dataesr/scientific-paragraphs-categorization huggingface dataset:
The corresponding tables containing the datapoints of the spectra are provided as supplemental material to this paper. These spectra can be used to compute the antineutrino energy spectra of future reactors loaded with various types of fuels in the frame of non-proliferation scenario studies. A big advantage of this set of spectra is the consistency brought by the use of the same model to compute the fission yields of all fissioning systems. This consistency could not be attained with the current evaluated fission yield datasets for the variety of fissioning systems that are needed for non-proliferation studies because of the lack of underlying experimental data and because of the remaining problems in the data listed in the paragraphs above. In the frame of the comparisons of diversion scenarios versus legitimate use of a given type of nuclear reactor  [29] , the fission yield covariance matrices provided by GEF make it possible to computate the uncertainties for the corresponding antineutrino emissions.
Systolic interarm differences in blood pressure have been associated with all-cause mortality and cardiovascular disease. We undertook individual participant data meta-analyses to (1) quantify independent associations of systolic interarm difference with mortality and cardiovascular events; (2) develop and validate prognostic models incorporating interarm difference, and (3) determine whether interarm difference remains associated with risk after adjustment for common cardiovascular risk scores. We searched for studies recording bilateral blood pressure and outcomes, established agreements with collaborating authors, and created a single international dataset: the Inter-arm Blood Pressure Difference -Individual Participant Data (INTERPRESS-IPD) Collaboration. Data were merged from 24 studies (53 827 participants).
METHODS Data-sharing agreements were signed with each collaborating study lead author, and an Independent Monitoring Group oversaw conduct. Under terms of the data-sharing agreements the INTERPRESS-IPD dataset is not freely available but can be discussed with the corresponding author on reasonable request. The study protocol is registered with the international prospective register of systematic reviews (PROSPERO) and has been published. 22 This study was conducted in accordance with the PRISMA-IPD (Preferred Reporting Items for Systematic Reviews and Meta-Analyses of IPD) statement. 23.
Novelty and Significance What Is New? • To determine the impact of differences in blood pressure between arms on cardiovascular outcomes (eg, heart attack, death), data from over 57 000 individuals from 24 studies around the world were collated into one large dataset and analyzed. • Results suggest that interarm differences above 5 mm of mercury (mm Hg) predict all-cause mortality, cardiovascular mortality and cardiovascular events. • Results are based on a single pair of measurements with blood pressure measured in one arm and then the other, and challenge advice that interarm differences in blood pressure should be confirmed using a simultaneous method (blood pressure measured in both arms at the same time). What Is Relevant? • Interarm differences in blood pressure identify additional risk of having a cardiovascular event beyond that predicted by existing cardiovascular risk scores alone; this could inform clinical decisions regarding primary prevention of cardiovascular disease.
Data Extraction and Preparation Eligible study authors were invited to share data; nonresponders received 2 reminders. Anonymized data were securely transferred to the University of Exeter server and cleaned with reference to published data and author discussion, where necessary. Datasets were combined using Stata v14. 0 (StataCorp, TX). For randomized controlled trials, we excluded interaction effects on outcomes between systolic IAD and treatment before including both trial arms in the IPD dataset. To overcome varying classifications of ethnicity, we mapped participants to the most commonly observed ethnic groups, seeking author and expert advice where required (Table  S2 ). Two authors (S. T. J. McDonagh, L. Cloutier, or C. E. Clark) independently assessed study quality using a modified Quality Improvement in Prognostic Studies tool (Table  S3 ).
Initial 2-stage IPD meta-analyses, modeling IAD for each study as both a continuous variable and serially as dichotomous variables with 1 mm Hg increments up to 20 mm Hg,  25  were used to estimate heterogeneity. This was categorized according to I 2 values as low (<25%) moderate (25%-50%) or high (>50%). 26 All models were adjusted for baseline systolic BP, age, and sex. One-stage random-effects flexible parametric models failed to converge (probably due to low between-study heterogeneity). Therefore, fixed-effect one-stage Cox proportional hazards modeling, stratified by study, was adopted, as per protocol. 22 For prognostic modeling, 4 cohorts (4264 participants) were reserved as a validation dataset. Validation cohorts were chosen nonrandomly to include participants of both sexes covering the full age range.
27 9] [30] [31]  One-stage Cox proportional hazards models, stratified by study, used the remaining cohorts (the derivation dataset) to investigate effects of participant-level covariates on outcomes with inclusion of IAD within the model. Baseline HDL cholesterol was missing for 7 studies, so was omitted as a candidate covariate in model development. Statistically significant covariates (threshold P<0. 1) were entered into multivariable models and retained if P values were <0. 05. Excluded covariates were individually returned to check whether inclusion improved model fit, tested with likelihood ratio tests, and the Akaike Information Criterion. 32 inal models were validated using calibration slopes and internal-external cross-validation analysis, with goodness of fit assessed using Harrell C statistic.
33 Kaplan-Meier curves for derivation and validation cohorts were compared using cutoffs approximating to 10-year mortality risks groups of <10%, 10% to 20%, and >20%. For participants without preexisting cardiovascular disease, systolic IAD was adjusted using Cox proportional hazards models stratified by study for 4 guideline-recommended cardiovascular risk scores: Atherosclerotic Cardiovascular Disease (ASCVD),  9, 34  Framingham,  8, 35  QRISK cardiovascular disease risk algorithm version 2 (QRISK2),  6, 36  and the European Systematic Coronary Risk Evaluation score (SCORE). 7, 37 ] [36] [37]  Inclusion bias was assessed by comparing outcomes in the INTERPRESS-IPD dataset with published cohorts not in the Collaboration. Publication bias was assessed visually using a funnel plot and quantified with Egger test.
38 ollowing multiple imputation of missing baseline data by chained equations, risk modeling was repeated and compared with models derived using observed data only; no missing outcome data were imputed. A total of 32 imputed datasets were produced, imputing baseline data across the pooled dataset, accounting for study in the imputation algorithm. A 3-person Public Advisory Group, facilitated by K. Boddy, was fully integrated throughout the study, contributing to bimonthly project meetings and facilitating dissemination of findings in accessible ways.
Study quality and assessment of bias Modified Quality Improvement in Prognostic Studies assessments judged 19 datasets to be at low risk of bias; 5 were unclear and none high (Table  S10 ); Quality Improvement in Prognostic Studies judgments were nondiscriminatory for outcomes. One published IAD study was not included in the INTERPRESS-IPD dataset; heterogeneity between HRs for it and included studies was low (I 2 =0%; Figure  S1 ). 21 We found no evidence of publication bias (Egger P=0. 657; Figure  S2 ).
Cardiovascular Mortality Cardiovascular deaths were reported for 1435 (3. 0%) of 48 261 participants from 21 studies (Figure  S3 ). Continuous systolic IAD, adjusted for age, sex, and baseline systolic BP was significantly associated with cardiovascular mortality (HR, 1. 07 [95% CI, 1. 03-1. 12]; I 2 =11%) per 5 mm Hg of IAD (Figure  S6 ). The derivation dataset for cardiovascular risk modeling consisted of 18 studies (33 650 participants with complete case data). The final model included systolic IAD (HR, 1. 06 [95% CI, 1. 02-1. 11] per 5 mm Hg) and, apart from total cholesterol, the same coefficients as the all-cause mortality model (Table  2 ). Harrell C statistics were good and overlapped for the derivation (0. 80 [95% CI, 0. 79-0. 82]) and validation (0. 83 [95% CI, 0. 79-0. 87]) cohorts.
Study Limitations Comprehensive searches were undertaken for cohorts that reported either IAD or ankle-brachial index. Additional data were identified through author contacts. 77, 81, 82, 88, 89 In total, we combined data from 24 of 34 confirmed eligible studies, representing two-thirds of potentially eligible participants, and comparison with the one published eligible IAD study that we did not obtain showed no evidence of inclusion bias. These IPD meta-analyses have required considerable data cleaning and preparation; closure of data inclusion in 2017 means that we could have excluded relevant recent publications, however, continued search updates to April 2020 have only identified one new potentially eligible study. 98 Our search strategy could have missed cohorts or registries holding bilateral BP data if no linked IAD or ankle-brachial index publications exist.
Perspectives Associations of sequentially measured systolic interarm differences in BP with all-cause mortality, cardiovascular Downloaded from http://ahajournals. org by on January 7, 2021 mortality, and events were investigated in a multinational individual participant data meta-analyses including 53 827 participants. Adjusted associations of IAD with all-cause mortality and cardiovascular mortality were confirmed, and models incorporating IAD were developed and validated. Interarm differences were also associated with higher risks of cardiovascular events after adjustment for commonly used cardiovascular risk scores. BP should be measured in both arms during cardiovascular assessment. Further research is needed to determine whether selection of intensive cardiovascular risk reduction strategies, on the basis of a detected interarm difference, can modify outcomes for people with an interarm difference.
Acknowledgments We thank the following for their involvement in the Inter-arm Blood Pressure Difference -Individual Participant Data ( INTERPRESS-IPD) Collaboration: Independent Monitoring Group: Alun Hughes  (chair),  Gary Collins , and  Tom Fahey. Data management Group:  Tim Eames  and  Sofia Sanabria (Exeter Clinical Trials Unit). Public Advisory Group :  J. Goddard ,  M. Turner , and  N. Reed. Ellie Kingsland  provided administrative support throughout the study. We thank  Professor Gerry Fowkes  (ankle-brachial index collaboration) and  Professor Colin Baigent  ( Antithrombotic Trialists' and Cholesterol Treatment Trialists ') for their assistance in establishing that existing triallists' collaborations could not address these research questions. Morwenna Rogers  assisted in development and piloting of the search strategy,  Francesco Cappuccio  (European Society of Hypertension Working Group on Hypertension and  Cardiovascular Risk  assessment in subjects living in or emigrating from Low Resource Settings) advised on ethnic classification across cohorts.
Sources of Funding This study was funded by the  National Institute for Health Research (NIHR) Research for Patient Benefit Programme  ( PB-PG-0215-36009 ). C. E. Clark was supported by an  NIHR   Clinical Lectureship award. R. J. McManus  receives support from the  NIHR Oxford Collaboration for Leadership in Applied Health Research and Care. The views expressed are those of the authors and not necessarily those of the  NIHR , the  National Health Service , or the  Department of Health.
14 software, removal of reads with length <100 bases, as well as primer clipping (up to three mismatches per primer), were performed by the sequencing company (LGC Genomics GmbH). Further processing of the sequence reads was done in R using the DADA2 pipeline  (Callahan et al. , 2016)  applying settings given in the instructions 1  , with some modifications (Rcode is available in https://github. com/angelrainf/gesifus. cryo. dada2). A total of 700 amplicon sequence variants (ASVs) were identified across all sequenced samples and the obtained ASVs were taxonomically assigned using the Genome Taxonomy Database (GTDB)  (Parks et al. , 2018). Raw sequence data are available at European Nucleotide Archive under the accession number PRJEB38947.
Differences in variance among compared datasets violate assumptions for two-sample t-tests. We accordingly used a one-sample t-test (two-tailed) to statistically compare the Shannon diversity between initial and resuscitated communities. For this purpose, we used the mean Shannon diversity of the technical replicates from the initial community as a reference value and we subtracted Shannon diversities from the triplicate incubations from this mean. The one-sample t-test was then applied to test the null hypothesis that the difference in diversity between the initial sample used as inoculum and the incubations did not differ from zero. To assess and subsequently visualize the similarity of resuscitated communities to the initial starting communities in dependence of the growth medium and inoculum (experiment 2), pairwise Bray-Curtis distance indices between all communities were calculated.
ACKNOWLEDGMENTS We thank the  Service d'Observation en Milieu Littoral (INSU-CNRS, Banyuls-sur-Mer)  for providing data, as well as the facilities to collect water samples. The facilities for sampling the Warnow River were provided by the  Leibniz Institute for Baltic Sea Research  and we thank  Uwe Hehl  for his support during sampling. We thank  Olivier Crispy ,  Jenny Jeschek , and  Madleen Dierken  for their help in the analysis of dissolved organic compounds and nutrients. We are grateful to the Bio2Mar platform (http://bio2mar. obs-banyuls. fr)for providing technical help and support. We also thank  BIOPIC platform at the Oceanological Observatory of Banyuls-sur-Mer  for access to flow cytometry facilities and  Philippe Catala  for performing the counting of heterotroph flagellates. We thank  Inessa Lacativa Bagatini  for her help in the data analyses and her input on earlier versions of the manuscript.
FUNDING The study was supported by a grant from the  German Science Foundation (DFG)  awarded to SB ( BE 5937/2-1 ). AR-F was funded by the  National Agency for Research and Development (ANID) / Scholarship Program/DOCTORADO   BECAS CHILE/2017-72180448. GPM was funded by the  doctorate scholarship program  of the  Coordination for the Improvement of Higher Education Personnel (CAPES) ,  Brazil , and a financial support via the above-mentioned DFG funding.
https://benjjneb. github. io/dada2/tutorial. html Sequencer 3130XL (Applied Biosystems R , Foster City, CA, United States). Terminal restriction fragments (T-RFs) were identified using the GeneMapper software (Applied Biosystems R , Foster City, CA, United States) and the areas of the identified T-RF peaks were imported into the T-Rex (T-RFLP analysis EXpedited) online software for noise filtering, alignment, and binning by default parameters(Culman et al. ,.
If this is the case, funding authorities may have a role in either negotiating better APC rates with publishers or considering increasing specific OA funding. Some European initiatives are being launched on this purpose, such as the Horizon 2020 programme, which demands to ensure OA to all scientific publications related to this program (https://ec. europa. eu/ programmes/horizon2020/en). Scientific knowledge is advancing very fast and the ways of doing science are changing with it. Alternative options to share scientific results are growing, such as public repository sites like Arxiv (https://arxiv. org/) or Biorxiv (https://www. biorxiv. org) as well as private companies hosting websites where to share publications and other datasets such as Re-search Gate or Academia. edu. Nevertheless, there is still much work left to promote and extend their use, as a large percentage of researchers do not to use any (Fig.
Milanovic (2018) adds 13 social tables to this database, including one from Sarah Merette (2013a, b) on colonial Southern (Cochinchina) and Northern (Tonkin) Vietnam in 1929. The social table approach entails a number of limitations. First, the method is mostly based on labor earnings linked to occupations, and often ignores (or fails to capture) other relevant sources of income (rents, profits). 6 Second, inequality within occupational groups can be severe, and can change across time. Third, the contents of each occupation may vary across space and time. Fiscal sources allow for a more direct approach that can complement social tables and address some of the drawbacks of missing survey data. After World War I, Britain and France gradually introduced progressive personal income taxes in their colonial empires, shortly after the enactment of these taxes in the metropolis.
In Algeria, until 1945, lump-sum deductions for family charges applied, and reported income was net of these deductions. Before 1937, only the number of children under 18 years old mattered. Starting in 1938, deduction schedules changed somewhat: married couples benefited from an initial deduction even in the absence of children, and the total deduction was modulated according to the level of declared income. These deductions mechanically generate an underestimation of the top income shares. As household composition is most likely correlated with income, it is very difficult to implement a refined correction. We can still compute a coarse estimate for the resulting underestimation. We used population census data for Algeria in 1948 to estimate the distribution of family types (married couples or single parents according to the number of children).
This method was chosen according to the data that was available, but also for the purpose of comparing our results with those of Atkinson (2014; 2015a, b, c), who made the same choice when analyzing British colonies. In the case of Algeria, and for Europeans in Indochina, we can confirm that this option is consistent with a more refined analysis of demographic structures, as we detail below. 22 lgeria. The total numbers of "Muslims" and "non-Muslims" are drawn from population censuses of 1931, 1936, 1948, and 1954 (Annuaire Statistique de l'Algérie, 1955). A population estimate for 1960 is taken from the United Nations Demographic Yearbook for 1966. The number of non-Muslims is drawn from the Annuaire Statistique de la France for 1962. These figures were double-checked with demographic studies from CICRED (1974a), and from Fargues  (1986).
The 1950s are also specific in that remittances (not included in GDP but included in household income) from Algerian migrants in France were already quite significant. For these reasons, taking 1950 as the base year is safer than taking an average of the 1950s. French Cameroon. The GDP of the French part of Cameroon was estimated for the years 1947 and 1953 by the French administration in charge of the overseas territories. From 1947 back to 1945, we use Cogneau, Dupraz and Mesplé-Somps' (2018) extrapolation, based on trade data. We translate the constant franc estimate to current prices using the consumer price index for Brazzaville (the capital city of French Congo, the nearest place where price data were collected between 1938 and 1951). We take 50% of GDP to obtain household income. Indochina / Vietnam.
The income shares of the top 1% and 0. 1% groups across time Table  2  presents our estimates of top income shares for Algeria (Table  2A ), Tunisia (Table  2B ), French Cameroon (Table  2C ), and Vietnam (Table  2D ). Figure  1  shows the income share of the top 1% in Algeria and Tunisia, compared to France, and in South Africa and India compared to the United Kingdom, between 1920 and 1960. In Algeria, this share starts from a very high 22% in 1932, then drops steadily throughout the 1930s, and drops further during WWII, like in France. It stabilizes at around 15% in the 1950s. Measured between 1946 and 1956, the figures for Tunisia are slightly lower but close, at around 14%. The inequality levels in Algeria in the early 1930s appear to be among the highest recorded in WID. world. 30 Even if the available data do not allow for a precise estimate, this suggests a top 10% income share of at least 60%, if not higher.
9% of the estimated non-European income. In any case, this would put the extrapolated top 1% far below 6%. We also get a 1. 2% share for the top 0. 1%, again much lower than the share for France (2%). In Vietnam, all European tax units are in principle recorded, as well as their taxable income. We thus just use the reported figures alone to individually calculate the top shares for Europeans and for non-Europeans. For the latter, this is only possible when they are supposed to be taxed, i. e. 38 Data for Europeans in 1948 is incomplete and cannot be used. 39 Obtained again as 60% of the 15+ population, assuming a 75% share for this age group, applied to the figure of 3,600 Europeans reported in the Statistical Abstract for 1936-1945. only for Cochinchina after 1937. 40 Before WWII, we find rather high top shares, especially in 1940 when the number of non-European taxpayers reached a local maximum (before an additional inflow of Chinese migrants during the 1940s, see above).
* We acknowledge the financial support that the  French National Agency for Research  ( ANR-11-BSHS1-006 ), the  Institute for New Economic Thinking , the  ESRC-DFID Joint Fund  (Grant  ES/I033114/1 ), and the  European Research Council  (Grants  340831  and  856455 ) provided at different stages of this research. We are grateful to  Louise Guillouët  and  Augustin Bergeron  for excellent research assistance at different stages of the project. Jean-Pascal Bassino  kindly gave us access to his data on Vietnam. We thank  Sylvie Lambert  for helpful comments, as well as conference participants at  University of Sussex  ( AEHN 2016 ), the  Paris School of Economics , and  MIT  (WEHC.
Field data collection Cities were visited twice from mid-May to mid-July 2018. Bat recordings and nocturnal insect collections were conducted simultaneously at the 12 selected UGAs of each city, except the 32 sites in Zurich, which were split in two sampling periods due to limited equipment. Samplings were scheduled for 5 consecutive nights per site for each of the two sampling periods, in order to avoid insufficient data sizes and reliably sample the species community. Samplings were carried out under favourable weather conditions, that is, without rain and temperatures above 12 °C  (Hutson, Mickleburgh & Racey, 2001).
As organism type or size may influence its quality as bat food  (Hutson et al. , 2001) , each order was assigned a relative factor according to its body length and a wet biomass index calculated per sample. We assumed insect shapes to be ellipsoids for the calculations. Some orders were additionally divided into size classes (e. g. Lepidoptera: <5 mm, 5À20 mm, >20 mm; see Appendix A: Table  2 ). Note that insect variables could not be estimated for all the sites, as several of the initial light-traps were vandalised in the course of the sampling. From the initial 56 sampling patches, we obtained unbalanced data for 12 sites in Paris, 11 in Antwerp and 26 in Zurich.