Every study, searchable

Foster and Rahmstorf (2026) remove El Niño, volcanic and solar influences from five global temperature records with a regression fitted to those records. They then report high confidence that warming accelerated, using critical values that treat the adjusted series as raw data. We ran their whole procedure, adjustment and tests together, on 2000 records per dataset, simulated under their own null: a straight-line trend since 1970 plus stationary noise. With noise fitted under that null, their changepoint test at nominal level 0.05 rejected 0.059 to 0.07 of null records, and the observed statistics had full-pipeline p of 0.0005 to 0.0025 in every record. Their quadratic test fared worse, with full-pipeline p up to 0.054. Under long-memory noise the changepoint evidence weakened, to p of at most 0.04, but held. The acceleration finding rests on the changepoint test.

• Published, Reproduced, Reviewed.Published, Reproduced, Reviewed: A numpy port of Foster and Rahmstorf's 2026 adjustment for El Niño, volcanic and solar influences reproduces their published adjusted NASA, NOAA, HadCRU, Berkeley and ERA5 series to within 0.0001 °C, and the changepoint test of Beaulieu et al. (2024) on the adjusted annual means for 1970-2024 gives maximum slope-change statistics from 4.91 to 6.68 with the change in 2013 or 2014 and nominal AR(1) Monte Carlo p-values from 0.00014 to 0.0033, against 0.041 to 0.19 for the unadjusted series.
• Published, Reproduced, Reviewed.Published, Reproduced, Reviewed: When 2,000 records per dataset are simulated under Foster and Rahmstorf's null hypothesis (a linear trend since 1970, their fitted exogenous terms, and ARMA(1,1) noise fitted under the null) and passed through their whole adjustment and changepoint test, the nominal 5% test rejects 5.9% to 7.0% of null records, and the observed statistics have full-pipeline p-values from 0.0005 to 0.0025 in all five records, so the changepoint finding is robust by the pre-registered rule.
• Published, Reproduced, Reviewed.Published, Reproduced, Reviewed: Under the same full-pipeline null, Foster and Rahmstorf's quadratic test with the ARMA(1,1) correction of Foster and Rahmstorf (2011) rejects 5.8% to 8.0% of null records at nominal 5% on monthly values and 7.2% to 9.2% on annual means, and the observed quadratic terms have full-pipeline p-values from 0.0015 to 0.054 (monthly) and 0.0005 to 0.032 (annual), below 0.02 in only three records for monthly values and two for annual means, so the reported confidence above 98% in all five records does not hold and the finding is mixed by the pre-registered rule.

1 more claim

Ketamine relieves depression within a day, but patients feel its psychoactive effects, so its benefit could depend on that experience. We asked whether ketamine and esketamine lower depressive symptoms when every dose is given under general anesthesia. Under a registered plan, we coded randomized trials' dosing from their Methods alone and pooled effects (Hedges' g) one to three days after surgery or childbirth. In 23 masked trials (3202 participants), depressive symptoms were lower with the drug than with an inert control (g -0.61, confidence interval -0.81 to -0.4; Holm-adjusted p < 0.001), though trials varied widely. Across trials in surgery, awake dosing did not show a larger effect (difference 0.12, -0.24 to 0.48), but this compares trials, and the one trial that randomized timing favored awake dosing. Masked doses also lowered pain. Ketamine can lower depressive symptoms after surgery without reaching an awake patient; whether experiencing the drug adds to that, these trials can't say. This version corrects counts, checks, and a reading that the first version's reviews found wrong.

• Published, Reproduced.Published, Reproduced: In 23 randomized trials (3,202 participants) that gave every dose of ketamine or esketamine under general anesthesia, between its induction and the end of surgery or cesarean delivery, depressive symptoms 1 to 3 days after the procedure were lower than with an inert control (Hedges' g −0.61, 95% CI −0.81 to −0.40; Holm-adjusted p < 0.001), though the trials varied widely (95% prediction interval −1.51 to 0.29). The difference remained among the 17 of these trials that passed the pre-registered integrity screen (g −0.48, −0.63 to −0.33), among the 13 whose last dose is known to have ended more than 30 minutes before the end of surgery (g −0.67, −1.04 to −0.30), among the 19 whose doses all began after induction rather than with it (g −0.60, −0.85 to −0.36), and with any single trial left out (largest p < 0.001).
• Published, Reproduced.Published, Reproduced: Among trials in surgery (37 comparisons: 22 masked, 15 with some dosing while patients were awake), trials that dosed patients while they were awake did not show a larger effect on depressive symptoms 1 to 3 days after surgery than masked trials: allowing for trials that enrolled only depressed patients, the coefficient of awake dosing in a meta-regression across trials was +0.12 (95% CI −0.24 to 0.48; p = 0.51), and the pooled effects in surgery were g −0.62 with masked and −0.45 with awake dosing. The comparison is between trials, which also differ in drug, dose, procedure, and patients, not between patients randomized to one timing or the other.
• Published, Reproduced.Published, Reproduced: The masked doses were pharmacologically active: masked ketamine lowered pain scores on day 1 or 2 after the procedure (g −0.55, 95% CI −0.97 to −0.14; 22 trials; g −0.38, −0.54 to −0.22 without the one trial whose printed dispersions can't be standard deviations), as awake ketamine did (g −0.33, −0.54 to −0.11; 32 trials; awake minus masked in surgery +0.32, −0.21 to 0.86). Masked ketamine still raised the odds of each trial's most frequent psychotomimetic adverse event (Mantel–Haenszel odds ratio 2.11, 1.48 to 2.99, in 22 trials), and awake ketamine raised them more (3.63, 2.97 to 4.43, in 40 trials): p = 0.0083 for the difference in the registered comparison, and p = 0.21 when the odds ratio may vary between trials (ratio of odds ratios 1.69, 0.74 to 3.84).

4 more claims

Many papers relate one National Health and Nutrition Examination Survey (NHANES) variable to one condition without correcting for multiple testing (Suchak et al. (2025)). We sampled 40 associations, reproduced each on its paper's cycles, and under a registered plan re-ran it on 2021–2023. There, 14 replicated as computed, and 10 with the survey design where papers ignored it. Most tests lacked power, but were every association true, 19 would replicate; counts this low are rare (0.047, or 0.011 with the design), so many effects are smaller than published or absent. In 36 papers the computation reproducing the estimate differs in some respect from the one described; as described, 6 fall outside the published interval.

• Published.Published: Of 40 associations sampled at random from those eligible among 341 papers that each relate one NHANES variable to one health condition (eligible meaning, among other rules, that its full text was open and that NHANES August 2021 to August 2023 could build its exposure, outcome, and population), 14 (35%; exact 95% confidence interval 21% to 52%, which treats the associations as independent) replicated in that cycle, meaning their estimate there has the published sign and a Benjamini-Hochberg-corrected p below 0.05, with each analysis computed as its paper's was. Of the 40, 15 were analyzed unweighted, ignoring the survey design, as their papers were; with those 15 fitted with the survey weights and design, as the plan also registered, 10 of the 40 replicated (25%; exact 95% confidence interval 13% to 41%).
• Published.Published: At an unadjusted two-sided alpha of 0.05, which bounds the power of the corrected replication decision from above, 13 of the 40 replication tests had at least 80% power to detect the published effect, taken as true, in the 2021 to 2023 cycle, and 10 of those replicated (77%; exact 95% confidence interval 46% to 95%); at the Bonferroni level of 0.05 over 40, which bounds it from below, 8 had that power, and 5 of them replicated. With the variance of the analyses re-run unweighted taken from the survey design, 13 tests had that power at the unadjusted level, 12 of them among the 13 above, and 9 of them replicated.
• Published.Published: On the analysis scale (log odds ratio or regression coefficient), the 2021 to 2023 effects are a median 0.77 times the published ones over all 40 associations (distribution-free 95% confidence interval 0.32 to 1.05) and 0.91 times (0.32 to 1.16) over the 13 informative ones; the intervals treat the associations as independent. The median is 0.53 over the 15 analyses re-run unweighted and 0.79 over the others; with the registered weighted fits in place of the unweighted ones, it is 0.85 (0.35 to 1.08).

12 more claims

An observational atlas linked GLP-1 receptor agonists to dozens of outcomes beyond glucose, weight and the heart. We pooled the serious adverse events that 234 placebo-controlled trials, with 161309 participants, posted to ClinicalTrials.gov, coded blind to the results under a registered plan. Reports of serious bacterial infection (ratio 0.88, confidence interval 0.83 to 0.93) and pneumonia (0.87) were less frequent on the drugs, as were pneumonia counted in people and death. The reductions held under random effects, with any trial left out, if infection terms overlapped within participants as all serious events do in each comparison, and in trials not coded as recording only on treatment. Of the atlas's associations with enough events to test, the trials found three in its direction with no difference from it detected, contradicted none, and left most inconclusive, among them every neurological and psychiatric one; its large hazard of drug-induced pancreatitis concerns a subtype the trials don't separate.

• Published, Reproduced.Published, Reproduced: Of the 539 interventional studies of GLP-1 receptor agonists with results on ClinicalTrials.gov on 2026-10-12, 234 randomized trials compare a GLP-1 receptor agonist with placebo, in 257 comparisons with 161,309 participants in their adverse-event tables (93,765 on the drugs and 67,544 on placebo). In them, reports of serious gallbladder and biliary disease, counting a participant once for each of the outcome's terms they had, are more frequent on the drugs (Mantel-Haenszel ratio 1.38, 95% CI 1.21 to 1.57), and so are reports of serious nausea and vomiting (2.02, 1.52 to 2.68), as the trials' own publications and the drugs' labels lead one to expect.
• Published, Reproduced.Published, Reproduced: Reports of serious bacterial infection, counting a participant once for each of the outcome's terms they had, are less frequent on GLP-1 receptor agonists in these trials (2,409 among 93,765 participants against 2,532 among 67,544 on placebo; Mantel-Haenszel ratio 0.88, 95% CI 0.83 to 0.93, q < 0.001), and so are reports of serious pneumonia (1,442 against 1,568; 0.87, 0.81 to 0.93, q = 0.0020); both stay significant under random effects and with any single trial left out. Counted in participants, pneumonia under its own preferred term, which no group reports in more than one row, is less frequent (807 against 880; 0.86, 0.79 to 0.95). If infection terms overlapped within a participant as all serious terms do in the same arm of the same comparison, bacterial infection would be 0.89 (0.82 to 0.97).
• Published, Reproduced.Published, Reproduced: In the registered scan without hypotheses of 27 organ systems and 518 preferred terms, with the same variance as the other claims, four organ systems (cardiac disorders; infections and infestations; metabolism and nutrition disorders; respiratory, thoracic and mediastinal disorders) and two preferred terms (acute myocardial infarction and cellulitis) are lower on the drugs with q < 0.05 in a random half of the trials and p < 0.05 in the held-out half, and one organ system, hepatobiliary disorders, is higher (1.36 in the first half, 1.24 in the second), as the gallbladder positive control leads one to expect. Among those lower are serious infection reports as a whole (0.88 and 0.89) and cellulitis (0.64 and 0.75); cardiac disorders and myocardial infarction are events many trials record as adjudicated end points rather than in these tables.

6 more claims

Whether antidepressants raise suicidal behaviour in adults is disputed: the FDA's pooled trials found no increase past young adulthood, and re-analyses of approval data on newer drugs found more suicide attempts. We tested this in the serious adverse events placebo-controlled trials posted to ClinicalTrials.gov, under a registered plan. This version corrects the first, which counted a participant listed under two of an outcome's terms twice. With the posted tables and the trials' public reports, serious suicidal behaviour affected 43 adults on antidepressants and 17 to 18 on placebo (risk ratio 1.86 to 1.96): nominally significant, but at the least favourable count not after the registered correction for two endpoints. Without the trials the re-analyses also used, the excess is similar but less certain. Pediatric trials showed more serious suicidality on antidepressants, as the FDA found. These data lean toward more suicidal behaviour in adults but cannot establish it.

• Published, Reproduced.Published, Reproduced: Of the 932 interventional studies naming an antidepressant with results posted on ClinicalTrials.gov on 2026-10-13, 393 randomized trials compare an antidepressant with placebo in 413 comparisons with 99,730 participants at risk in their adverse-event tables. In the 306 adult trials that randomized participants from the start (71,643 participants), nausea is more frequent on antidepressants in the 207 comparisons whose non-serious tables list it (57,580 participants; Mantel-Haenszel risk ratio 2.08, 95% CI 1.98 to 2.18), as these drugs' known side effects predict.
• Published, Reproduced.Published, Reproduced: In those adult trials, the posted tables bound the participants with serious suicidal behaviour (completed suicide, suicide attempt, intentional self-harm or intentional overdose) at 43 to 46 of 40,564 on antidepressants and 16 to 19 of 31,079 on placebo, and those with any serious suicidality at 71 to 81 against 36 to 43, because a participant listed under two of an outcome's terms may be one person or two. Across every count the tables allow, the Mantel-Haenszel risk ratio is 1.76 to 2.20 for serious suicidal behaviour (p 0.0097 to 0.0502) and 1.29 to 1.73 for any serious suicidality (p 0.0078 to 0.20). At the counts least favourable to an increase neither is significant after the registered Holm correction across the two co-primary endpoints (adjusted p 0.10 and 0.20); at the most favourable both are (0.016 and 0.016).
• Published, Reproduced.Published, Reproduced: Public reports of the trials (their publications, regulators' reviews and the sponsors' documents) settle how many participants share terms in 11 of the 17 adverse-event groups the tables leave open and narrow it in 3 more. With them, serious suicidal behaviour affected 43 of 40,564 adults on antidepressants and 17 to 18 of 31,079 on placebo (Mantel-Haenszel risk ratio 1.86 to 1.96; p 0.025 to 0.036), and any serious suicidality 72 to 77 against 40 to 41 (1.38 to 1.51; p 0.041 to 0.12). After the registered Holm correction across the two, the adjusted p-values are 0.071 and 0.12 at the counts least favourable to an increase and 0.0496 for both at the most favourable.

6 more claims

Why did United States heart-disease deaths increase between 2018 and 2024? An updated analysis of final national death counts and published population estimates averages all six orders of a demographic decomposition. Known-age deaths increase by 28135. Population size contributes 25981.4 deaths, age composition 26512.2, and age-specific rates -24358.5. Demographic contributions outweigh the negative rate contribution. These are arithmetic contributions conditional on the source denominators, which change estimation methods during the period, rather than causal effects or individual mortality risks.

• Published, Reproduced, Reviewed.Published, Reproduced, Reviewed: In the bundled CDC WONDER D158 final snapshot for United States residents with known age, 2018-2024 heart-disease deaths increase by 28,135. Averaging all six orders of the population-size, age-share-vector, and age-specific-rate-vector decomposition yields contributions of +25,981.4, +26,512.2, and -24,358.5 deaths, respectively (display-rounded), so positive demographic contributions outweigh the negative rate contribution. The statement is conditional on the published population denominators and specified age bins, and is arithmetic rather than causal or an individual clinical-risk estimate.

People with depression report that time drags, yet laboratory timing studies disagree. A temporal account of depression predicts that time is dilated within the experienced present moment (targets up to a few seconds) and accelerated beyond it. In a time production task, dilation means producing intervals that are too long and acceleration producing them too short. In a registered meta-analysis of 11 case-control production studies, re-extracted from the primary papers where possible, the depression effect was 0.4836 standardised units higher above the 3 s boundary than below it (one-sided p = 0.03380082115675086), with overproduction at short targets (g = -0.2851) and underproduction at longer ones (g = 0.1985), neither significant alone. The effect rose by 0.6302 per tenfold increase in duration. Re-extraction also showed that an earlier meta-analysis coded a 30-s production as longer than 10 min.

• Published, Reproduced, Reviewed.Published, Reproduced, Reviewed: Across 11 case-control studies of time production in depression, the pooled standardised difference (positive = depressed participants produce shorter intervals than controls) is 0.4836 higher for target intervals above 3 s than for targets of 3 s or less (95% CI -0.0431 to 1.0103; one-sided p = 0.034), with pooled g = -0.2851 at 3 s or less (3 studies; 95% CI -0.6917 to 0.1214) and g = 0.1985 above 3 s (10 studies; 95% CI -0.1378 to 0.5348).
• Published, Reproduced, Reviewed.Published, Reproduced, Reviewed: Across the per-interval effects extracted from 7 primary papers (20 target intervals from 1 to 240 s), the depression effect on time production rises by 0.6302 standardised units per tenfold increase in target duration (cluster-robust SE 0.2128; one-sided p = 0.016), from overproduction at the shortest targets toward underproduction at the longest.
• Published, Reviewed.Published, Reviewed: The time production task of Kitamura and Kumar (1983) asked participants to produce 30 s, but the meta-analysis of Thones and Oberfeld (2015, Table A2) coded it as a long (over 10 min) interval, where it was the only production effect; no case-control production study of depression in that meta-analysis or since has used a target over 10 min.

An account of depression as accumulated physical load distributed across body systems predicts that each system's state tracks the symptoms it produces, that bodily measures relate to somatic symptoms beyond mood, and that socially distributed load appears somatically. We tested these predictions in a registered analysis of 14722 adults in the US National Health and Nutrition Examination Survey (NHANES), relating limitation in load-bearing tasks, slow bowel transit, body mass index (BMI), an allostatic load index and income to PHQ-9 depressive symptom items. Load-bearing limitation, BMI and allostatic load related to somatic symptoms beyond cognitive-affective ones, and more than to cognitive-affective symptoms beyond somatic ones. But the predicted system-specific pattern did not appear: load-bearing limitation tracked sleep, energy and appetite (0.2946) far more than slow transit did (0.0404). Low income related more to cognitive-affective than to somatic symptoms (difference -0.0996). Bodily load is linked to somatic depression generally, not system by system.

• Published, Reproduced, Reviewed.Published, Reproduced, Reviewed: In the same adults, load-bearing limitation, body mass index and a nine-marker allostatic load index are each associated with somatic depressive symptoms after adjusting for cognitive-affective symptoms (partial Spearman 0.1916, 0.0558, 0.08) and more strongly than with cognitive-affective symptoms after adjusting for somatic ones (differences 0.0711, 0.0399, 0.0646; Holm-adjusted p 0.002, 0.016, 0.002), whereas slow bowel transit shows no such specificity (difference 0.0074, 95% CI -0.0209 to 0.0347).
• Published, Reproduced, Reviewed.Published, Reproduced, Reviewed: In 14722 NHANES 2005-2010 adults, limitation in load-bearing tasks is associated with the PHQ-9 psychomotor item more than slow bowel transit is (age-, sex- and cycle-adjusted partial Spearman 0.2038 vs 0.0296), but also with the sleep, energy and appetite items more than slow transit is (0.2946 vs 0.0404; difference -0.2542, 95% CI -0.2762 to -0.2312), so the predicted double dissociation between a structural and a transport system does not appear.
• Published, Reproduced, Reviewed.Published, Reproduced, Reviewed: Lower family income relates more to cognitive-affective than to somatic depressive symptoms in these adults (partial Spearman 0.1275 vs 0.0279; difference -0.0996, 95% CI -0.1299 to -0.0695), opposite to the prediction that socially distributed load expresses itself somatically.

Embodied accounts of affect propose that position, weight, tension and shape are bodily templates for mood and evaluation: down and heavy for sadness, up and light for happiness, tension for anger, crookedness for badness. If such mappings are deep, many unrelated languages should use one word for both senses. In two registered analyses of the CLICS4 colexification database, "light" and "easy" share a word in {{R2.pairs.LIGHT (WEIGHT)|EASY.families_colexifying}} of {{R2.pairs.LIGHT (WEIGHT)|EASY.families_with_both}} language families, mirroring "heavy" and "difficult", and "straight" shares a word with "correct" in {{R1.pairs.STRAIGHT|CORRECT (RIGHT).families_colexifying}} families. By contrast, "up" or "high" and "happy", "down" or "low" and "sad", and "light" and "happy" share a word in at most one family each, and "hard" never shares one with "angry". Across languages, weight is lexicalised as effort and straightness as rightness, but emotion is kept lexically apart from posture and weight.

• Published, Reproduced, Reviewed.Published, Reproduced, Reviewed: In CLICS4 v1.0, one word means both 'light (in weight)' and 'easy' in 11 of 50 language families with words for both (rate 0.22), more often than 'light' pairs with any of its reference concepts, and 'heavy' and 'difficult' share a word in 7 of 57 families, so weight words double as words for ease and difficulty at both poles.
• Published, Reproduced, Reviewed.Published, Reproduced, Reviewed: 'Straight' shares a word with 'correct' in 38 of 155 families and with 'good' in 7 of 200, whereas 'crooked' shares a word with 'wrong' in 4 of 113 and with 'bad' in 1 of 171, so the straightness-evaluation link is lexicalised mainly at its positive pole.
• Published, Reproduced.Published, Reproduced: Words for vertical position or weight almost never double as words for happiness or sadness: up-happy 1 of 66 families, high-happy 0 of 145, down-sad 0 of 21, low-sad 0 of 108, light-happy 0 of 146.

1 more claim

Accounts that treat the felt weight of depression as more than metaphor rest on a premise: that distress is described in gravitational terms across unrelated languages. We tested a lexical form of that premise in a registered analysis of the CLICS4 database of cross-linguistic colexifications, asking whether languages use one word for "heavy" and for difficulty, grief, sadness or tiredness more often than for other concepts, and whether gravity-related words share words with sadness more than other physical-property words do. "Heavy" and "difficult" share a word in 7 of 57 families, more often than "heavy" pairs with any of its 1374 reference concepts, and "heavy" and "grief" in 2 families. "Heavy" never shares a word with "sad" (zero of 156 families) or "tired", and gravity-related words show no detectable excess with sadness (permutation p = 0.2232). Lexically, weight is widely the word for burden, not for sadness.

• Published, Reproduced, Reviewed.Published, Reproduced, Reviewed: In CLICS4 v1.0 (3447 varieties, 247 families), a single word means both 'heavy' and 'difficult' in at least one language of 7 of the 57 families that have words for both (family-level rate 0.1228), higher than the rate for every one of 'heavy's 1374 reference partner concepts, whose 95th percentile is 0.0051; 'heavy' and 'grief' share a word in 2 of 69 families (rate 0.029, percentile 0.9964).
• Published, Reproduced, Reviewed.Published, Reproduced, Reviewed: In the same data, 'heavy' never shares a word with 'sad' (0 of 156 families) or 'tired' (0 of 73), and 'sad' or 'grief' shares a word with any of 15 physical-property concepts in at most 2 families each, so gravity-related words (heavy, low, down) do not colexify with sadness detectably more than other physical-property words (difference in family-level rates 0.0031, one-sided permutation p = 0.2232).

Psychological states are often described in mechanical terms: under pressure, crushed, torn, worn down, flowing, stuck, off course. If such mappings were deep and widespread, unrelated languages should often use one word for the mechanical and the psychological sense, as many use one word for "heavy" and "difficult". In a registered screen of the CLICS4 colexification database, we paired words for compression and tension, shear and friction, flow and momentum, and paths with concepts of emotion, cognition and effort. None of these pairs met the registered criterion of recurring across several families more often than the word's other pairings; each class had 0 hits against about 0.477 to 0.719 expected for random concepts. Only the load words reached it, through "heavy" with "difficult" and "light" with "easy" (2 hits, p = 0.0633). Across languages, weight is the one mechanical domain lexicalised as a psychological quantity.

• Published, Reproduced, Reviewed.Published, Reproduced, Reviewed: Weight words are the exception: 'heavy' and 'light (weight)' give 2 hits among five load words (with 'difficult' and 'easy'), against 0.304 expected for random concept sets, though the class-level excess does not reach the registered 0.05 level (p = 0.0633), so the class test has limited sensitivity.
• Published, Reproduced.Published, Reproduced: Across CLICS4 v1.0, none of 37 words for compression and tension, shear and friction, flow and momentum, or paths shares a word with any of 30 emotion, cognition or effort concepts in at least three language families at the 95th percentile of the word's other pairings; each of the four classes has 0, 0, 0, 0 hits, against a mean of 0.48 to 0.72 for random concept sets of the same size.

Does maintaining normal tissue mechanics prevent cancer initiation? A focused evidence map separates the endpoints of 4 selected papers and reanalyzes an available tumor-count table. The selected contrasts occupy 4 classes, with 0 papers testing incident cancer after a mechanical intervention in humans. The table concerns already malignant cells: blocking-antibody pretreatment gives risk ratios of 0.4667 and 0.5 against its controls. These endpoints answer different questions; neither a smaller transplantation risk nor a human biomarker change establishes primary prevention. This is a bounded, exploratory source audit, not a systematic review or a new biological experiment.

• Published, Reproduced, Reviewed.Published, Reproduced, Reviewed: Reanalysis of Weaver et al. (1997) Table II gives 7/16 tumor-bearing mice after blocking-antibody pretreatment of already malignant cells versus 15/16 untreated and 14/16 control-antibody mice, risk ratios 0.4667 and 0.5000, with two-sided Fisher p-values 0.005923 and 0.023333 (Holm-adjusted 0.011846 and 0.023333), while conservative simultaneous 95% risk-ratio intervals are 0.1612 to 1.1614 and 0.1625 to 1.3304 under independent-binomial sampling.
• Published, Reproduced, Reviewed.Published, Reproduced, Reviewed: In a purposively selected four-paper evidence map, the coded contrasts occupy four distinct endpoint classes, only one paper starts its reprogramming program with normal primary cells, and none measures incident cancer after a mechanical intervention in humans, with all counts restricted to this selected set rather than the literature.

How does transmission heterogeneity govern whether an emerging pathogen dies out or causes an epidemic? In negative binomial branching processes at reproduction number 2.5, increasing overdispersion from the Poisson limit to geometric variance to superspreading raises the extinction probability from 0.107355 to 0.4 to 0.860685. Epidemic certainty at 95 percent requires 2 introductions under Poisson transmission, 4 under geometric variance, and 20 under superspreading. Otter-Dwass evaluation establishes that solitary index cases account for 0.721943 extinction probability, or 83.88 percent of extinctions. Cluster size summation through 500 cases recovers the extinction probability with an absolute residual of 0, confirmed across 50000 simulations. Superspreading erects an extinction barrier against solitary introductions while concentrating epidemic risk in explosive clusters.

• Published, Reproduced, Reviewed.Published, Reproduced, Reviewed: In negative binomial branching processes with reproduction number R0 = 2.50, increasing transmission overdispersion from the Poisson limit (k -> inf) to geometric variance (k = 1.00) to SARS-like superspreading (k = 0.10) increases the single-case extinction probability from 0.107355 to 0.400000 to 0.860685, expanding the critical introduction threshold for 95% epidemic certainty from 2 to 4 to 20 independent introductions.
• Published, Reproduced, Reviewed.Published, Reproduced, Reviewed: For SARS-like superspreading dynamics (R0 = 2.50, k = 0.10), closed-form Otter-Dwass evaluation establishes that solitary index cases without secondary spread account for 0.721943 extinction probability (83.88% of all extinctions), and cumulative summation over finite cluster sizes s from 1 through 500 recovers the complete extinction probability with an absolute residual under 1.0e-6, verified across 50000 independent branching process simulations.

Adding floating-point numbers one after another loses a little at each step, and how much depends on the order of the additions. For sums of up to a million doubles, drawn uniformly from $[0, 1)$ and from $[-1, 1)$, this study measures how far four common methods land from the correctly rounded sum, in units in the last place (ulps). Adding left to right, the error averages 196.86 ulps for a million positive values and grows with their number $n$ as about $n^{\alpha}$ with fitted $\alpha$ = 0.545, close to the square root that independent rounding errors predict. Pairwise summation stays within 2 ulps at every length. Kahan's and Neumaier's compensated sums return the correctly rounded sum in all 400 draws, including mixed-sign sums that nearly cancel. Every value is reproducible exactly, on any machine and Python version.

• Published, Reproduced, Reviewed.Published, Reproduced, Reviewed: Adding 1000000 doubles drawn uniformly from [0, 1) from left to right lands a mean of 196.86 units in the last place (at most 589) from the correctly rounded sum over 50 draws, and the mean error grows as about n to the power 0.545 for n from 1000 to 1000000, near the square-root growth that independent rounding errors predict.
• Published, Reproduced, Reviewed.Published, Reproduced, Reviewed: Pairwise summation of the same positive draws stays within 2 units in the last place of the correctly rounded sum in all 200 draws for n from 1000 to 1000000, with a mean error of 0.4 units at n = 1000000.
• Published, Reproduced, Reviewed.Published, Reproduced, Reviewed: Kahan's and Neumaier's compensated sums equal the correctly rounded sum in all 400 draws, positive and mixed-sign, including mixed-sign sums of 1000000 values whose median condition number is 991.

1 more claim

Turning 35, and again turning 40, moves a pregnancy into the "advanced maternal age" category, which carries more surveillance and more advice to deliver early. Using every US singleton first birth at term from 2016 to 2024, we compared mothers just below and just above each age with a preregistered regression discontinuity, using confidence intervals that stay valid when age is recorded in whole years. Labor induction jumps by 1.19 percentage points at 35 (honest 95% CI 0.7 to 1.68). The preregistered verdict is inconclusive, because the share of first births also jumps at 35, by 0.4 points. At 40, induction jumps by 2.86 points (2.05 to 3.67), and that secondary test meets its confirm rule. Delivery at 39 weeks rises at both ages. Cesarean delivery, NICU admission, low Apgar scores, severe maternal morbidity and term stillbirth show no jump at either age. The label appears to move delivery timing without changing these measured outcomes.

• Published, Reproduced, Reviewed.Published, Reproduced, Reviewed: Among singleton first births at 37-44 weeks in US natality files for 2016-2024, labor induction rises by 1.19 percentage points at maternal age 35 (honest 95% CI 0.70 to 1.68, bias bound 0.1 pp per year squared), but the preregistered decision rule classifies this primary test as inconclusive because the all-parity nulliparity share also jumps at age 35 (0.40 pp, Holm-adjusted p = 0.023).
• Published, Reproduced, Reviewed.Published, Reproduced, Reviewed: In the same population, labor induction rises by 2.86 percentage points at maternal age 40 (honest 95% CI 2.05 to 3.67), with no covariate discontinuity at 40 after Holm adjustment, so the preregistered secondary test at 40 meets its confirm rule.
• Published, Reproduced, Reviewed.Published, Reproduced, Reviewed: Delivery at 39 weeks rises by 1.00 percentage points at maternal age 35 (honest 95% CI 0.52 to 1.48) and by 3.64 percentage points at age 40 (honest 95% CI 2.85 to 4.44) among singleton first births at 37-44 weeks.

2 more claims

How much does scheduling a daily batch by historical grid intensity help when computing capacity and idle power are explicit? We evaluate seasonal clock-time profiles trained on 2023 public Great Britain data, with equal work and equal energy in every comparison. At 50% utilization, assigned carbon falls by 11.63% in 2024 and 12.81% in 2025 relative to uniform processing. Setting idle power to 50% of maximum power reduces these advantages to 3.88% and 4.27%. Some dates have higher assigned carbon after shifting. These results describe a fixed historical workload benchmark and an accounting signal; they do not measure avoided emissions.

• Published, Reproduced, Reviewed.Published, Reproduced, Reviewed: In the frozen Great Britain daily-batch benchmark, seasonal profiles trained on 2023 data reduce assigned carbon versus uniform processing by 11.63% and 12.81% at 50% utilization with zero idle power in 2024 and 2025, but by 3.88% and 4.27% when idle power is 50% of maximum; at 87.5% utilization with zero idle power the reductions are 3.37% and 3.64%, while the half-utilization policy increases assigned carbon on 16 of 364 and 10 of 365 evaluation dates, respectively, conditional on the specified processor model and estimated average intensity data.

Do historical electricity schedules help when jobs arrive throughout the day and cannot wait indefinitely? We evaluated fixed seasonal profiles and recent-history profiles on public Great Britain grid records. For the primary workload, seasonal scheduling lowered mean assigned carbon intensity by 11.24% in 2024 and 12.01% in 2025 relative to immediate execution. Yet 17.61% and 15.55% of jobs received higher intensity. Recent-minus-seasonal mean differences were -0.21 and -0.07 g CO2/kWh. These are retrospective accounting results for a synthetic workload, not measured emissions avoided.

• Published, Reproduced, Reviewed.Published, Reproduced, Reviewed: Conditional on the frozen Great Britain national intensity snapshot and a uniform half-hour arrival workload, three-hour jobs allowed up to six hours of start delay have 11.24% and 12.01% lower mean assigned intensity under 2023-trained seasonal profiles than immediate execution across 17,424 and 17,472 jobs in 2024 and 2025, respectively, while 17.61% and 15.55% have higher intensity and a lagged recent-history profile changes the seasonal mean by -0.21 and -0.07 g CO2/kWh, without establishing causal emissions avoided or future workload performance.

Does adaptive therapy consistently outperform continuous maximum tolerated dose treatment? Simulating stochastic birth-death-mutation processes and competitive dynamics reveals a sharp bifurcation governed by baseline clonality. Under de novo mutation, continuous treatment achieves complete clonal eradication in 96.6% of simulated tumors, whereas adaptive therapy achieves 0% eradications due to mutational influx during drug holidays. Conversely, with 2% pre-existing resistance, adaptive therapy prolongs mean time to progression from 68.6 days to 74.8 days while reducing cumulative drug exposure by 18.1%. In competitive dynamics, the adaptive-to-continuous progression-time ratio contracts from 1.11 under a 20% fitness cost to 1.09 under neutral resistance. Thus, adaptive therapy requires established pre-existing resistance to provide clinical benefit.

• Published, Reproduced, Reviewed.Published, Reproduced, Reviewed: In the presence of 2.0% pre-existing resistant cells at baseline, adaptive therapy prolongs mean time to progression from 68.6 days to 74.8 days while reducing cumulative drug exposure by 18.1% compared to continuous therapy.
• Published, Reproduced, Reviewed.Published, Reproduced, Reviewed: Across deterministic competitive dynamics, the time-to-progression ratio of adaptive relative to continuous therapy contracts from 1.11 at a resistant fitness cost of 0.20 to 1.09 at zero fitness cost, confirming that carrying-capacity crowding alone provides only modest progression delay in the absence of a competitive proliferation penalty.
• Published, Reproduced.Published, Reproduced: Under de novo mutation without pre-existing resistance, continuous maximum tolerated dose therapy achieves complete stochastic clonal eradication in 96.6% of simulated tumors, whereas adaptive therapy achieves 0.0% eradications and incurs progression across all trials due to mutational influx during treatment holidays.