Adversarial review. C1 minor_issues; C2 major_issues; C3 major_issues; C4 major_issues; C5 major_issues; C6 minor_issues; C7 sound.

The numerical analysis reproduces, but its composite outcomes do not identify the claimed numbers of unique participants. This is a substantive measurement problem rather than a numerical reproduction failure. All six supplied result JSON files (R1-R5 and S) equal a fresh isolated rerun. The rerun used Python3.12, NumPy2.1.3 and SciPy1.14.1, not the declared newer library versions; it is therefore corroboration in an alternate environment, not certification of that exact environment. Execution took about one second with no network or host mounts. All575 downloaded files were checked against supplied SHA256 and byte counts. I read the paper, claims, analysis and summarization code, outcome definitions/codebook, and inspected the structured source records implicated by the attacks. I did not independently manually adjudicate all932 trials or4029 terms. No author identity was sought or learned.

PRIMARY COUNTEREXAMPLE (C2,C4): code/analyze.py count() sums participants affected across terms and caps only at participants at risk. Different terms may refer to the same person. More decisively, NCT02942004 group EG001 has two included serious-suicidality term counts of1 each but only1 participant with ANY serious adverse event. The composite cannot contain2 unique participants. NCT01226511 group EG000 has the same contradiction in the pediatric analysis. These are internal contradictions with the supplied registry records, not a speculation about unobserved overlap. C2's82 drug participants and C4's38 drug participants cannot be unique-participant counts if the source totals are correct. At most81 and37 respectively are consistent with those totals.

The attached audit derives per-original-event-group set-union bounds directly from seriousEvents.stats.numAffected, respecting seriousNumAffected. Lower=max(largest included term count, overall serious participants minus the sum of all excluded term counts); upper=min(sum of included term counts, overall serious participants). Bounds are summed only across disjoint treatment groups. Repeating event-group IDs across adult comparisons was checked and none were found. This avoids mistakenly allowing overlap across different arms or ignoring constraints supplied by non-outcome terms. Conditional on the recorded totals and term coding, the count bounds are:
P1 adult behaviour: drug43-46, placebo16-19.
P2 adult any suicidality: drug71-81, placebo36-43.
A6 depressive-disorder behaviour: drug34-37, placebo11-12.
A18 pediatric behaviour: drug13-14, placebo4-5.
A19 pediatric any suicidality: drug34-37, placebo7-8.

Plugging low drug/high placebo unions into the submitted MH estimator gives P1 RR1.76244, CI0.999622-3.10739,p0.050153; P2 RR1.29285,CI0.872303-1.91615,p0.200745; A19 RR2.15291,CI0.998116-4.64377,p0.050565. These are sensitivity configurations, not recovered patient data, new confidence bounds for an unidentified estimand, or proof of no drug effect. They show that the claimed nominal significance need not hold for unique participants. Conversely the opposite overlap configurations can strengthen the signal; no particular correction direction is identified. The registered variance inflation changes uncertainty but cannot repair the unknown numerator or biased point estimate. Registration does not resolve this measurement issue. Report term-participant sums honestly or obtain deduplicated unions; provide identification bounds and rerun all corrected analyses. Merely relabelling sums as 'events' also requires an appropriate model for recurrent/correlated event counts.

C1 minor_issues: corpus and comparison counts reproduce, and the nausea positive control does not depend on the disputed suicidal composite. Nausea is available in207 selected comparisons (57,580 participants), not the whole adult sample of71,643. Explicitly attach that denominator to the nausea estimate and retain the threshold-selection caveat. A plausible positive control does not validate screening or every outcome classification. I found no reused event-group IDs across the adult comparison rows. I have not independently established full search completeness.

C2 major_issues: the strong counterexample and bounds above require changing the participant counts and re-evaluating estimates. The correctly reported failure of the original Holm adjustment is a strength and prevents a claim of established harm, but does not fix the headline counts. Also both displayed random-effects analyses have I-squared0; with the DL variance component at zero, the change from MH reflects inverse-variance weighting/continuity correction rather than evidence that between-trial heterogeneity explains shrinkage. Explain this estimator sensitivity directly.

C3 major_issues: the numerical interval-to-point comparisons are true of the original outputs, but inherit the composite-outcome defect. They compare all eligible adults aged18+ (including older adults) and serious-event-only outcomes to a25-64 population and different ascertainment. Excluding an earlier point estimate is not itself a calibrated test of a difference between uncertain estimates. Nor is posting results after2008 evidence of independent trials: ten included adult comparison records have primary completion before2008 (listed in audit), including NCT01621009 in2002-10 and NCT00285584 in2004-03. This does not prove overlap with FDA trials, but it disproves the stated justification for independence. Match trial identities against the FDA datasets, or withdraw the independence assertion. The illustrative two-estimate normal comparisons in the audit still yield small p-values under independence/common-estimand assumptions; I do NOT claim that accounting for historical uncertainty alone eliminates a difference. Those assumptions themselves require justification.

C4 major_issues: the pediatric82/38-type counting defect is directly observed for the38 count; see NCT01226511 above and A19 sensitivity. The supplied q0.076 is appropriately non-significant. Do not interpret the original marginally significant unadjusted estimate as a deduplicated risk. The behaviour subclaim also uses unidentified unions, although its non-significant conclusion persists in the audited bounds.

C5 major_issues: the37 versus12 behaviour numerator is a sum across terms, with union bounds34-37 versus11-12. The low-drug/high-placebo configuration gives RR2.03000,CI1.01191-4.07241,p0.04623, so this particular attack does NOT erase its nominal signal; it does establish that the specific participant counts/estimate are not uniquely determined. Recompute multiplicity adjustments after resolving the outcome definition. Ideation and completed-suicide components must be separately described; the small completed-suicide count cannot establish an absence of effect.

C6 minor_issues: the narrow descriptive absence of significance in the coded trial-level regression reproduces. Its explicit warning against interpreting it as evidence against a participant-level age gradient is appropriate. mean_age() falls back to medians; eleven adult trial IDs have only median age measures (audit), so call the regressor a baseline age summary and report which influential/event-bearing rows use medians. The composite measurement correction can change regression inputs and should be propagated. Do not turn this null into equivalence or an individual-level conclusion.

C7 sound within stated scope: the source overall seriousNumAffected field provides a union count, unlike the composite endpoints above. No adult included event group was missing that total. These counts and estimates reproduce. Death is restricted to121 comparisons with reported totals (20,743 participants), as described; make that denominator prominent. This is a registry subset estimate, not evidence of equivalence or long-term mortality effects.

Significance: C1 minor (useful curated reproducible corpus/control); C2 moderate (potentially useful safety synthesis if measurement is repaired); C3 minor (limited cross-study comparison); C4 moderate (clinically relevant confirmation subject to outcome repair); C5 moderate (relevant subgroup evidence); C6 minor (limited ecological null); C7 minor (descriptive registry safety estimates). No finding here supports clinical prescribing or stopping advice.

Sources for interpretation: ClinicalTrials.gov Results Data Element Definitions, https://clinicaltrials.gov/policy/results-definitions (government submitted PDF also available at https://omb.report/icr/202302-0925-001/doc/128670601.pdf), distinguishes per-term affected counts from overall participants with one or more serious events. Stone et al., BMJ2009;339:b2880, https://www.bmj.com/content/339/bmj.b2880, reports age-stratified estimates; the present paper itself also specifies the incompatible age/ascertainment scopes. The main counterexamples require only the bundle's own raw registry records. Audit limits: term coding and eligibility retained as supplied; no participant-level adjudication, no independently reconstructed full historical trial overlap map, and no claim that all possible biases have been excluded.