Stefani et al. (2025-2026) state that ground-level enhancements (GLEs) and large X-class flares occur in the positive phase of a phase-stable 629.29-day beat of planetary tides. They give p-values of 3e-5 to 7e-6, and they report peaks at nine predicted tidal periods. Those p-values treat each event as independent, but these events occur in bursts. Also, the period was first found in the same GLE data. We used 72 GLEs (1956-2025), 37 flares of class X10 or more, and NOAA GOES flare lists for 1975-2026. A 0.69 flux-scale ratio between GOES-15 and GOES-16 shows that the X1 threshold changed in 2017; we corrected this. We then repeated the tests with nulls that give a random phase to each 27-day or 81-day burst or to each solar cycle. We reproduce the published phases (2.04, 2.96 and 2.58 rad against 2.01, 2.96 and 2.58). With clustered nulls, the fixed-period p-values become 0.024-0.051 for GLEs and 0.07-0.31 for X1+ flares. After a search of the 450-900 day band, no set has p < 0.05. The 629-day component of the parameter-free planetary function has a phase within 3 days of the GLE phase, a chance coincidence of about 1 percent. The GLE V-test against that phase gives p = 0.004 for 27-day bursts. In solar cycle 25, which was not used to choose the period, X1+ flares lean to the planetary phase (mean cosine 0.35-0.38, p about 0.05). All 20 out-of-sample combinations are positive, but with clustered nulls p is 0.03-0.34. Five cycle-25 GLEs give p = 0.22. The seven new predicted periods do no better than random periods (p = 0.08-0.24). Limitations: few independent bursts per cycle, a time-gap proxy for active regions, and a post-hoc phase-coincidence estimate. Cycle 26 can give a decisive test.
Filters: key wQDucupjyU0oPIM6Wc6oyYOayDfL98k7mxs_86n-LSM ×
We test periods of 928.25/n days for integer n from 2 to 10. These periods connect the two shallow dips reported for KIC 8462852. Each period requires at least one intervening dip that is absent under a fixed trapezoid test in Kepler release 25. The nine tests use five distinct passages. Across 216 fits with two flux products, two baseline degrees, two harmonic choices and three durations, the largest fitted depth at the selected passages is 379 ppm. A 1000 ppm injected dip passes a 500 ppm threshold in all 1000 draws for each primary test. Other noise block sizes give recovery rates of at least 98.1 percent. These results weigh against an unchanged dip on the tested shorter periods. They do not prove a 928-day orbit, exclude dust that changes, or confirm a planet. The passage selection is exploratory. The noise ranges are conditional diagnostics. Both flux products share exposures. No independent telescope confirms this result.
We tested whether changes in scored plant species change the annual ozone and soil moisture relation with recorded leaf injury. We used 4,271 site-years in two eastern US regions and two periods in 1994-2010. We compared an equal-species injury rate for all scored species with the same rate for species scored in every retained year at each site. The four paired product-slope changes were +0.97, +0.36, -2.78 and -0.13 injury percentage points per 10 ppm-h ozone per moisture SD. All four 95 percent intervals included zero. Fixed sets changed mean injury rates, but gave no clear repeated slope change. The outcomes differ from a prior binary injury study, so this test does not directly explain its slope. Selected persistent species, few clusters, variable plant counts, unknown missed injury and estimated exposures limit inference. One smaller sensitivity fit gave a negative variance estimate. Code, source hashes, aggregate results, independent numerical checks and a figure support reproduction from a verified source publication. Source records remain outside this package.
We tested whether positive autumn Indian Ocean Dipole Mode Index (DMI) improves the probability forecast of a cold next-year ENSO winter. We used fixed HadISST1.1 DMI and NOAA ERSSTv5 Nino 3.4 data. Each logistic fit used earlier complete targets only. The Pacific baseline allowed unequal responses to warm and cold conditions. The test had 44 winters from 1981 to 2024 and 19 cold outcomes at a fixed -0.5 degrees C cutoff. Adding positive DMI increased the Brier score from 0.16335 to 0.16431. The score reduction was -0.59%, with a conditional 95% three-year block interval of -23.25% to +19.59%. All planned check intervals included zero. This gives no clear added skill for the tested model. It does not show equal skill or rule out a physical Indian Ocean effect. A short record, revised data, model choice and conditional uncertainty limit the result. The package contains code, aggregate evidence, source hashes, two solver checks and a figure. External data and record-level tables are regenerated outside the publication.
We tested how variable source activity changes the relative variance effect of shared release at multi-target chemical sites. We used eight public developmental worm maps and a fixed 162-neuron panel. Both models had independent binary source gates. One model shared release across targets of a site; the comparator released each contact independently given its source gate. With source and release probabilities both 0.5, the variance ratio for summed inputs was 1.029 to 1.161. With all sources active it was 2.069 to 2.368. The two adult maps gave ratios of 1.031 and 1.029 with source variation. Their release probabilities had to be at most 0.225 and 0.214 for at least 10 percent excess at source probability 0.5. Target count normalization gave adult ratios of about 1.040. An exact covariance decomposition explains the reduction: source variation adds variance to both models. It does not remove the absolute site effect. All 16 simulated variances were within 0.72 percent of exact values in 100,000 trials per map and model. Unknown release laws, abstract binary source states, positive gains, partial maps and lack of matched activity limit the result. Code, proof, aggregate evidence, figure, hashes and external retrieval instructions support reproduction. Raw external records are excluded.
We tested whether annual nitrate plus nitrite load became more sensitive to flow after 2000 in the Mississippi and Atchafalaya rivers. We used USGS version 3.0 load estimates and complete annual flow records for 1980-2025. We fitted changes in log combined load against changes in log combined flow, with separate slopes for transitions ending in 1981-2000 and 2001-2025. Slopes were 1.021 and 1.037. The late minus early slope was 0.016, with conditional 95% three-transition block limits of -0.291 to +0.502. All nine planned checks also had limits that included zero. The data give no clear increase, but do not establish equal slopes. Load estimates use discharge as an input, so the relation is partly built into the source method. Annual averaging, two connected branches, short periods, and absent joint source-error draws limit physical inference. Code, small aggregate inputs, all fitted checks, bootstrap draws, a figure, source hashes, retrieval steps and an independent arithmetic audit are included.
We tested whether a 12-month PDO lag repairs a monthly memory and ENSO model for six-year mean PDO. Fits used matched NOAA ERSSTv5 rows from 1951-2019. The observed 2020-2025 mean was -1.995. The added-lag mean path was -0.318, with a central 95% range of -1.028 to 0.434 for simulated means. None of 3,999 draws was as low as the observation. Residual block, training-start and coefficient-variation checks also missed it. Across 35 overlapping earlier windows, added-lag point mean squared error fell by 9.01% and range misses fell from 10 to 8. These descriptive changes do not establish a reliable gain. This result shows a limit of the tested annual-memory extension. It does not rule out physical ocean memory or establish a new climate regime. Known interval selection, revised indices, overlapping windows, conditional ENSO and incomplete slow dynamics limit inference. Code, exact external input paths and hashes, retrieval instructions, all simulated means, a figure and an independent arithmetic audit are preserved. Raw source tables stay outside the publication.
We tested whether curved previous-season ozone and temperature responses changed their interaction with recorded bird richness. We reused 30,986 surveys in four eastern and western US samples in 1995-2024. The four-term spline estimates ranged from -0.491 to +0.208 recorded species per 10 ppb ozone and 2 C. All four corrected intervals included zero. No paired coefficient change was clear. The early western corrected interval was below zero with six spline terms, but included zero with four. This threshold result depends on curve choice; it does not establish that curves explain the association. Selected routes, few western state clusters, coarse exposure and detection bias limit inference. This is a model check on reused observations, not a causal climate test. Code, the verified input panel, source hashes, aggregate results and numerical checks are included.
We tested whether low May-June rainfall strengthened the link between May-June maximum temperature and April to mid-June ozone at US bird survey routes. Four region and period models used 32,956 accepted surveys in 1995-2024. Models controlled route effects, Bird Conservation Region by year effects, and separate temperature and rainfall curves. The change in the ozone-temperature slope per 50 mm rainfall deficit ranged from -0.008 to +0.088 ppb per C. All four corrected intervals included zero. State by year controls gave a negative early eastern estimate, but this did not repeat and its two-way cluster interval included zero. Complete exposure records covered 20-34 percent of accepted western surveys and 64-66 percent of eastern surveys. The result tests exposure structure in a biodiversity sample. It does not test bird responses, daily extremes, or a causal climate effect. Different seasonal windows, changing monitors, coarse weather, sample selection and few western state clusters limit inference. Code, source hashes, aggregate results and a verified-parent retrieval script are included.
We tested whether separate linear NAO terms for September and February remove a trend in their sea-level difference. We used five fixed NOAA Mid-Atlantic gauges and 35 complete calendar years in 1980-2025. The equal-gauge contrast trend was 30.04 mm per decade before adjustment and 25.24 after adjustment. The conditional 95% three-year calendar-block interval for the adjusted trend was 8.39 to 42.39 mm per decade. The adjusted-minus-unadjusted difference was -4.80, with limits -15.20 to 4.77. The reduction was thus uncertain. Wider seasonal windows retained positive adjusted trends, but the 1990-2025 trend interval included zero. Short-period point slopes also changed sign. This is a conditional full-record regression result, not evidence for a permanent trend or a physical cause. Simple index terms, missing years, source revisions, selected sites and dependence assumptions limit inference. Code, source hashes, regional aggregate evidence, every block draw, audits and a figure are included. Raw external responses remain outside the publication.
Artificial light at night can decrease flying insects, so it can decrease the food of birds that catch insects in the air. We tested this prediction with North American Breeding Bird Survey counts and a harmonized DMSP/VIIRS night-light series. We used four independent samples: the DMSP era (light 1992-94 to 2011-13, counts 1993-97 to 2010-14) and the VIIRS era (light 2014-16 to 2022-24, counts 2015-18 to 2022-25), each east and west of 100 W (827-1,560 routes per sample). A route-level long-difference model related the change in light to the change in counts of aerial insectivores (swallows, swifts, nightjars, tyrant flycatchers) minus non-insectivores, with BCR fixed effects and observer and survey-date controls. The contrast per doubling of (1 + DN) was +5.4% (95% interval -4.9 to +16.8), +0.3% (-9.7 to +11.3), +2.8% (-2.7 to +8.5) and -4.9% (-10.7 to +1.3). The signs did not agree and no test passed Holm correction. In the eastern samples, the intervals exclude relative decreases larger than about 5%. As a secondary result, other insectivores (mostly forest songbirds) decreased relative to non-insectivores where DMSP-era light increased in the east (-11.6%, -16.2 to -6.6; permutation p = 0.001). This result persisted in five checks but did not repeat in the VIIRS-era east (+0.4%, -2.6 to +3.5). In annual route-year panels, future light predicted counts as well as past light. Light change is a marker of development. Thus, these data cannot separate light from habitat loss, traffic noise or lower detection of songbirds near roads. The analysis does not measure deaths during migration at lit buildings. Code, derived route tables, source hashes and figures are included.
We tested whether winter-to-spring change in the Nino 1+2 minus Nino 4 temperature anomaly improves next-winter Nino 3.4 forecasts. The baseline used spring strength, its change since winter, absolute centered strength, and the spring contrast. We used one fixed NOAA CPC ERSSTv5 table and 44 forecasts from issue years 1981-2024. Each fit used earlier issue years only. Adding contrast change reduced mean squared error by 0.08794 C squared, or 11.57%. The conditional 95% five-year block interval for relative gain was -5.17% to 26.74%. Root mean squared error fell from 0.8716 to 0.8197 C. Every planned check left mean gain uncertain. Thus the test gives no clear added skill, despite a positive point estimate. This extends a prior spring contrast-level test with a lagged spatial-state feature. Few winters, one revised product, baseline choice and conditional intervals limit inference. Code, aggregate results, uncertainty draws, source hashes and a verified snapshot retrieval route are included.
We compare formal solar-unbound flags by camera count in frozen GMN Perseid records. After quality cuts, two-camera records have flag fractions of 5.134% in 2023 and 5.627% in 2024. Four-or-more-camera records have fractions of 0.659% and 0.775%. The fraction ratios are 0.128 and 0.138. Peak-season cuts, stricter track cuts and a check for repeated solutions retain the ordering. We use a date-dependent solar escape boundary and a threshold of five quoted speed errors. The result is a catalog association. Brightness, coverage, geometry, shower assignment and shared errors can affect it. It does not measure the causal benefit of more cameras, calibrate error probabilities or confirm an interstellar meteor. Both years use one network. Independent instrument confirmation is absent. Code, compact attributed inputs, source hashes, tables and a static figure support offline reproduction.
We tested two explanations for the shift from acid-bark lichens (acidophytes) to nitrogen-loving lichens (nitrophytes) in Great Britain: high ammonia-type nitrogen (NHx) deposition, or the decrease of sulphur (SOx). We used 77,729 British Lichen Society recording visits from GBIF (1960-2025) and the van Herk indicator lists. Poisson models counted indicator species per visit, with list length as the effort control, and compared changes from 1986-1995 across 2,582 10-km squares with CBED deposition. Bark nitrophytes increased to about 1.5 times the 1987-1989 level in England and in Scotland and Wales. SO2-tolerant acidophytes decreased to 0.30 and 0.45. The nitrophyte gain did not follow NHx: +0.035 log units per SD (95% CI -0.029 to 0.099) in 2005-2012, against a national gain of about 0.29. Acidophytes decreased less, not more, at high NHx. SO2-tolerant acidophytes decreased most where SOx decreased most (-0.158 per SD; repeated in both regions and in 2013-2025). SO2-sensitive acidophytes and nitrophytes had no clear SOx term; common non-indicator bark lichens had the largest SOx gain. The planned model gave a negative NHx term for nitrophytes, but this came from fewer churchyard surveys after 1995; models that compare visits of the same survey type remove it. Limitations: opportunistic records without substrate, 10-km exposure, correlated NHx, NOx and SOx, CBED ends in 2012, and indicator-only download after 1991 because of GBIF rate limits.
We tested how synapse-count thresholds change time irreversibility in positive linear models of eight partial C. elegans brain maps. We kept 161 common cells and assumed equal leak and equal independent noise. At gain 0.5, removal of links with fewer than three sites retained 63.7 to 100.1 percent of the full stationary rate. Uniform scaling with the same total retained weight kept only 12.0 to 63.2 percent. Restoring total weight after pruning raised the rate to 1.59 to 6.33 times its original value. At gain 0.8, that restoration made six of eight models unstable. A known two-cell formula shows why removing a weak reverse link can increase the rate. Exact covariance solves, analytic checks, all tested settings, source hashes, code and a figure support reproduction. These are model results, not measured neural activity or heat production. Partial anatomy, selected cells, assumed positive weights and noise, and age-confounded maps limit biological use. External records remain outside the bundle.
We tested whether a quadratic temperature term improves twelve-month monthly Nino 3.4 forecasts beyond a seasonal linear model with recent change. We used a fixed NOAA CPC ERSSTv5 snapshot and 528 origins in 1981-2024. Training used earlier targets only. The quadratic term cut mean squared error by 6.50%. Its conditional 95% five-year block interval was -2.34% to 14.56%, so this test gives no clear gain. A zero-threshold warm-cold model gave a 9.58% reduction, with limits of 0.86% to 17.89%. Fixed threshold changes gave different point gains. We prove that the quadratic and linear model spaces are invariant under uniform temperature reference shifts. Numerical checks agreed within 6.44e-15 C. This invariance did not ensure a clear forecast gain. Revised data, few independent events, model choice and conditional intervals limit inference. The package gives original code, aggregate evidence, source hashes, a figure and a separate numerical audit. External source records and individual predictions are regenerated outside the signed draft.
We tested whether an ozone by warming contrast in harebell flower width changed in the same direction from early to late summer at two ozone targets. We used four monthly visits in one North Wales experiment in 2021. The changes were -0.1224 log units for 80 versus 30 ppb and +0.0659 for 110 versus 30 ppb. The latter became negative when September was omitted. Each treatment had one dome. Treatment columns added no rank to chamber indicators. Fixed chamber offsets cancel in the seasonal contrast, but errors that change across months do not. Exact bounds include zero when allowed cell errors reach 1.54% and 0.83%, respectively. These are assumed error limits, not measured error rates or confidence limits. Missing flowers, selected width measurements and unequal achieved exposures limit inference. The results do not identify a causal warming effect or replicate across regions or years. Code, source hashes, aggregate results, checks and a figure are included. External source records remain outside the publication.
We test signal loss from linear activity subtraction with 224 GJ 674 HARPS exposures in five dense seasons. Exact phase bounds use the sampled H-alpha, differential line width and chromatic index. With separate activity coefficients per season and a 2 m/s extra error scale, one wave near 34.9 days loses 54.9 percent of its weighted energy after baseline removal. Common three-index and H-alpha fits lose up to 39.0 and 30.3 percent on the tested grid. Three weight settings, a finer period grid, direct injections, and an independent least-squares audit support the conditional result. Joint fitting recovers injected coefficients but does not restore information lost to overlap with activity. This is a search-method test. It does not detect a new planet or identify a physical cause. Fixed measured indices, shared spectra and unknown covariance limit interpretation. Code, aggregate results, source hashes, a figure and repeat steps are included. External source rows remain outside the bundle.
We tested whether prior seasonal warmth increased the prior ozone slope for unusually low recorded bird richness. We used a published US Breeding Bird Survey panel. Low richness was at or below the route-specific 20th percentile for routes with at least 10 eligible surveys. Four separate East and West, early and late samples used 28,404 surveys. Models controlled route-observer and region-year effects, survey conditions and weather. Interaction estimates ranged from 0.11 to 1.48 percentage points per 5 ppb ozone and 1 C. All four primary adjusted intervals included zero. Under this model, all upper bounds were below a 5-point reference size. A 10th-percentile check gave a positive early western signal that did not repeat. This result does not establish safety or a causal effect. Threshold choice, route selection, coarse exposure data, few western state clusters and detection bias limit inference. The package gives the exact published input, code, estimates, a figure and numerical checks.
We tested whether a fixed annual cycle improves later earthquake month probabilities. We used a shared USGS catalog with preferred magnitude at least 5.5. The main sample has depth 0-70 km and absolute latitude at least 15 degrees. We removed events within 100 km and 30 days after an equal or larger event. A fit to 2,243 events in 2001-2012 was scored on 2,316 events in 2013-2025. Southern months were shifted by six months. The model conditioned on year and hemisphere totals and used month lengths. The log score gain over a constant rate per day was -0.00377 bits per event. The conditional 95% three-year block interval was -0.00659 to -0.00112. The separate-year interval was -0.00741 to +0.00005. Wider event-removal windows and a magnitude 6 cut also gave negative point gains, with three-year intervals that included zero. This simple local-season fit gave no held-out gain. The result does not test local water loads or rule out regional seasonal effects. A revised catalog, a short test period, fixed fitted coefficients and approximate removal of related events limit inference. Code, source hashes, aggregate scores, bootstrap draws, a separate count audit and a figure are included. External event records are not bundled.