated from the archive. Declared; no entry.

Krishnamurthy on Madsen — every prevalence figure transmits, and a study that never measured thromboembolism is reported as having found none

16. The prevalence figures and the timing finding transmit exactly

Krishnamurthy reports that Madsen’s cohort of 1073 transgender men receiving care in Amsterdam found 24% with a single haematocrit measurement above 50%, 11% with erythrocytosis on at least two laboratory assessments, and 2.2% with at least one value above 54%, with the largest increase in the first year of therapy [21].

Madsen: 1073 trans men were included; a single measurement above 0.50 L/L occurred in 24.0% and 11.1% had it twice; values above 0.54 L/L were seen in 2.2% once and 0.5% twice; and the largest increase in haematocrit was seen in the first year after initiation [28].

Four figures, a cohort size, a setting and a timing finding, all clean. ⚠ One thing is not carried: the operator. Krishnamurthy writes greater than 54% here and at or above 54% two sentences later, for the same Madsen stratum, and Madsen’s own form is >0.54 L/L. The figures transmit exactly; the criterion does not, and the drifted form is the one carrying the subgroup null recorded below (correction 507).

⚠ And the once-versus-twice distinction survives, which is the part most easily lost. Madsen defines erythrocytosis as exceeding 0.50 L/L twice, and reports single-measurement and repeat-measurement prevalences separately at every threshold. Krishnamurthy uses the twice figure for the 11% and the once figure for the 2.2% and labels each correctly. A citing author who mixed them would have produced a figure no reader could reconstruct. Recorded because a verified non-instance weighs equally with an instance, and because this one is more than arithmetic — it is the definitional structure carrying across intact.

⚠ Direction: against this review’s standing direction. The transmission is better than the ledger expected.

17. A study that never measured thromboembolism is reported as having found none of it

Krishnamurthy: The authors did not note any thromboembolic events, even in the group with hematocrit levels at or above 54% [21].

Madsen’s stated aims are two: the prevalence and determinants of erythrocytosis, and the time relation between duration of testosterone therapy and haematocrit. Its outcome variable throughout is haematocrit crossing a threshold. Thromboembolic events are never an outcome, never ascertained, never counted and never reported.

Where thromboembolism appears in Madsen, it appears as background and as a gap: the Context names it as what erythrocytosis may increase the risk of; the Introduction cites the general-population associations, including the Tromsø unprovoked-venous-thromboembolism figure; and the Discussion names the risk of thromboembolic events with different cutoff values as a subject for further research [28].

There is no null result here to transmit. There is no result. Did not note any describes a study that looked and found nothing; Madsen did not look. Absence of measurement is reported as absence of events — the fourth of this review’s recurring error patterns, and the one Amendment D exists to prevent this review from committing itself.

⚠⚠ AND A FIFTH PATTERN IS RECORDED HERE: OPERATOR DRIFT. It is listed by instance, with the two passages compared and a verification status for each, and NO AGGREGATE COUNT IS GIVEN, because the counting unit — source, passage pair, or affected claim — has not been fixed (correction 545). A threshold is stated with one operator in one place and another in another, for the same quantity, and the drifted form travels. ⚠ [20] states its criterion as ≥52 in its conclusions, >52 in its methods and results, and pairs >52 with <52 in a figure caption, leaving exactly 52 in neither arm. [21] writes greater than 54% and at or above 54% two sentences apart for one stratum of [28], and reports its own sixty patients at ≥54% in its abstract and >54% in its results. And this review’s own §9.2.2 metric table carried operator-bearing column headers over rows whose sources used different operators (correction 514). ⚠⚠ THESE INSTANCES INCLUDE INCONSISTENCIES WITHIN SOURCE REPORTS AND AN OPERATOR MISMATCH INTRODUCED BY THIS REVIEW’S OWN TABLE. They are recorded separately, and no aggregate count is reported until the counting unit is defined (correction 562). ⚠ Automated screening can identify candidate symbol differences; sweep_operators.py does that. Human review must establish whether the passages describe the same quantity and whether the difference is substantive, because only the source can say which form is right (corrections 515, 562).

And the stratum the null is localised to is the one Madsen declares unanalysable

⚠ Krishnamurthy does not merely assert the absence. It strengthens it — even in the group with hematocrit levels at or above 54% — which asserts a subgroup finding.

Madsen’s Limitations state that the group size of trans men with haematocrit above 0.54 L/L was not sufficient to perform analysis on [28]. The 0.54 stratum is 2.2% of 1073 on a single measurement and 0.5% twice.

The one subgroup Krishnamurthy names is the one Madsen names as too small to analyse. ⚠ This is what raises the limb above a loose did not note. A reader is given a stratified null in a stratum with no analysis behind it, and Madsen says so in terms, in the same paper, under a heading announcing its limitations.

⚠ DIRECTION: AGAINST §6.1’s STANDING DIRECTION, and the entry says so rather than being filed with the ones that fit. The error makes testosterone look safer, not more dangerous. It is aligned with Krishnamurthy’s own thesis — a paper titled Erythrocytosis Is Rare With Exogenous Testosterone in Gender-Affirming Hormone Therapy whose abstract hopes to allay concerns and whose conclusion hopes to temper fears. §6.1’s mechanism fits exactly — optimisation for the compelling finding — and here the compelling finding is the reassuring one.

18. Madsen’s headline time-relation finding does not travel, and it is the one that cuts the other way

Madsen’s second stated aim produces a cumulative-incidence curve: the cumulative risk of exceeding 0.50 L/L is 8% after one year, 38% after ten years, and 50% at the end of follow-up at fourteen years, in a subgroup of 776 participants restricted to those with baseline measurements [28].

Madsen’s own recommendation follows from it: the probability of first-time erythrocytosis keeps rising, a person stable on testosterone for years can still develop it, and regular control of haematocrit is warranted as long as people use testosterone.

Krishnamurthy carries the prevalences and the first-year timing finding. It does not carry the 50%.

⚠ An introduction need not report everything, and this entry does not rest on selection alone. It rests on what the selection is later used for. Krishnamurthy’s Discussion makes a comparative claim — that prior studies found a more impressive effect than its own, which it attributes to the size of its cohort. The single most directly comparable figure in the paper it is comparing itself against is the one omitted.

Madsen’s 24% single-measurement prevalence and its 50% fourteen-year cumulative incidence are the same cohort answering two different questions. Only the smaller number travels.

⚠ Direction: against §6.1’s standing direction, toward the reassuring reading. ⚠ This is the strongest limb in the block, and it is the one the register did not anticipate at all.

Population and design flags — mandatory

Madsen is 1073 trans men on gender-affirming testosterone and is the review’s own population. No §9.5 bridge is required and none should be implied. Median age at start 22.5 years, mean body mass index 24.5, 38% reporting tobacco use, 8.6% with a condition associated with erythrocytosis. Drawn from the Amsterdam Cohort of Gender Dysphoria, 6793 people seen between 1972 and 2015, with laboratory data from 2004 to 2018.

What limits it is design, and Madsen states each limit itself. Data are retrospective and chart-derived. 894 of 2398 trans men in the source database were excluded for lack of laboratory results and 431 for a missing start date — so the analysed cohort is under half the available one. Measurement frequency varied by person, which prevented any analysis of how long high haematocrit persisted. The route of administration was unknown for 2120 of 5528 haematocrit measurements, and the unknown-route category itself carries an adjusted odds ratio of 1.9 (1.3–2.8) for exceeding 0.50 L/L — so the route findings sit on incomplete exposure data, which Madsen names.

⚠ The centre’s own upper limit of normal for men is 0.50 L/L, which is the threshold the prevalence figures are built on. A prevalence at a locally defined limit is not transferable to a review that treats the threshold as the open question — §9.2 and §9.3’s problem, flagged here rather than assumed away.

Limbs examined and declined, recorded at equal weight

Declined 1 — a single measurement. Krishnamurthy’s phrasing could be read as exactly one rather than at least one. ⚠ It does not hold: the phrasing is Madsen’s own — A single measurement of hematocrit > 0.50 L/L occurred in 24.0% of trans men. Inherited, not introduced. Amendment C’s exonerating limb, for the third time this sitting.

Declined 2 — the unit conversion. Madsen reports throughout in L/L and Krishnamurthy in per cent. Numerically identical and no qualification is lost. Recorded so it cannot be raised later as unexamined.

19. Krishnamurthy on the phlebotomy threshold — three cited sources, two of them open and checked, and neither states it

Krishnamurthy attaches to the 54% threshold the clause the level at which it is recommended that phlebotomy be initiated, citing three sources together [21]. ⚠ The three are verified from Krishnamurthy’s own reference list, read first-hand and not carried from the register: reference 13 Bhasin 2018; reference 14 Dohle GR, Arver S, Bettocchi C, Jones TH, Kliesch S, Punab M. EAU Guidelines on male hypogonadism. Eur Assoc Urol. 2016;13:33; reference 15 Madsen 2021 [21].

The entry has three limbs and they do not carry equal weight. Two are ruled. The third cannot be checked at all, and the entry says so rather than assuming it.

Limb one — Madsen does not state it

Madsen’s Introduction reports the guideline as advising a cessation of testosterone therapy, a dose reduction, or therapeutic phlebotomy when haematocrit exceeds 0.54 L/L — three options. Its Discussion and its Limitations each restate the same threshold as testosterone therapy being discontinued until levels normalise, with no mention of phlebotomy. And Madsen’s own Recommendations for Clinical Practice contain no phlebotomy advice at all — for the 0.50 to 0.54 band it advises considering a transdermal switch, weight loss above a body mass index of 25, smoking cessation, and treatment of chronic lung disease or sleep apnoea [28].

One of three options is selected and printed as the recommendation, and Madsen’s own two restatements select a different one. ⚠ The charge is precise and is not that Madsen is the authority on the guideline’s content. Under Amendment C Madsen is a secondary source for guideline content. The charge is that Krishnamurthy cited Madsen for a proposition Madsen does not state. That is rulable with both open, and it is ruled.

Limb two — Bhasin 2018 does not state it either, and its formal recommendation names three actions of which phlebotomy is not one

Bhasin 2018 10.1210/jc.2018-00229 id 163, gate-checked and opened untruncated in the step every figure here was written, 30 July 2026 13:48:08 UTC, truncated: false, 156,695 characters, the whole guideline including its reference list.

The number is right. The guideline defines testosterone-induced erythrocytosis as haematocrit >54% [7].

The action is not. ⚠ The word phlebotomy occurs exactly once in the entire document — counted across all 156,695 characters, not judged by impression. It sits in the Erythrocytosis evidence narrative, and it sits after the guideline’s own instruction: clinicians should evaluate men who develop erythrocytosis, withhold testosterone until haematocrit has returned to the normal range, and then resume at a lower dose. Phlebotomy follows as an alternative that is also effective in managing the condition [7]. An efficacy statement about a second option, attached to no threshold.

⚠ And the guideline’s formal recommendation contains no phlebotomy at all. Table 9, Monitoring Men Receiving T Therapy, gives the haematocrit row as: check at baseline, 3–6 months, then annually; if haematocrit is >54%, stop therapy until it decreases to a safe level, evaluate the patient for hypoxia and sleep apnoea, and reinitiate therapy at a reduced dose [7]. Three prescribed actions. Phlebotomy is not among them.

The same shape as limb one, in a different source. One of two management options is printed as the recommendation, and the document’s own formal recommendation names a different one — three of them, in fact, and not this one.

⚠ The charge is precise. It is not that Bhasin 2018 forbids phlebotomy; the guideline says plainly that it works. The charge is that a clause presenting 54% as the level at which phlebotomy is recommended to be initiated is not supported by the source cited for it.

⚠ A finding from the same passage that runs the other way, recorded at equal weight. The guideline states in its own voice that the haematocrit level at which the risk of neuro-occlusive or cardiovascular events increases is not known, and that neuro-occlusive events among trial participants who developed erythrocytosis have been very infrequent [7]. This strengthens the review’s own position and is recorded beside the finding that does not.

⚠ Residual, closed rather than declared. The archive’s extraction of this document contains no >, <, ≥ or ≤ character anywhere; four ASCII substitutes carry them and the mapping is established by positive control — . → >, , → <, $ → ≥, # → ≤ — with unambiguous controls at men $65 years of age, haemoglobin to $17.5 g/dL and PSA level #4.0 ng per milliliter. So every operator above is >, not ≥, decoded from a calibrated mapping rather than read off a glyph. The calibration holds for this document’s extraction and licenses nothing about any other.

Limb three — Dohle cannot be checked at all, and that is stated rather than assumed

⚠ This limb is not ruled and is not counted. It is recorded because an entry that charged all three sources without checking the third would overclaim.

Krishnamurthy’s reference 14 carries no DOI [21]. source_has therefore cannot be run against it — the source is structurally ungateable, not merely absent from the archive. source_search on Dohle, run 30 July 2026 at 14:44 UTC, returns two documents and neither is the guideline.

⚠ That is a WEAK negative and the scope is stated accordingly. What is established is that no held document is identifiable as the Dohle guideline by this search, and no identifier exists by which to gate it. ⚠ It is not established that the archive does not hold it.

⚠ A weak negative is not a closure, and if Dohle does recommend phlebotomy at 54% then Krishnamurthy’s clause is substantially supported by one of its three citations and this entry’s wording changes.

What this entry claims, and what it does not

It claims: that Krishnamurthy cites three sources for a proposition, that two of the three are held, were opened untruncated, and neither states it, and that in both cases the source’s own formal recommendation names a different action.

⚠ It does not claim that none of the three states it. The third cannot be opened, and the entry carries that hole visibly rather than closing it by assertion.

Direction

⚠ Toward the more striking finding, and §6.1’s standing direction fits this one. Phlebotomy-at-54% reads as a more interventional and more alarming standard of care than stop, look for hypoxia and apnoea, restart lower, which is what both checkable sources actually prescribe.

⚠ A structural observation recorded separately, not as part of this entry’s charge. Krishnamurthy attributes >54% as the level of increased cardiovascular mortality to the Framingham Heart Study, its reference 26 — Gagnon 1994, publicly indexed at 10.1016/0002-8703(94)90679-3 and not obtained through this review’s access routes. Cannarella’s Discussion routes its elevated-haematocrit claim through the same paper, its reference 58. Two of this tranche’s sources rest the threshold’s evidentiary basis on one 1994 paper this review cannot open.

Population and design flags — mandatory

Bhasin 2018 writes for hypogonadal men. Its thresholds, its Table 7 initiation bar and its Table 9 monitoring instruction were not set with masculinising hormone therapy in a trans population in view, and nothing in this limb transfers without the argument made in §9.5. Madsen is the review’s own population and needs no bridge, as limb one records.

20. Antun — two fold-changes in one sentence, resting on two different definitions of the outcome, with neither named

Antun’s Abstract reports that transmasculine participants experienced a 7-fold higher rate (95% CI: 4.1–13.4) of erythrocytosis relative to matched cisgender males, and an 83-fold higher rate (95% CI: 36.1–191.2) compared with cisgender females [31].

Antun’s Methods state that erythrocytosis was defined as haematocrit above 52% in the analyses using cisgender male referents and above 48% in the analyses using cisgender female referents. Table 2’s footnote states the same. [31]

The Abstract names neither cutoff.

The two ratios are not one outcome measured against two referent groups. They are two different definitions of erythrocytosis, applied to two different at-risk subsets of the same cohort — 398 participants never above 52% before hormone therapy, and 311 never above 48%. Presented side by side in one sentence, they read as a single analysis in which the female comparison simply came out larger.

What the 83-fold actually records is that a female haematocrit threshold classifies people taking testosterone as abnormal. ⚠ It is the paper’s own conclusion — Antun recommends that haematological parameters be interpreted against affirmed gender rather than sex documented at birth, and states it in the Abstract’s Conclusions, in the Discussion and in the closing paragraph.

So the paper both produces the figure and supplies its own correct reading of it. What the Abstract’s headline sentence does not do is attach the reading to the figure.

The exonerating material is in the same abstract, and the entry records it

⚠ This entry is weaker than R27.19’s and the difference is stated rather than smoothed. Antun’s Abstract carries the affirmed-gender conclusion four sentences after the fold-change sentence. A reader of the Abstract alone has what they need to interpret the 83-fold correctly, provided they read to the end of it.

What is lost is not the interpretation but its attachment. The qualification survives in the passage and travels no further than the passage. ⚠ Krishnamurthy is the demonstration: it took the ratios from that sentence, did not take the conclusion, and the source-side block records that as Krishnamurthy’s loss rather than Antun’s. Two entries, two authors, one sentence — and they are filed separately because the failures are different.

FM1, and this is the sharper half of the entry

Neither fold-change in the Abstract carries an absolute rate. Antun’s Table 2 carries every one of them [31]:

• Cutoff above 52%, cisgender male referents: transmasculine 26 events, 7.5 per 100 000 person-days; referents 17 events, 1.4 per 100 000 person-days.
• Cutoff above 48%, cisgender female referents: transmasculine 91 events, 39.3 per 100 000 person-days; referents 6 events, 0.39 per 100 000 person-days.

⚠ The 83-fold rests on six events in its comparison group, which is why its interval spans 36.1 to 191.2 — a range of more than fivefold. A fold-change of eighty-three built on six referent events is a fragile quantity presented as a precise one, and the Abstract gives a reader no way to see it.

FM1 in force: the 83-fold does not appear anywhere in this review without those rates inline. This is the clearest case in the tranche of the rule FM1 exists for.

Direction, stated explicitly

⚠ Toward the more striking finding — the standing direction — and qualified. The Abstract’s compression makes the female-referent comparison look like a risk statement, which is the more alarming reading. But the same abstract carries the correction, so this is a failure of attachment rather than of content, and it is recorded at that weight.

Population flag

Antun is 424 transmasculine and 559 transfeminine members of the STRONG cohort across three integrated health systems, matched to cisgender referents on birth year, race or ethnicity, site and enrolment year. This is the review’s own population and no §9.5 bridge is required. ⚠ The cisgender referent groups are internal to the study, not imported — which is what makes the dual-cutoff structure a design choice of Antun’s rather than a transfer problem.
***
## Madsen — an abstract that disagrees with its own Results on one figure, and reports crude odds ratios as the findings

Two intra-source instances, both rulable from a single read, because both sides of each sit inside Madsen. [28]
***
### 21. One figure disagrees between the abstract and the Results

Madsen’s Abstract states that the probability of developing erythrocytosis continued to rise — 10% after 1 year, 38% after 10 years.

Madsen’s Results state that after one year the cumulative risk of developing haematocrit above 0.50 L/L was 8%, after ten years 38%, and at the end of follow-up at fourteen years 50%.

The ten-year figure agrees. The one-year figure does not: 10% in the abstract against 8% in the Results.

A two-point discrepancy on a small number, and the abstract’s version is the larger one. Nothing turns on it clinically. ⚠ It is entered because the abstract is the surface most readers and most citing authors meet, and because this section’s own evidence is that abstract figures are what propagate — Krishnamurthy took Madsen’s prevalences and Antun’s fold-changes from exactly this surface.

⚠ Declared residual. Figure 2 is a Nelson–Aalen curve and its internal values are not present in the archive’s text extraction, so which of the two numbers the curve supports cannot be settled here. What is established is that the paper states both.

Direction: toward the more striking finding, and marginally. The standing direction, at negligible magnitude.
***
### 22. The abstract’s determinants are crude odds ratios, presented without that word, in a paper whose Discussion says adjustment changed them

Madsen’s Abstract lists five determinants associated with haematocrit above 0.50 L/L: tobacco use (OR 2.2; 95% CI, 1.6–3.3), long-acting undecanoate injections (OR 2.9; 1.7–5.0), age at initiation of hormone therapy (OR 5.9; 2.8–12.3), body mass index (OR 3.7; 2.2–6.2), and pulmonary conditions associated with erythrocytosis and polycythaemia vera (OR 2.5; 1.4–4.4).

Every one of those five is the CRUDE odds ratio from Table 2. The adjusted column, in the same rows, gives 2.0 (1.3–3.0) for tobacco, 3.1 (1.7–5.6) for long-acting intramuscular, 3.1 (1.4–7.2) for the age band, 3.1 (1.7–5.6) for the highest body-mass-index category, and 2.0 (1.1–3.8) for a positive medical history. The abstract does not say the figures are unadjusted.

⚠ Two of the five are also band maxima presented as whole-variable effects. Age at initiation is given as 5.9, which is Table 2’s 40–50 band against a reference of under 18; the bands run 2.5, 3.9, 5.9 and then 2.6 at 50 and over, so the relationship is not monotone and the abstract reports its peak. Body mass index is given as 3.7, which is the 30 and over category.

⚠ The paper’s own Discussion is what makes this an instance rather than a presentational choice. Madsen writes that after adjustment for body mass index and smoking the odds of high haematocrit in the 18-to-40 range was much lower, and concludes that higher weight and smoking partially caused the increased odds in that age range. The paper knows the adjustment matters to the age finding specifically, and the abstract carries the unadjusted age figure as its headline determinant.

What is lost is not accuracy but attribution of cause. A reader of the abstract takes away that starting testosterone later multiplies the odds nearly sixfold. A reader of Table 2 and the Discussion finds that most of that is weight and smoking, that the adjusted figure is 3.1, and that the oldest band shows almost no effect at all.

⚠ Amendment C in the form R27.6 established: qualification lost inside one paper, between its own table and its own summary, with no second author involved. Direction: toward the more striking finding. The standing direction.
***
### Population flag

Madsen is 1073 trans men on gender-affirming testosterone — the review’s own population, median age at start 22.5, with twenty years of follow-up. ⚠ No §9.5 bridge is required, and the determinants are the review’s own subject matter, which is why the adjusted-versus-crude distinction is load-bearing rather than pedantic: §9.5’s formulation argument and §9.1’s dose argument both rest on which column is read.
***
## 23. Luo — the Abstract asserts causation that the paper’s own conclusion hedges twice

Luo’s Abstract, Conclusions states that endogenous haemoglobin is the key red-blood-cell trait causing venous thromboembolism [43].

The paper says it three further ways, and each of the three is hedged:

• Clinical Perspective box: which suggested endogenous haemoglobin is the key factor [43].
• Discussion, opening: suggests haemoglobin is the red-blood-cell trait most relevant to venous thromboembolism [43].
• Conclusion, final paragraph: suggests haemoglobin could be the trait most relevant to venous thromboembolism [43].

Two hedges are dropped and one word is upgraded. Suggests disappears; could be disappears; and relevant to becomes causing.

⚠ The Abstract is the only place in the paper that asserts causation flatly, and the paper’s own concluding sentence is doubly hedged. The body is not overclaiming. The Abstract is.

Why this is filed here and not against a citing author

Under R28.1 an abstract is secondary even inside a primary paper — the same rule that governs the Antun dual-cutoff entry and the Madsen abstract entry above. ⚠ And the abstract is what propagates.

The consequence is exculpatory and the entry records it as the finding, not as a caveat. ⚠ Any citing author who wrote that Luo showed haemoglobin causes venous thromboembolism transmitted the Abstract faithfully and is exonerated. The failure is inside Luo, between its own Abstract and its own Conclusion, and it is not available as a charge against anyone downstream.

This is R28.1’s flagship limb repeating in a second paper, and the repetition is the point: the mechanism §9.6 documents is not that citing authors are careless. It is that the compressed, confident version of a finding is the version that travels.

A second Luo limb was examined and DECLINED

⚠ Recorded because a declined limb weighs equally with an entered one (symmetric standard). R30.4 examined Luo’s consistent estimates against the body’s directionally consistent and DECLINED it. It is not an instance and does not enter the count.

Direction, stated explicitly

⚠ WITH §6.1’s standing direction. The entry weakens a causal claim, and this review is sceptical of the haematocrit threshold — so the finding is convenient to the review’s own position. That is precisely why it rests on four quoted passages from one open document and on no inference at all.

⚠ And it should be read against R31.14: §6.1’s standing direction is now understood as a property of this review’s own process rather than of the literature. This entry is a genuine intra-source instance and survives that refinement — but the refinement means an entry running WITH the review’s position carries a higher burden, not a lower one.

Population flag

Luo is a Mendelian randomisation study in a general-population genetic sample. It is not a transgender cohort. ⚠ A §9.5 bridge is required before any Luo finding is applied to this review’s population, and this entry does not supply one.
***
## 24. Defreyne — an inequality that changes across the paper, at the one boundary where the paper’s own figures show it changes the count

Defreyne’s ENIGI cohort is the origin of the haematocrit-threshold prevalences this review carries, and both figures it uses transmit exactly [22]. The defect is not in those figures. It is in the operator attached to them.

The paper states its own thresholds with an inequality that is not stable. Across the Summary, the Results, Table 4’s title, Table 4’s two sub-headings, Table 4’s note and the Discussion, the same two thresholds are written sometimes as at or above and sometimes as above: ≥50.0%, ≥50% and ≥52% in the Summary; ≥50.0% and ≥52.0% in the main Results; >50% in the Results’ body-mass-index comparison; >50.0% and >52.0% in Table 4’s title; ≥50.0% and ≥52.0% in Table 4’s two sub-headings; above 50% and 52% in Table 4’s note; and >50 Hct % and >52 Hct % throughout the Discussion.

In most papers this would be a typographical inconsistency and nothing more. ⚠ In this one the paper’s own reported figures show that it is not.

Defreyne reports that after three months of treatment, eight trans men — 4.3% of those measured — had a serum haematocrit ≥50.0%, with a stated range of 50.0–51.0%. The range opens exactly at 50.0. At least one of those eight therefore sits on the line: under ≥ they are counted, and under > they are not. The count is eight on one operator and at most seven on the other, and this is established from the paper’s own two sentences, not from any assumption about how the measurements were rounded or recorded.

⚠ This is the only instance in this ledger whose defect is provable by arithmetic rather than arguable by reading. Every other entry rests on a judgement about what a passage means. This one rests on a range that opens on its own boundary.

What the entry does not claim. ⚠ It does not claim that any figure this review cites from Defreyne is wrong. The two figures the review carries — 22 of 192 at ≥50.0% and four at ≥52.0% — verify exactly against the paper. Nor does it claim the drift is deliberate, or that any single one of the eight locations is the incorrect one. The paper does not say which operator it means, and this review does not pick one; picking one would conceal the defect rather than record it.

Direction: WITH this review’s position. ⚠ It therefore carries the higher burden, not the lower one, and it is recorded here on eight locations read in situ and a boundary case established from two of the paper’s own sentences.

⚠ RESIDUAL, DECLARED AND NOT RESOLVED. Table 4’s three treatment sub-rows sum three short of its total row — 186 against 189 at baseline, and 181 against 184 at three months. Undocumented rather than wrong: participants may appear in the total without a recorded formulation, and the paper does not say. Recorded so this entry cannot harden past its evidence. ⚠ The denominator structure is otherwise sound — 872 measurements less the 189 at baseline is exactly the stated 683 — so Table 4 is not generally unreliable and must not be characterised as such.
***
## 25. ARIC on Warny — the charge was that a null had been rendered as a shape, and the source’s own figures make the shape and the null both true

The charge carried at R30.8 was that ARIC had converted Warny’s null result into a U-shaped association. Both papers were opened together and it does not hold.

ARIC’s Introduction reports that a large Danish population-based cohort found a U-shaped haematocrit–venous thromboembolism association that was overall not statistically significant, at its reference 8 — verified from ARIC’s own reference list as Warny [45].

Warny’s Figure 2C, multivariable adjusted [40]:

Both tails sit at 44 against a reference of 30. Both are elevated. Neither is significant. That is a symmetric U, and it is what the published category estimates show.

Warny’s own text states that neither high platelet counts nor high haematocrit was associated with risk of venous thromboembolism [40].

The two statements are not in tension, and the absence of tension is the finding. A set of category estimates can trace a shape while no single estimate reaches significance, and a paper can report the second without the first ceasing to be true of its own numbers. ARIC said both things in one sentence — the shape and the non-significance — and both are accurate against Figure 2C.

⚠ Warny never uses the word U-shaped of its own result. The word occurs once in Warny, at §4.2, and there it describes a different study’s platelet–stroke finding. The tempting entry is that ARIC put a word into Warny’s mouth. ⚠ It does not hold: reading a shape off a published set of category estimates is not mischaracterisation. Declined as an instance, and recorded because it ran toward the more striking reading and did not survive the check.

And the characterisation happens three times, not twice — and all three are accurate

R30.8 recorded two characterisations. There are three [45]: the Introduction, above; the Discussion, that the Copenhagen Study reported a nonsignificant U-shaped association; and a third in the anaemia paragraph — that although not statistically significant, spline analysis in the Copenhagen Study suggested venous thromboembolism may be elevated in those with low haematocrit.

The third is the one most easily got wrong, and it is right. It addresses the low tail specifically, and Warny’s low tail is 1.24 (0.98–1.55). May be elevated and not statistically significant are both fair against that interval, and the hedge is placed where the interval requires it.

⚠ A citing author producing three separate characterisations of one source and getting all three right is a stronger result than a single accurate sentence, and it is recorded at that weight. The undercount was this review’s, not ARIC’s.

Direction, stated explicitly

⚠ AGAINST this review’s own case. The entry removes a charge rather than adding one.

⚠ It is also one of the Parts behind §6.1’s fifth refinement (R31.14). The charge this register pointed at most firmly dissolved on opening the paper, and what had been unreliable was the register’s own account of what ARIC said — not ARIC. A verified non-instance weighs equally with an instance, and this section’s credibility rests on that being true in practice and not only in its preamble.

Population flag — mandatory

Warny is 108,521 individuals of Danish descent, mean age 58, with 2524 venous thromboembolic events. ARIC is 13,891 people with 800 events. ⚠ Neither cohort was designed to include people on exogenous testosterone, and neither reports any. Nothing in this entry transfers to a trans-masculine population without the argument made in §9.5.

FM1 is satisfied on the face of the entry: every ratio above sits beside its absolute rate per 10 000 person-years, and no fold-change travels bare.
***
## 26. Okano on Tatarian — a citing author who transmitted fifteen figures without one error, and who demonstrably read the correction

R2.7’s propagation limb — that citers working from the pre-correction text carry Tatarian’s withdrawn intramuscular determinant — stood as a live candidate from Part 2 through four sittings. Krishnamurthy’s half dissolved on a reference list: it never cites Tatarian at all. Okano is the half where a citer does cite Tatarian, and it dissolves on content.

Okano cites Tatarian at its reference 20 [36].

Against Tatarian open in the same step, Okano transmits the following without a single error [36], [30]:

• 511 patients, of whom 113 (22%) experienced polycythaemia
• 77% under 40; 56% with a body mass index above 30; 44% with nicotine dependence
• 46%, 66% and 84% at one, two and three years
• 69% subcutaneous against 31% intramuscular
• body mass index 31.72 against 29.21 (p = 0.0022)
• overall dose 95.42 against 88.46 (p = 0.0085)
• intramuscular dose 101.00 against 88.29 (p = 0.023)
• subcutaneous dose 93.24 against 88.30 (p = 0.049)
• serum testosterone 860 against 737 (p = 0.014)
• band means 873, 856 and 828

Fifteen figures and four p-values.

⚠ The withdrawn intramuscular determinant appears nowhere in Okano’s account of Tatarian. Okano attributes to Tatarian only body mass index, dose and serum testosterone — all three retained after the December 2022 correction. Its one route statement is descriptive, and it points away from the withdrawn claim rather than toward it.

The decisive evidence is a number that moved down a row rather than being replaced

Tatarian’s December 2022 note records three changes, marked in the Results text of the page carrying Table 5 ⚠ (an earlier version read two changes to Table 5; the note also corrects the ≥ 54·0 group’s mean dose, which formerly read 84·42 — correction 218) that are easy to read as one [30]: the polycythaemia-group mean formerly read 873 and was corrected to 860; the 50.0–51.9 band mean formerly read 888 and was corrected to 873.

⚠ The two values did not replace one another. 873 moved down a row.

Okano carries 860 in the group slot and 873 in the band slot — which is corrected Table 5 exactly, on all five values.

A citer working from the August 2022 text would carry 873 in the group slot. Okano does not. ⚠ The figure that looks at first sight like the withdrawn value is the proof that the correction was read — the same two numbers, in the other order, would have been the instance this ledger was looking for.

The citing author did the harder thing. ⚠ And the entry records that the shape of the evidence was the reverse of what the charge predicted.

A residual examined and declined, recorded so it is not re-derived

Tatarian writes that specific testosterone esters included cypionate and undecenoate, said of the whole polycythaemia group; Okano attaches the esters to the intramuscular subgroup specifically.

⚠ A narrowing, not a reversal, and it bears on nothing this review cites. Recorded, not entered.

Direction, stated explicitly

⚠ AGAINST this review’s own case, and it closes the oldest live charge in the tranche.

Nothing false was ever written from R2.7, because R27.11 refused to enter it on plausibility and R13.3a and R18.1a held the line until a receiving paper could be opened. ⚠ What the rule caught was the cost of deferral, not an error of assertion — the charge could have been closed at any point by opening two reference lists.

⚠ And the exonerating evidence was the source’s own published correction, faithfully transmitted. That is the sharpest available counter-example to this section’s own thesis, and it is recorded first rather than buried.

Population and design flags — mandatory

Tatarian is 511 transmasculine patients on gender-affirming testosterone at a single United States centre — the review’s own population. What limits it is design: retrospective, single-centre and chart-derived.

⚠ OKANO’S OWN DESIGN AND POPULATION ARE NOT CHARACTERISED HERE, AND THE GAP IS DECLARED RATHER THAN FILLED. Any statement about what Okano is requires a fresh read under R13.3a. ⚠ This does not weaken the entry — the charge is about citation, not about population transfer — but the section must not imply a population flag it has not established.

FM1 is satisfied: the absolute base rate travels with every ratio above — 113 of 511, 22%.

Residual, declared, and this entry’s decisive limb rests on it

⚠ The archive’s extraction of Tatarian flattens at least four distinct correction markers into the same character. So the binding of each superseded value to its position rests on order-matching — the order of the correction list matching the order of the marked values, which holds across all five values in that note, three of them verifying independently against Tables 2, 4 and 5 — and not on a distinct symbol.

Settling it beyond order-matching needs the publisher’s typeset PDF. ⚠ Recorded prominently because the 860/873 binding above is this entry’s decisive limb, and it must not harden past its evidence.
***
## This review’s own two — the subsection R22.2 created and R33.5 confines

Basis for this subsection. R18.2a asked whether a characterisation failure committed by this review enters §9.6, given that §9.6 had catalogued characterisation-of-source failure in other people’s work. Eden ruled it in at R22.2. Two such failures qualify: both are the same failure class the section catalogues, both occurred at a known point of vulnerability, and both were caught by the review’s own gate rather than by a reviewer. Entering them answers, permanently, the objection that the ledger scores points its subjects cannot answer, and it is what the symmetric-bias standard requires. The section is therefore a catalogue of characterisation failure, not a catalogue of other people’s characterisation failure — the entries below carry the same evidential standard as every entry above, and no lighter one for being the review’s own.

R22.2 sent two of that session’s four recorded errors here and two to the §6 failure log, and the split is load-bearing. The §6 pair — a block inversion and an inference drawn from a directory listing — are process errors, not characterisations of a source’s content. Placing all four here would dilute what this section is. What enters §9.6 is a failure in which the review’s own text misstated what a source says; nothing else.
***
## 27. An accurate quotation about myocardial infarction was placed in a list about venous thromboembolism, its referent silently changing on the way

Warny’s §4.4 reads that individuals with high haematocrit had a 1.5-fold risk of arterial thrombosis in the heart, and that it seems unlikely that adding haematocrit would improve risk prediction in this disease. The referent of this disease is myocardial infarction — the outcome the surrounding paragraph is about.

The venous block, §9.4.2, placed that sentence fourth in a list whose other three items are all about venous thromboembolism. Read in that position, Warny appears to say that haematocrit would not improve prediction of VTE. Warny makes no statement about VTE risk prediction anywhere in the paper.

The decisive detail is that the same sentence is deployed correctly elsewhere in the review. At §9.4.1, where the surrounding outcome is myocardial infarction, the sentence sits in its proper context and says what Warny says. So the defect is not a misreading of Warny — the quotation is accurate and its arterial use is sound. The defect is transplantation: an accurate quotation whose referent changed the moment it moved between two of the review’s own blocks, because the receiving block was about a different outcome and nothing marked the sentence as belonging to the outcome it was carried from.

Entered as an internal instance. This is the section’s own failure class — a source’s statement about A represented as a statement about B — occurring not in a downstream paper but between two sections of this review, and surviving until the review’s gate reached it. It was corrected when caught. It is entered here not because it remains live but because a ledger that omits the one instance it committed itself is not a ledger of characterisation failure, it is a ledger of other people’s.

⚠ Direction is not assigned. The transplant does not run cleanly toward the more striking finding: a this will not improve prediction verdict is if anything the unstriking reading, and what moved it was proximity in a list rather than a reach for a sharper claim. A referent that slips because two sections sit adjacent is a structural failure, not a rhetorical one, and is recorded that way.
***
## 28. A qualification the source log preserved was lost at the first internal step, and an open item inherited the flattened version

The Warny defect log records §4.3 of Warny with its ambiguity intact. Warny’s §4.3 says the review cannot exclude that the lack of statistical significance for arterial thrombosis in the heart in those with high haematocrit represents a type II error — a sentence that contradicts the paper’s own headline 1.46 (1.06–2.00) for the haematocrit–MI association. The log gives two readings and declines to disambiguate: a slip for platelet count — which the log calls the more likely, since platelets predict brain (1.77) and not heart (0.82) — or a reference to the competing-risk analyses, where the sub-hazard ratios are borderline at 1.38 (1.00–1.90), p = .05 in Figure 2B and 1.34 (0.98–1.82) in Figure 5B.

The register’s own recommendation, R18.4, carried neither reading. It compressed §4.3 to a Limitations sentence describing the haematocrit–MI association as lacking statistical significance, and recommended that the review state the contradiction. Open item 31 was written against R18.4 and inherited the compression.

The transmission runs log → register → open item, entirely inside the review’s own records, and the qualification is lost at the first step. What the log holds as a sentence with two possible referents, the more likely of which is a typo for a different variable becomes, one step downstream, the paper contradicts its own principal result — and the open item then proposes to publish that flattened form.

Entered as an internal instance. This is the qualification-loss-in-transmission pattern the section documents in source-to-source citation, occurring across three of the review’s own artefacts. It was caught by re-reading the log against the register (R21.5), and item 31 has since been reworded: the review may state that §4.3’s referent is undisambiguated and name the log’s preferred reading, but the flat version must not be used.

⚠ Direction: the standing one. The flattened form is the more striking of the two. The compression discarded exactly the material that made the finding less dramatic — which is the mechanism §6.1 names, operating here on the review’s own text.

29. The corrigendum that fixed one error introduced another — the only statistically significant change is false against the paper it corrects

⚠⚠ THIS ENTRY IS ADDED AT v4.7 AND RENUMBERS NOTHING. §7’s constraint holds: the ledger’s order is fixed and later instances are appended. ⚠ Both sides were opened untruncated in the step it was written — Toorians 10.1210/jc.2003-030520 id 106 and the corrigendum 10.1016/j.thromres.2026.109804 id 192.

The section’s front matter already records that a 2025 literature review misstated Toorians’s finding as a decrease in the level of activated protein C, and that a corrigendum of 21 August 2026 replaced that with a decrease in the normalised activated protein C sensitivity ratio, reflecting reduced acquired resistance, with the original authors concluding a mild antithrombotic effect.[66]

The replacement clause is correct on the quantity. It is wrong on the scope. The corrigendum states that this was the only statistically significant change induced by 4 months of testosterone treatment.[66]

Against Toorians’s Table 3, read row by row, it was not. In the fourteen trans-masculine participants:

• normalised APC sensitivity ratio 2.0 ± 0.8 → 1.3 ± 0.7, p < 0.001;
• total protein S 105 ± 22 → 118 ± 19%, p = 0.037;
• free protein S 83 ± 20 → 97 ± 15%, p = 0.012;
• protein C 110 ± 18 → 100 ± 15%, p = 0.070 — not significant;
• prothrombin 106 ± 18 → 107 ± 16%, p = 0.826 — not significant.[48]

Three haemostatic variables reached significance, not one. The paper’s own summary figure plots the protein S rise beside the APC-resistance fall, and its abstract names testosterone’s antithrombotic effect without restricting it to a single variable.[48]

The direction of the loss is the interesting part, and it does not run the way this review’s errors run. ⚠⚠ Both omitted changes point the SAME WAY as the one reported. A rise in protein S, an anticoagulant, is an antithrombotic change; so is a fall in acquired APC resistance. The corrigendum therefore understates the breadth of the antithrombotic signal in the paper it is correcting — which is the less striking reading, not the more striking one.

⚠ This is the third instance in this ledger running against the standing failure direction, and it is recorded at the same weight as the instances that run with it.

⚠⚠ AND IT IS THE SECOND TIME A FORMAL CORRECTION IN THIS CORPUS HAS INTRODUCED A NEW DEFECT WHILE REPAIRING AN OLD ONE. Entry 10 records a correction that moved an error down a row rather than removing it. This one repairs a quantity and breaks a scope in the same sentence. A published corrigendum is not self-certifying, and this section now has two demonstrations of that.

⚠ A limb examined and declined. The corrigendum’s phrase might be read as scoped to the specific variables that review discusses rather than to Toorians’s whole result. It does not survive: the sentence says the only statistically significant change induced by 4 months of TT, which is a claim about the treatment’s effects and not about the review’s selection.[66] Declined.

Population flag — mandatory

Toorians is 14 trans-masculine people at the Vrije Universiteit gender clinic in Amsterdam, mean age 26, on testosterone esters 250 mg every two weeks for four months — the review’s own population, and no §9.5 bridge is required. ⚠ What limits it is size, duration and the absence of any control group; one participant carried factor V Leiden.[48] ⚠⚠ It is also the earliest member of the overlapping Amsterdam assay lineage mapped at §9.4.6, and it must not be counted as an independent cohort alongside the later analyses from that clinic.


## 30. Cannarella — three of eight checked rows misreport the initiation value, and two do it by substituting the on-treatment one

⚠⚠ THIS ENTRY IS ADDED AT v4.16 AND RENUMBERS NOTHING. ⚠ Both sides open: Supplementary Table 7 and its Discussion[6], against seven society guidelines read first-hand — AUA and ISSAM untruncated, EAA, BSSM, AACE and SIAMS/SIE truncated inside their reference lists after the last recommendation.

Supplementary Table 7’s action column is headed Recommendation against TRT. Against the primaries, five of the eight checked rows report that society’s initiation position accurately — BSSM at >54%, EAA at >48–50%, ISSAM at >52%, SIAMS/SIE as no number with global cardiovascular risk considered, and ICSM at >50%, which is the accurate number carrying a defect of a different kind: the tabulation cannot carry ICSM’s own attribution of that value to another body’s 2010 guideline or its statement in the next sentence that such contraindications must be re-evaluated.

Three do not:

• The Endocrine Society row is printed Hematocrit<48%. The guideline sets >48%, or >50% at altitude, as a relative contraindication.[7] The operator is inverted and the altitude qualifier is absent. (This limb is entry 8; it is named here for completeness and not counted twice.)
• The AUA row is printed >54%. AUA’s initiation instruction is that above 50%, clinicians should consider withholding testosterone until the aetiology is formally investigated; 54% is its on-treatment intervention level.[12] The on-treatment value is printed under the initiation heading and the initiation value is absent.
• The AACE row is printed >50%. AACE names polycythaemia a relative contraindication with no number; >50% is its on-treatment decrease-or-discontinue level.[9] Same substitution.

⚠⚠ AND THE STRUCTURAL FINDING IS LARGER THAN THE THREE ROWS. Seven of the eight verified Cannarella rows set an on-treatment action level DIFFERENT from their initiation position, and not one of those values appears in the tabulation this review reads them through. Where a society distinguishes two objects, the instrument can carry only one.

⚠⚠ WHAT IS LOST IS NOT A NUMBER BUT AN AGREEMENT. Seven of the eight set the on-treatment level at 54%. A reader of the table sees a 48–54% spread and concludes the societies disagree; a reader of the guidelines finds the disagreement is about INITIATION and that the on-treatment threshold is very nearly unanimous. ⚠ This review made that error itself and carried it for many versions.

⚠ Direction: MIXED, and both halves are stated. The three row-level defects make the societies look more permissive about STARTING treatment than they are — which runs against the treatment-sceptical reading. The structural loss made the guideline literature look more disordered than it is — which runs WITH this review’s position, and which this review reproduced.

Population flag — mandatory

Every guideline in this entry writes for hypogonadal men. ⚠ Not one threshold in Supplementary Table 7 was set with masculinising hormone therapy in a trans population in view, and nothing in this entry transfers without the argument made at §9.5.

31. A randomised analysis titled for harm in a stratum whose own estimates are null — and the citing letter is exonerated

⚠⚠ ADDED AT v4.18; RENUMBERS NOTHING. ⚠ Both sides opened untruncated in the step: the primary[59] and the letter commenting on it[67].

The primary is titled Canagliflozin may increase thromboembolic events in males with erythrocytosis but not in females, and its Key Points state that baseline haematocrit modified the effect on thromboembolism, indicating harm in males with erythrocytosis.[59]

Against its own Figures 3 and 5, in erythrocytotic males: the primary composite estimate is 1.08 (95% CI 0.68–1.69) and the myocardial-infarction estimate is 1.77 (0.76–4.13). ⚠⚠ Both cross unity. The significant interaction is carried substantially by the anaemic stratum, where canagliflozin is protective at 0.64 (0.49–0.85) and 0.53 (0.34–0.83). The paper states that no adjustment was made for multiplicity.[59]

⚠ The finding of effect modification is real and the paper’s Discussion is careful. It says the harm is potential, that the analysis is exploratory in trials not designed for heterogeneity, and that power in the erythrocytosis stratum is limited. ⚠⚠ What the title and Key Points carry forward is the harm and not the null strata, and it is the title and Key Points that travel.

⚠⚠ AND THE CITING LETTER IS EXONERATED, WHICH IS THE POINT OF ENTERING THIS. The letter reports that the primary showed benefit limited to anaemic patients and may confer thrombotic risk in those with baseline erythrocytosis, particularly males.[67] That is a faithful transmission of the primary’s own summary, hedged exactly as the primary hedges it. Under Amendment C an abstract, title and Key Points block are secondary characterisations even inside a primary paper, so a citing author reading them transmits what is there. ⚠ The defect is intra-source, as at entries 20, 21, 22 and 23, and it is not available as a charge against anyone downstream.

⚠ Direction: AGAINST this review’s position. A headline asserting harm from erythrocytosis is the reading this review is sceptical of, and correcting it toward null strata strengthens this review’s own case — which is why it rests on four figures quoted from one open document and on no inference.

Population flag — mandatory

Older people with type 2 diabetes and established or high cardiovascular risk, median age 63. ⚠ Not a transgender cohort, and sodium-glucose cotransporter 2 inhibition raises haematocrit by a different mechanism from testosterone. Nothing transfers without the argument at §9.5.

32. A position statement’s account of a regulator’s action — one category widened, one half omitted, one verb collapsed

⚠⚠ ADDED AT v4.52; RENUMBERS NOTHING. ⚠ Both sides opened untruncated in the step: the regulatory release[50] and the position statement citing it at its reference 45[54].

The release states two sets of changes, driven by two evidence bases and carrying two different regulatory verbs. Led by the TRAVERSE trial, the agency is RECOMMENDING: adding the trial’s results to all testosterone products; retaining “Limitation of Use” language for age-related hypogonadism; and removing from the Boxed Warning the language on an increased risk of adverse cardiovascular outcomes. Led by the ambulatory blood-pressure studies, it is REQUIRING: product-specific information on increased blood pressure for products with completed studies, and a new warning about increased blood pressure for products which currently carry none.[50]

The position statement reproduces three bullets. Set against the release:

• Its first and third bullets transmit without error.
• ⚠⚠ Its second reports the Limitation of Use as retained for FUNCTIONAL hypogonadism. The release says AGE-RELATED. The statement’s own definitions make these different objects — it defines functional hypogonadism as arising from inflammatory conditions, obesity or chronic illness, and as most often, but not exclusively, found in older men. Functional is the broader category; reporting the restriction as covering it widens the population it is described as restricting.[54]
• ⚠⚠ The REQUIRED blood-pressure half is absent. A reader learns that a cardiovascular warning was removed, and not that new blood-pressure warnings were required class-wide in the same action.
• ⚠ And the verb is collapsed: the statement says the agency is implementing changes, where the release recommends one set and requires the other.

⚠ No intent is attributed. A statement whose subject is cardiovascular safety may have judged the blood-pressure limb out of scope. ⚠ What is recorded is that its abstract states the document is in agreement with recent United States Food and Drug Administration label changes, and that the half omitted is the half imposing a new warning.[54]

⚠⚠ AND THIS REVIEW READ THE PRIMARY AND DID NOT COMMIT THE ERROR. §9’s paragraph carries the age-related wording, carries the separate blood-pressure requirement, and says requested rather than any completed-action verb. Recorded as a verified non-instance at the same weight a defect would carry.

⚠ Direction: FOR this review — a source it cites is shown to be less reliable than it appears — and BOUNDED: the defect is in the transmitting document and not in the regulator’s action, and it touches no haematocrit threshold.

33. A guideline’s account of a sensitivity analysis — “even higher at 54%” where the printed odds ratio is below unity

⚠⚠ BOTH DOCUMENTS READ UNTRUNCATED IN THE STEP THIS WAS WRITTEN, per R13.3a. The transmitting document is the EAU Sexual and Reproductive Health (2026) guideline (EAU-SRH-2026, held in the archive, added 2026-09-11; included as reference [70]; held as archive id 332, correction 496). The transmitted document is Ory et al. (2022).[20]

The EAU-SRH-2026 guideline at §3.5.6 cites a multi-institutional database study of 48,671 hypogonadal men on testosterone therapy — 5,887 of whom developed polycythemia (haematocrit ≥52%) — and reports that it demonstrated higher MACE/VTE risk at the ≥52% threshold. It then states:

“The risk was even higher when a haematocrit threshold of 54% was considered, whilst no risk was observed when a 50% threshold was applied.”

The description matches Ory et al. precisely: n=5,887 who developed polycythemia (haematocrit ≥52%), TriNetX database, MACE/VTE primary outcome, sensitivity analyses at 50% and 54%.

What Ory’s printed sensitivity analysis actually shows:

The paper reports: “Odds of MACE remained higher in the polycythemia group when defined as hematocrit >54% (OR 0.78, 95% CI 0.62–0.98, p=0.03).”

⚠⚠ The printed odds ratio is below one, whereas the accompanying prose describes higher odds. Under the orientation established for the MAIN analysis, those would disagree. ⚠⚠ THE REPORT DOES NOT ESTABLISH WHETHER THAT ORIENTATION WAS IMPLEMENTED IN THE SENSITIVITY ANALYSIS, SO THE DIRECTION CANNOT BE RESOLVED FROM THE PRINTED NUMBER ALONE (correction 556).

The EAU guideline transmits the prose interpretation (“even higher”) without noting the contradictory printed OR. A reader of the guideline learns that risk escalates at 54%, without being told that the paper’s printed figure and its own prose disagree. ⚠ This review infers no direction from that odds ratio, and does not place the figure in a table: it is printed as prose in the results (correction 556).

⚠ This review’s own manuscript (§9.4.1) declined to interpret the direction of that result, and its caution is confirmed here. The MAIN analysis’s orientation is established: its Methods state the comparison and its own event counts pin it arithmetically. ⚠⚠ THAT DOES NOT ESTABLISH THE ORIENTATION IMPLEMENTED IN A SEPARATELY GENERATED SENSITIVITY ANALYSIS. Under the main orientation the printed sensitivity odds ratio would indicate lower odds while its own prose states higher odds, and the report does not resolve whether the figure, the prose, or that analysis’s contrast labelling is wrong (correction 532). What cannot be determined is which of the paper’s own limbs is in error — the printed figure or the sentence reporting it (correction 499).

⚠ No intent is attributed to either document. Ory’s internal contradiction may reflect an error in one of its own numbers, a comparison-group change between the main analysis and the sensitivity analysis that the prose does not make explicit, or a typographical transposition. The EAU guideline cannot have resolved a contradiction it did not identify. What is recorded is the state of the evidence as transmitted. ⚠ The §9.4.1 treatment — reporting the result with both the printed OR and the prose, noting their disagreement, and leaving the direction ungraded — is the correct handling and is unchanged.

⚠⚠ Direction: AGAINST this review’s argument in its strongest form — the result at ≥52% in Ory supports threshold-linked risk, but the ≥54% sensitivity analysis cannot be read as additional support for the 54% guideline threshold. The review should not build on either reading. ⚠ And against the guideline transmission: the EAU-SRH-2026 characterisation of “even higher risk at 54%” is not supported by the printed OR.

Population flag — mandatory

[20] was conducted in hypogonadal men in the United States (mean age 53·4, predominantly Caucasian, high comorbidity burden). The EAU guideline is written for men with hypogonadism. ⚠ Neither document concerns trans-masculine people or gender-affirming hormone therapy, and nothing in this entry transfers without the argument made at §9.4.1 and summarised at §12.

Population flag — mandatory

Both documents write for men with hypogonadism. ⚠ Neither concerns trans-masculine people or gender-affirming hormone therapy, and nothing in this entry transfers without the argument made at §9.5.

⚠ §9.6 ENTRY COUNTS — GENERATED FROM THE SECTION’S OWN HEADINGS. Thirty-seven entries are numbered. Twenty-six carry their own top-level heading; eleven are nested beneath a source-level heading.

⚠ §9.6 ENTRY COUNTS — GENERATED FROM THE SECTION’S OWN HEADINGS. ⚠ SUPERSEDED, RETAINED BY QUOTATION. An earlier version of this sentence read thirty-two entries have been numbered; entry 9 was subsequently withdrawn, leaving thirty-one live.

⚠ §9.6 ENTRY COUNTS — GENERATED FROM THE SECTION’S OWN HEADINGS. ⚠ SUPERSEDED, RETAINED BY QUOTATION. An earlier version of this sentence read, in the separately issued v4.52, the ledger has numbered thirty-two entries; eleven of those numbers are vacant, entry 9 by withdrawal, leaving twenty-one live. ⚠⚠ §9.6 defines vacant nowhere. The term is retired rather than defined, and that reading’s arithmetic was CORRECT. ⚠ Entry 29 was appended at v4.7, entry 30 at v4.16, entry 31 at v4.18 and entry 32 at v4.52; none renumbered anything. ⚠⚠ Because the withdrawal altered the prior partition, no failure-to-clean-transmission split is reported until the ledger is mechanically reclassified.

34. One trial, two registrations that do not reconcile — and the divergence is at registration, before any citation chain begins

⚠⚠ The randomised pilot at [62] and [65] is registered twice, and the two records do not reconcile. Which of them describes what was done is not established here.

⚠ The published report does not state masking in its abstract, and this review has not obtained its full text, so which record is correct is not established here.[81]

⚠⚠ This entry records a divergence between two records. It does NOT allege an error by any author. The two registries use different masking vocabularies and name non-corresponding parties, so part or all of the difference may be a schema difference rather than a contradiction. A registration that is filed and never updated is also common, and the observation here is about what the records say, not about the conduct of the trial.

⚠ Why it belongs in this ledger. Every other entry above traces a qualification lost as a finding passes from one document to another. This one occurs earlier: the two primary descriptions of the study disagree before any citing author has read either. A reader combining fields from the two registrations without reconciling their definitions and dates could produce an inaccurate or unsupported composite description, and the records should be compared with the protocol and the publication before they are treated as interchangeable accounts (correction 541).

⚠ Direction: against this review’s own use of the source. [62] is cited four times and [81] supplies the clinical result. Recording this weakens material the review relies on, which is the reason to record it.

Population and design flags — mandatory

⚠ The trial is in trans men with testosterone-induced erythrocytosis, which is this review’s population; nothing in this entry transfers without the argument made at §9.5.

35. A threshold-specific risk statement whose cited support contains no such estimate — and one end of it this review has not opened

⚠⚠ ENIGI reports threshold-crossing frequencies and NO OBSERVED THROMBOEMBOLIC EVENTS in the relevant cohort. It does not provide a comparative estimate showing that a haematocrit above 48 per cent increases thromboembolic risk.[22] Its zero-event observation neither establishes such an association nor excludes one — what it bounds depends on person-time, and the paper does not report it: it gives a visit schedule, per-visit denominators and 683 measurements on treatment, but no person-years, so the zero cannot be converted into a rate or an upper bound.

⚠⚠ WHAT THIS ENTRY CLAIMS. If [22] is the sole source cited for a threshold-specific risk statement, it does not support that statement. That is established from [22] alone, which this review holds and has read untruncated.

⚠⚠ WHAT THIS ENTRY DOES NOT CLAIM, AND IT IS NARROWED TO SAY SO (correction 486). An earlier form stated that [25] writes that haematocrit above that level is associated with thromboembolic complications and cites [22] for it, and called the pair verifiable from both ends. This review has not obtained [25]’s full text. Its citing sentence and reference number have not been placed beside [22]’s Results by this review, and until they are, no transmission error is alleged against [25].

⚠ An earlier form also wrote that [22] does not establish what happened to those people because nothing happened that it recorded. That is wrong as written: a reported zero IS outcome ascertainment, not an absence of outcome information. The defect worth recording is the absence of a COMPARATIVE ESTIMATE, not an absence of observation.

⚠ Direction. Neutral for this review’s central argument, which concerns whether the thresholds are outcome-calibrated and does not rest on [25]’s characterisation either way.

Population flag — mandatory

⚠ Both sources are trans-masculine cohorts on testosterone, which is this review’s population; nothing in this entry transfers without the argument made at §9.5.

36. Two arithmetically impossible values in a systematic review’s data table, and a third count that contradicts itself four sentences later

⚠⚠ [36]’s Table 1 gives Jarin’s haematocrit over six months as 15·0. A haematocrit of 15·0 is clinically discordant with the reported context and requires verification against Jarin’s own table. It must not be used as a reliable haematocrit value, and it must not be silently replaced by an inferred one: a unit error or a haemoglobin column are equally available explanations, and severe anaemia at that level is survivable rather than impossible. An earlier form of this entry called it incompatible with life and almost certainly 45 — a number repaired by plausibility, in the section that catalogues other people’s transmission errors (correction 664).

⚠⚠ The same table gives Defreyne’s gel arm a six-month median of 36, sitting between 43 at three months and 46·5 at twelve — a seven-point fall and recovery that the source study does not report.

⚠ And the same paper describes Nolan as eight patients (6%) and, four sentences later, as six individuals.

This review reproduces none of the three, and that is the point at which the defect would otherwise have passed unrecorded: not reproducing a wrong figure leaves no trace, so a reader cannot tell whether it was caught or never met. The entry exists so that it is the first.

⚠ Why it is entered rather than logged as a source defect. §9.5.2 already records that this paper characterises a randomised trial as a small observational study, and that its own discussion endorses two contrary accounts in different passages. These are the same paper’s fourth, fifth and sixth internal inconsistencies, in a source cited sixteen times and used as the citation route to [82], which this review holds only through it.

⚠ Direction: against a source this review relies on. No claim of this review changes. What changes is the weight [36]’s unverified characterisations can carry — including the one at §9.5.2 that supplies [82].

Population flag — mandatory

⚠ The review is of testosterone formulations in transgender individuals, which is this review’s population; nothing in this entry transfers without the argument made at §9.5.

37. A relative measure’s confidence interval reported as an absolute risk, then subtracted from a hundred

[36] reports a hazard ratio for erythrocytosis in trans men relative to cisgender men, citing [31]. Its Discussion also presents an absolute-risk range whose endpoints resemble that ratio’s confidence limits.

⚠⚠ Its Discussion then writes that the absolute risk ranges between roughly 4% and 13%, which means that 87%–96% of patients do not have erythrocytosis, which might suggest that the data are not worrisome for this analysis.

The asserted absolute-risk range is not justified by the reported hazard-ratio interval. Numerical similarity suggests that the two may have been conflated; it does not prove the calculation that produced the statement. A confidence interval for a relative measure cannot be converted into the proportion unaffected by subtraction from 100%.

The threshold-specific eligible denominators, event counts and observed proportions are set out in §9.2.2. Dividing those event counts by the full baseline cohort would answer a different question and would not reproduce the source’s first-crossing proportions. No confidence interval for absolute risk is derived from a hazard-ratio interval here.

Why this matters. A relative effect measure and an absolute risk are different statistical objects. The unsupported conversion can produce misleading reassurance. The observable reporting defect is recorded here without attributing an unverified calculation or motive to the source authors.

⚠ Direction: this entry runs AGAINST a passage this review carried in its own disfavour. §9.2 quoted the sentence as a source conceding that its own headline ratio was not alarming. The concession was real as a statement and unfounded as arithmetic, and removing it removes a qualification this review had imposed on itself — which is recorded here so that the removal cannot be read as a gain.

This issue was identified during claim checking. Faithful quotation and scientific validity are separate adjudications: confirming that [36] contains the sentence establishes fidelity, not the validity of its risk interpretation. The full claim census remains incomplete.

Population flag — mandatory

⚠ [36] is a systematic review of testosterone formulations in transgender individuals, which is this review’s population. The defect is in its handling of [31]’s statistics, not in a transfer across populations.

Closing — after the count sentence, unnumbered (R33.4, R34.2)

A closing note about this section’s own reliability, which the section’s method obliges it to record.

This ledger catalogues qualification lost as findings pass between papers. In the course of building it, the same failure was found operating inside this review’s own records. Six times in succession, the charge these records pointed at most firmly dissolved when the paper itself was opened — three of those on nothing more demanding than the cited paper’s reference list, where the accused author had not cited the paper they were said to have misrepresented. Once more, the exonerating material sat in the cited paper’s own abstract — Antun’s — meaning the citing author had transmitted it faithfully. The conclusion this section draws about itself is therefore narrower and less comfortable than the one it set out to draw: the sources examined here have been about as careful as sources are, and the least reliable transmitter in the set was the review’s own intermediate record of what they said. Two entries above are this review’s own, and they are there for that reason.

⚠⚠ AND THE SECOND EXTERNAL REVIEW OF THE ASSEMBLED MANUSCRIPT EXTENDS THAT CONCLUSION ONE STEP FURTHER, AND IT IS THE LESS COMFORTABLE STEP. The failures found in v4.4 were not transmission failures between records. They were failures of the assembled document to describe itself — a subsection carrying three incompatible accounts of its own evidence base, a map describing a corpus its own Part V had superseded, and a count about the correction record that the correction record contradicts. ⚠ This section’s method — open both sides, verify in the step, weigh the clean case equally — catches none of them, because there is no second document to open. That is a limit on what this ledger can be offered as, and it is recorded here rather than left for a reader to find.

⚠⚠ §9.7 — RECOMMENDED FOR DELETION, NOT REWRITING

It restates §3’s definitions and duplicates §10’s table inside Part V. The v0.2 text is not reproduced here.

⛔ §9.8 — REMOVED, NOT CARRIED

v0.2’s What this section establishes, and what it does not is deleted from this assembly rather than reproduced with a warning. ⚠⚠ It contradicted the rebuilt sections on three counts: it described the mechanism as an early erythropoietin rise with hepcidin suppression, which §9.1 grades NOT ESTABLISHED; it stated a haematocrit-to-venous-thromboembolism association in a form that flattens the four-cohort division §9.4.2 establishes; and it referred to the only cohort that counted events when §9.4.3 now integrates five.

⚠ An expressly superseded but still-live bottom-line section cannot be excluded from the scientific claim by a marker. It is removed. ⚠ A replacement is owed, written from the rebuilt sections rather than from v0.2.

PART VI — EVIDENCE QUALITY, AND THE BOTTOM LINE

10. The grading, gathered — REBUILT

Every mark is the mark the section assigned. Read the last two columns together.

⚠⚠ THIS TABLE AND ITS COUNTS ARE GENERATED FROM THE GRADING BLOCKS OF PART V OF THIS ASSEMBLY, AFTER IT IS BUILT. ⚠ An earlier build generated it from the PREVIOUS version’s Part V, so it was reliably one version stale — which is why external review kept finding marks in Part V that the table did not carry. That fault is fixed here, not patched.

⚠ §9.4.3’s row was added at v4.5. It carried a graded block in Part V and no row here, which is the mirror of the defect §9.4.6 was created to fix.

⚠⚠ THREE ROWS MOVED AT v4.6 AND ALL THREE MOVES ARE AXIS CORRECTIONS RATHER THAN RE-READINGS OF ANY SOURCE. The pooled randomised row’s causal mark rises from LIMITED to MODERATE, because sparse events are an imprecision fact and belong on the effect-size axis; the haemostasis row’s causal mark falls from MODERATE to LIMITED, because prospective uncontrolled measurement establishes within-person change and not causal attribution; and the management row’s causal mark falls from MODERATE to LIMITED, because provenance caps what an abstract-plus-registry read can support. ⚠ One rises and two fall. Corrections 48, 49 and 50 record them.

Generated counts: 3 ROBUST cells · 8 rows marked direct · 1 NONE IDENTIFIED · 12 NOT ESTABLISHED · 14 LIMITED · 7 MODERATE · 7 NOT ESTIMABLE. ⚠ A dash means the source block assigned no mark on that axis, not that the axis is favourable.

Reading it down

⚠⚠ THE COUNTS ARE PRINTED WITH THE TABLE AND ARE COMPUTED FROM IT. They are not restated here, because an earlier version restated them by hand and the two disagreed.

Every ROBUST cell in the table is about the blood count rising. None is about anything happening as a result.

Two rows carry evidence AGAINST the proposition rather than merely an absence of evidence: ⚠ against close quantitative correspondence between haematocrit and red-cell mass in the one measured referral subset, and against a consistent pro-thrombotic shift in the GLOBAL COAGULATION AND FIBRINOLYSIS assays measured after masculinising therapy. ⚠ Both were previously carried as unknowns. They are answered, in the direction opposite to the intuition, in the populations where they were measured.

⚠⚠ THAT SECOND ROW IS SCOPED TO ONE ASSAY FAMILY AND MUST NOT BE READ ACROSS THE CORPUS. The platelet and inflammatory family is a SEPARATE row and is graded LIMITED. ⚠⚠ It is no longer directionless: the identified whole-blood analysis has been obtained, and the family is DIVIDED. Two same-lineage analyses disagree on platelet activation — the earlier reports no change in platelet markers, while the later, measured in whole blood, is read by its authors as compatible with increased platelet activation and inflammation. ⚠ The later is small and reports significance in one of seven platelet markers and three of twelve inflammatory markers across nineteen tested without multiplicity correction, so it does not establish a coherent prothrombotic shift. ⚠⚠ What the two agree on is high-sensitivity C-reactive protein, which rises in both — an inflammatory rather than a direct haemostatic endpoint.

⚠ CORRECTION NOTICE, IN ITS OWN PARAGRAPH. An earlier version of this paragraph said the plasma-based assays, collapsing the two families back into the single synthesis correction 62 removed; and a later one continued to describe the platelet corpus as incomplete, the whole-blood analysis as unobtained and the family as directionless, after Part V had recorded all three as superseded. Corrections 62 and 96.

⚠⚠ And no patient-outcome claim in this review has an estimable clinical importance attributable to haematocrit — in a review about a monitoring threshold. The reason is stated rather than left as an omission: crossing a laboratory threshold is a surrogate, and no study identified in this review relates crossing it to an outcome in this population.

11. The apparatus in practice, and where the two questions meet again

This section describes a moving position and will date faster than anything else here. It is quarantined deliberately. Nothing in Parts I to V depends on it, and a reader who skips it loses no part of the argument.

§1 separates the evidence question from the policy question and keeps them apart for ten sections. This is where they come back together, because a clinician cannot keep them apart. A patient is in the room. A haematocrit is on the screen. Something has to be done, or not done.

What the apparatus asks for is not in dispute. As §5 sets out: across the guidance examined here, baseline and on-treatment measurement are widely recommended, and some guidance also names numerical initiation or on-treatment action levels. ⚠ Not all of it does. What varies is the level. The spread is wide enough that the same result triggers action under some guidance and not under others.

What follows from a threshold being crossed varies more than the threshold does. Reduce the dose. Alter the dosing schedule. Switch preparation. Use venesection or blood donation. Interrupt treatment. ⚠ The evidence for these differs sharply, and §9.5 sets out how: lengthening a fortnightly cypionate interval to monthly did not lower the haematocrit, while discontinuation did. These are not equivalent acts. Some are reversible within weeks and some end the treatment, and guidance that names a number does not always name which response it intends.

⚠⚠ AND ONE DISTINCTION BELONGS HERE THAT THE NUMBERS THEMSELVES DO NOT CARRY: A SINGLE EXCURSION AND A SUSTAINED ELEVATION ARE NOT THE SAME CLINICAL OBJECT. One value above a threshold, drawn once, under incompletely documented pre-analytical conditions is a different exposure definition from two or more confirmed values above it over months — and §9.2 shows the choice between those two definitions moving a published prevalence from 24% to 11% in the same cohort. ⚠ No study identified in this review quantifies cumulative time above threshold, so the evidence cannot tell a clinician how much that distinction matters. What it can say is that the apparatus’s numbers do not distinguish them, and that the studies underlying those numbers do not either.

A service’s answer is shaped by things that are not evidence at all. Whether a phlebotomy service is available. Whether switching preparation is fundable. Whether the person can attend for repeat testing. ⚠ Where less disruptive responses are unavailable, the practical consequence of crossing a threshold may be treatment interruption even where the source guidance permits several management options.

⚠ Reference-range interpretation is also exposure-dependent: the interval discussed at §6 was derived in people receiving stable testosterone and should not automatically be generalised to lower-dose regimens or to people who continue to menstruate.

⚠ Modification of gender-affirming hormone therapy is not a clinically neutral response to a laboratory threshold. Current trans-specific professional guidance frames dose reduction or interruption as a shared decision and recommends a harm-reduction approach where treatment changes are being considered.[5] That principle bears on how a threshold crossing is managed with a patient; it is not evidence about where the numerical threshold itself should be set.

What §6 establishes bears on this, and it should not be overstated in either direction. No appraisal identified here grades the threshold-and-outcome question this review asks. ⚠ The numerical thresholds appear in guidance, including trans-specific guidance, but their individual derivations have not been reconstructed; no threshold-specific outcome validation in trans-masculine people was identified. That is not a finding that the thresholds are too low, too high, or unnecessary. It is a finding about what is known. A clinician who must act today acts under that uncertainty whatever this review says.

Where it leaves a reader depends on which error they more fear, and this review does not tell them which to fear. A raised haematocrit left unwatched in someone who turns out to be at risk. Or a treatment interrupted on a number that may carry no risk at all.

⚠ IRON STATUS BELONGS IN THE CLINICAL PICTURE TOO. Iron deficiency does not exclude testosterone-driven erythropoiesis. Menstrual blood loss may change after treatment begins, and The harms of repeated venesection themselves remain largely unstudied. ⚠ A clinically important downstream consequence is that iron-restricted erythropoiesis and microcytosis are possible after repeated venesection, such that haematocrit may no longer straightforwardly represent the underlying erythropoietic drive; a rising haematocrit may be replaced by a picture in which iron depletion limits red-cell production. ⚠ This review prescribes no testing; it observes that iron status belongs to the clinical context when the phenotype or the management history suggests it.

⚠⚠ AND OTHER CAUSES NEEDS SAYING RATHER THAN GESTURING AT. A raised haematocrit in someone taking testosterone is not automatically caused by the testosterone, and the literature reviewed here investigates that badly: the systematic review of drug-induced erythrocytosis found that formal assessment of alternative causes was rarely reported.[18]

Depending on the clinical picture, assessment may take in: repeat measurement under stable hydration and sampling conditions, since relative haemoconcentration produces the same number ⚠ — hydration is the limb with a documented mechanism here, and time of day is recorded as a sampling condition worth standardising rather than as an established determinant of threshold classification, because no source examined here establishes that its effect is large enough to move a value across a cut-off; smoking and carbon-monoxide exposure, which is the strongest single predictor of haematocrit in one of the general-population cohorts examined here;[68] obstructive sleep apnoea and chronic hypoxic pulmonary disease; altitude; other drugs associated with erythrocytosis, including SGLT2 inhibitors, which account for a substantial share of the drug-induced erythrocytosis literature and of the secondary-erythrocytosis cohorts;[18][64] renal or other erythropoietin-driven causes where clinically indicated; and investigation for a primary myeloproliferative disorder where the phenotype warrants it.

⚠ This review does not prescribe a work-up. Which of these applies is a clinical judgement about a particular patient. What the review can say is that the evidence base it has examined largely did not make these distinctions, so a prevalence figure for erythrocytosis on testosterone is not the same quantity as a prevalence figure for erythrocytosis caused by testosterone.

Both errors are real, and they trade off imperfectly rather than one-for-one. ⚠ Repeat measurement and assessment of sampling conditions and alternative causes can reduce misclassification. Dose or preparation changes can lower the haematocrit in some settings. ⚠⚠ Whether those strategies reduce thrombotic events, or improve the overall balance of benefit and harm, is not established. What the evidence in Part V cannot say is which error is the more common, or how far any of those strategies reduces either.

Confidence: not applicable — this section describes a policy environment, not a finding.

12. The bottom line — REBUILT

What is firm, and it is more than it was. Testosterone raises haemoglobin and haematocrit in trans-masculine people by about four and a half percentage points over the first year — pooled across seven studies and 312 people, with low heterogeneity, ⚠ with a prospective 192-person cohort reporting a similar estimate; because its inclusion in the pooled set has not been checked here, it is not counted as independent confirmation. ⚠ An active-comparator randomised trial found rises in both testosterone-formulation arms; experimental identification against no testosterone comes from placebo-controlled trials in men outside gender-affirming care, not from that trial. ⚠⚠ This is the best-evidenced directly measured statement in the review. ⚠ It is not the only direct one — §10 marks eight rows direct — but it is the strongest. In the prospective longitudinal cohort evidence reviewed here, the largest increment occurred during the first three months of testosterone therapy.

The average rise does not, by itself, place a typical participant near the intervention thresholds. Individual participants nevertheless do cross them. In 192 prospectively followed trans men, four people reached at least 52%; one reached exactly 54·0%, and no value above 54·0% was reported. In the retrospective cohort of 511 patients, twelve (2·3%) had a recorded haematocrit of at least 54%.

What is better characterised than the argument allows. ⚠⚠ Reported frequencies cannot be interpreted as directly comparable cohort risks without aligning thresholds, observation periods, ascertainment and analysis. Some studies count participants with any observed crossing, others describe a single sampled value, and one estimates time-specific cumulative risk with censoring. The relative contributions of these differences and of underlying population or treatment differences have not been established (correction 568). ⚠ An earlier version of this review treated the spread as substantive.

And the mechanism is not settled. The human datasets disagree on erythropoietin, and on the best-characterised alternative route, clamping hepcidin constitutively high prevented the response in one genetic model while eliminating hepcidin did not prevent it in another — contributory but non-essential in at least some experimental settings; its quantitative contribution in humans is uncertain.

Two things the review can now answer, and both answers are the opposite of the intuition.

⚠ Haematocrit cannot be assumed to quantify absolute erythrocytosis accurately. In the subset of one secondary and idiopathic polycythaemia referral cohort who underwent nuclear red-cell-mass measurement, it explained eight per cent of the variance; a randomised trial and a prospective cohort make the same point in their own words. ⚠⚠ The corresponding relationship has not been measured in testosterone-treated trans-masculine people, and an earlier version of this section generalised it to them. That was an overreach and is withdrawn.

⚠⚠ AND A DEEPER POINT THAT UNDERCUTS THE FRAMING, NOT ONLY THE MAGNITUDE. Haematocrit is itself a concentration measure and a major determinant of whole-blood viscosity — which also depends on shear conditions, plasma composition, and erythrocyte deformability and aggregation. ⚠⚠ Haematocrit is not viscosity, and an earlier version of this passage came close to identifying the two. Weak agreement with total red-cell mass therefore does not by itself invalidate haematocrit as the hypothesised exposure, and this review no longer argues that it does.

⚠ And serum testosterone is a poor predictor of the blood count between individuals at steady state — under three per cent of the between-person variance in 6,670 people, and no relationship within a cohort of 79 on stable therapy. ⚠⚠ That constrains the use of a single contemporaneous serum testosterone concentration as a proxy for an individual’s haematocrit. It does not establish how effectively changing that individual’s dose changes the blood count — §9.5’s management evidence bears on that, and these correlations do not. ⚠ And it says nothing about causation: an earlier version of this section used those figures to argue the treatment effect itself was weak, which contradicts the ROBUST grade this review gives it from randomised and longitudinal evidence. The argument is withdrawn.

⚠⚠ AND THE DIRECT MECHANISTIC EVIDENCE IS HETEROGENEOUS BY ASSAY FAMILY. IT IS NOT SUMMARISED IN ONE DIRECTION, AND AN EARLIER VERSION DID SUMMARISE IT IN ONE DIRECTION. Larger plasma-based global analyses lean neutral-to-anticoagulant: in 348 trans men, endogenous thrombin potential or peak thrombin generation fell after six of seven modalities and the lag time lengthened, and in a nested analysis of 320 of the same participants, fibrin formation decreased and clot lysis increased.[26][58] Conventional coagulation-factor work runs the same way in a much smaller and older cohort: in 14 trans men over four months, activated-protein-C resistance fell and protein S rose, and the authors conclude a mild antithrombotic effect.[48] In 47 trans men on transdermal gel, no statistically significant changes were detected in the measured platelet-activation and coagulation markers at twelve months; the estimates do not establish equivalence or exclude potentially relevant changes ⚠⚠ while high-sensitivity C-reactive protein ROSE by 71% (95% CI 19; 145), which runs the other way and is reported at equal weight.[57] In smaller separate work, Factor VII falls by 15%, the international normalised ratio lengthens slightly, and no statistically significant changes were detected in fibrinogen, aPTT, PAI-1, D-dimer, Factor VIII or antithrombin.
>
> ⚠⚠ THESE ARE ONE OVERLAPPING AMSTERDAM/GHENT ASSAY PROGRAMME AND NOT INDEPENDENT REPLICATIONS: 348, 320, 47 and 14 are not 729, and the first two are one dataset analysed twice. ⚠ §9.4.6 carries the overlap map, and participant-level overlap has not been reconstructed. ⚠⚠ AND A NEW EVIDENCE CLASS BEARS ON THE REVIEW’S CENTRAL QUESTION FROM OUTSIDE THIS POPULATION. In drug-induced erythrocytosis, where an agent raises haematocrit and thrombotic outcomes are counted, three designs bear on the question. The randomised analysis finds SIGNIFICANT TREATMENT-EFFECT HETEROGENEITY BY BASELINE HAEMATOCRIT IN MALES, holding under categorical and continuous representations, while sparse erythrocytotic-stratum events, absent multiplicity adjustment and the absence of haematocrit randomisation or any mediation analysis prevent attributing that heterogeneity to the haematocrit itself;[59] ⚠ an earlier version summarised it by the subgroup null alone, which is the characterisation correction 120 rejects. Correction 125. a 6,787-patient cohort finds no gradient of thrombosis across the elevation range and attributes its events to established prothrombotic conditions;[60] and among people who had ALREADY had a thromboembolic event, those whose erythrocytosis had no identified cause showed better recurrence-free survival than those with no elevation, while those whose erythrocytosis had an identified cause did worst.[61] ⚠ Groups defined by whether investigation found a cause differ by construction, and the comparison is within an event-selected population. ⚠⚠ Read together they make underlying phenotype and comorbidity a PLAUSIBLE COMPETING EXPLANATION for the association between erythrocytosis and thrombotic events, and they do NOT establish either mediation by the haematocrit rise or its absence. ⚠ None randomises haematocrit; none performs a mediation analysis; the absence of a gradient does not exclude a non-linear, thresholded or treatment-obscured causal relation; and the Swedish cohort begins with people who have already had an event, so it cannot speak to incident thrombosis at all. The distinction §9.4.1 leaves open is therefore still open. ⚠ The mechanism differs from testosterone, the populations differ, and one of the cohorts did not evaluate testosterone use at all. Nothing transfers directly; the shape of the answer is what is carried.