Every study, searchable

How sensitive is evidence for recent global warming acceleration to the final years included and the model of natural variability? A registered audit uses a frozen NASA global annual temperature series. A fixed-breakpoint model estimates a warming-rate increase of 0.232 °C per decade, with a confidence interval from 0.106 to 0.358. A search over possible breakpoints gives p = 0.0271 with fitted autocorrelation, but p = 0.4078 when the recent endpoint years are excluded. Stronger assumed autocorrelation also weakens the test. These are conditional statistical results about an unadjusted series, with limited simulated power for smaller slope changes.

• Published, Reproduced, Reviewed.Published, Reproduced, Reviewed: In the frozen NASA GISTEMP annual series for 1970-2025, a fixed-2015-knot model estimates a warming-rate increase of 0.232 degrees Celsius per decade with a 95% HAC interval of 0.106 to 0.358, while the registered breakpoint-search p-value is 0.0271 through 2025, 0.4078 through 2022, and 0.1726 through 2025 when the AR(1) lag coefficient is fixed at 0.6.

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.

How sensitive is a focused meta-analysis of walking and jogging for depressive symptoms? A registered audit of a fixed public extraction pools 21 direct-comparison trials and 994 participants. Post-treatment Hedges g is -0.762, with a confidence interval from -1.203 to -0.321. Every leave-one-trial-out interval excludes zero, but the usual-care-only interval and the between-study prediction interval include zero. No included trial has an overall low-risk-of-bias rating. These conditional computations support neither a uniform treatment effect nor a treatment recommendation.

• Published, Reproduced, Reviewed.Published, Reproduced, Reviewed: In the fixed Noetel et al. public extraction, the preregistered post-treatment direct comparison of walking/jogging with its specified active controls pools 21 trials and 994 participants at Hedges g=-0.762 (95% modified Hartung-Knapp confidence interval -1.203 to -0.321), all 21 single-trial-omission intervals remain below zero, but the overall prediction interval (-2.613 to 1.088) and usual-care-only confidence interval (-1.650 to 0.237) include zero and the overall-low-risk-of-bias subset has no trials, conditional on unchanged source outcomes and ratings.

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: 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.
• 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.

We provide an explicit rational parameter for every finite iteration cutoff. Its critical orbit remains uniformly below the usual escape radius throughout the cutoff, although an elementary growth argument proves that it eventually escapes. The contribution is a deterministic adversarial test family, a short uniform proof, and a certified finite benchmark. Integer enclosures and Arb ball arithmetic agree on 7 cases, with first-escape counts [23,61,212,815,3228,12879,51482]. Slow escape near the real parabolic cusp is established in the literature; we claim no discovery of that phenomenon or of its asymptotic law.

• Published, Reviewed.Published, Reviewed: For every integer T >= 1, the real parameter c_T = 1/4 + 1/[16(T+1)^2] has critical iterates z_0 = 0 and z_(n+1) = z_n^2 + c_T in [0,17/32] for 0 <= n <= T, but an unbounded critical orbit.
• Published, Reproduced, Reviewed.Published, Reproduced, Reviewed: The exact-rational benchmark parameters generated from cutoffs [1, 4, 16, 64, 256, 1024, 4096] have first critical-orbit escape iterations [23, 61, 212, 815, 3228, 12879, 51482], with all 7 counts certified and matched by integer dyadic enclosures and Arb ball arithmetic, and all 49 exact-rational control checks passing.

Published aggregate follow-up tables can support a transparent sensitivity calculation without reconstructing participant records or fitting the original trial model. We examined 6 long-term recovery rows from PRINCIPLE, keeping the primary and larger sensitivity cohorts separate. For the primary six-month row, the respondent-only treatment-minus-control recovery difference is 3.071 percentage points. Unrestricted binary completion of unavailable responses gives a range from -16.54 to 21.764 percentage points, whereas assigning both arms any common missing-response recovery rate gives a positive range from 2.329 to 2.896 percentage points. These are conditional bounds, not confidence intervals. They quantify how assumptions affect a crude contrast and do not identify a clinical treatment effect.

• Published.Published: In six published PRINCIPLE long-term recovery rows, unrestricted binary completion bounds permit either sign of the crude recovery contrast, every common missing-response recovery rate preserves a positive contrast, and at six months in the primary cohort a missing-treatment recovery rate 13.206 percentage points below the respondent-control anchor erases that contrast, conditional on the stated completion model.

Can unknown overlap between adverse-event terms change the direction of published composite-event estimates? We reconstructed the frozen registry comparisons and calculated every possible range of distinct affected participants consistent with the selected term counts. For adult serious suicidal behaviour, the resulting Mantel–Haenszel risk-ratio bounds are 1.712 to 2.1962; for adult serious suicidality, 1.2154 to 1.8476. The corresponding pediatric bounds are 1.0045 to 1.4773 and 1.8144 to 2.8091. All remain above the null. Unknown overlap therefore changes the magnitude but cannot reverse these fixed-data point estimates under the source's cohort and outcome definitions. These are identification bounds, not confidence intervals or estimates of causal treatment effects.

• Published, Reproduced.Published, Reproduced: In the frozen antidepressant registry dataset, conditional on source cohort selection, denominators, term coding, and absent-term imputation, sharp Mantel-Haenszel risk-ratio bounds relative to selected within-group event-count marginals are 1.7120261533 to 2.1961879882 for adult serious suicidal behaviour, 1.2154409158 to 1.8475901940 for adult serious suicidality, 1.0045047792 to 1.4773331886 for pediatric serious behaviour, and 1.8144436696 to 2.8090979871 for pediatric serious suicidality. Thus unknown participant overlap alone cannot reverse the direction of these fixed-data point estimates; these are not confidence intervals or causal treatment effects.