dataesr/scientific-paragraphs-categorization huggingface dataset:
Acknowledgments We would like to thank  Dr Chengming Huang ,  Beijing Museum of Zoology, Chinese Academy of Science, Beijing ;  Professor Hu Yaowu ,  Institute of vertebrate Paleontology and Paleoanthropology of the Chinese Academy of Science, Beijing  and  Dr Josephine Lesur ,  Muséum  national d'Histoire naturelle, Paris-all of whom kindly made available modern osteological collections. Isabelle Carrère, EHESS, contributed to the production of Figs 1 and2.
The ERAnet (https://ec. europa. eu) Co-Reach  project  n°137  ( European-Chinese Bioarchaeological Collaboration, Euch-Bioarch ) led by  K. Dobney , and the  Chinese Academy of Social Science  funded this research. This research was also granted by the  Agence National pour la Recherche (ANR , http://www. agence-nationale-recherche. fr/), through the LabEx  ANR-10-LABX-0003-BCDiv , in the framework of the  Investissements d'avenir programme  ( ANR-11-IDEX-0004-02 ).
(9, 29, 30)  and Conville et al. (31). To perform the phylogenetic analysis, we obtained the sequences of the type strains of Nocardia species genetically closest to strain OFN 14. 177 T from the GenBank database with the exception of N. vulneris, N. altamirensis, N. iowensis, N. pseudovaccinii, and N. anaemiae, which we determined. The multiple-sequence alignment software Clustal X (32) and the MEGA6 software (33) allowed us to build phylogenetic trees by three different methods: maximum likelihood (34), maximum parsimony  (35) , and neighbor joining  (36). Branch robustness was calculated by a random resampling according to the bootstrap method (1,000 replications). DNA-DNA hybridization (DDH). The degree of DNA-DNA relatedness between strain OFN 14. 177 T and all the species belonging to the N.
6 kg concentrate/day in the parlor. The feed offered during milking is not included in the analysis reported here, and only one cow (which was not lame) did consistently yield >30 l/day during the experiment. Cows had ad libitum access to water from two water troughs, one at each end of the pen. Water intake was not recorded. The experiment started on June 23, 2014, hereafter referred to as day 1, and lasted for 24 days. The cows were permanently in the pen from day 6 and throughout the experimental period, except for short periods during normal farm routines such as milking and pregnancy detection. All 20 cows were equipped with neck-mounted accelerometers. However, two cows past 22 months in lactation were excluded from analysis, and accelerometer data could not be retrieved from two cows, leaving 16 cows in the final data set used for analysis.
accelerometers and rumination classification The cows were equipped with a neck-collar fitted with an accelerometer (Silent Herdsman Ltd. , Glasgow, UK) sampling in three dimensions at 12 Hz. The accelerometers were a prototype; therefore, data were stored on the SD-card of the accelerometer and transferred manually to a local computer when the accelerometers were taken off. To ensure data storage, two batches of accelerometers were used in serial. The first batch collected data from day 1 to 9 and the second batch collected data from day 10 to 25. Because days 9 and 10 were not full days of data collection, they were removed from the data set. Unfortunately, data from four accelerometers were corrupted either in the beginning or in the end of the experimental period. Therefore, the accelerometer data consisted of time-series from 16 cows with on average 19.
statistical analysis All analysis was performed using R version 3. 2. 0 (23). Coefficients of variation (CV) were calculated to express inter-and intraindividual variation. We applied the lme4-package  (24)  for linear mixed effects analysis. For the variables summarized on a daily basis (see Table  1 ), we first wanted to see if they were affected by the lameness status of the cows. Days in milk were included in this model because feeding behavior, in particular feed intake is known to change with days in milk. The resulting model is: Y ijk i j i j k ijk = + + + × + + µ LAME DIM LAME DIM cow ε (M1) Y represents the variables summarized on a daily basis, μ was the overall mean, lameness (LAME; i = lame; not lame), days in milk (DIM, j = 54, …, 384 days), and their interaction term were fixed effects, cow (k = 1, …, 16) was random effect, and ε was the error term.
acKnOWleDgMenTs The authors would like to thank Ivan Andonovich at Silent Herdsman Ltd. , Glasgow, UK and University of Strathclyde, Department of Electronic and Electrical Engineering, Glasgow, UK for providing the accelerometers. We would also like to thank Ainsley Bagnall, SRUC, Dumfries, UK for taking care of the cows and providing data from the barn and David Roberts, SRUC, Dumfries, UK for providing the cows and research facilities.
FUnDing VT has received the support of the European Union, in the framework of the Marie-Curie FP7 COFUND People Program, through the award of an AgreenSkills' fellowship (under grant agreement no. 267196). This work was part of the Dairy ICT project funded as an EU ERA-net (French research funding agency ID: ANR-12-ICTA-0001) within the ICT-AGRI program.
[30]. Scratches (i. e. superficial lesions) and deep wounds (i. e. haemorrhages) were differentiated. Behavioural observations in the home pen after weaning. On the day of weaning, and 15 and 27 days after weaning, behaviours of weaned piglets were scored live using 2-min instantaneous scan sampling for 6 h/day during 6 sessions of 1 h (8:00-9:00 h, 9:15-10:15 h, 10:30-11:30 h, 14:00-15:00 h, 15:15-16:15 h, 16:30-17:30 h). The behaviours (S1 Table ) were recorded using a Psion hand-held computer with the Observer 5. 0 software package (Noldus Information Technology B. V. , Wageningen, The Netherlands). Feed preference tests. The day after the end of the dietary treatment (i. e. 8 weeks of age) and 8 weeks later, the piglets were subjected to a two-choice feed preference test. The control starter diet and the HFS diet were distributed in excess to estimated intake in two adjacent feeders at 9:00 h.
The bucket was then left on the floor for 5 min. After each test, faeces were removed from the test arena, which was cleaned with water and cleanser, and dried with towels. The order of testing was balanced for treatment and room. During the tests, behaviours and postures (S2 Table ) were scored live using Psion hand-held computers with the Observer 5. 0 software package (Noldus Information Technology, Wageningen, The Netherlands). The tests were also recorded on video to perform a posteriori analyses of locomotion patterns in the arena using the Ethovision XT 10 video tracking software (Noldus Information Technology B. V. , Wageningen, The Netherlands). The variables calculated in Ethovision were: the time spent in the centre, walls, observer, entrance and novel object zones (Fig 1 ), the latencies (s) to enter the centre and novel object zones, and the total distance (m) covered during the tests.
Statistical analyses Statistical analyses were conducted using SAS version 9. 1. 3 (Statistical Analysis Software; SAS Institute, Cary, NC, USA). If the model residuals were not normally distributed, the statistical analyses were performed on (arcsin) square root or logarithmically transformed data. Two pens were excluded from the study because of a methodological error in replicate 2, resulting in n = 8 pens for CON/HFS and HFS/CON treatments, and n = 7 pens for HFS/HFS and CON/CON treatments. Data are presented as (untransformed) means ± SEMs. The upper limit for the statistically significant effect was set at p 0. 05, with 0. 05 < p 0. 10 considered as a trend. Effects of dietary treatment. Litter size at farrowing was analysed using a mixed model with prenatal diet and replicate as fixed effects.
Indeed, behaviours on this day differed profoundly from behaviours 15 and 27 days later, and were therefore analysed separately. Behaviours on the day after weaning were analysed using a mixed model that included the fixed effects of prenatal diet, postnatal diet, their interactions and replicate. Behaviours at 15 and 27 days after weaning were analysed using a repeated mixed model with prenatal diet, postnatal diet, day, their interactions and replicate as fixed effects. As preliminary analyses showed that the number of skin lesions on the day of arrival to the home pen significantly differed from the number of lesions 7, 14 or 21 days after weaning, these data were analysed separately. The numbers of skin lesions the day of weaning were analysed using a mixed model that included the fixed effects of prenatal diet, postnatal diet, their interactions and replicate.
Acknowledgments The authors are very grateful to  Fleur Bartels ,  Monique Ooms ,  Jan Wijnen ,  Anouschka Middelkoop  and  Welmoed Wehkamp  for all the help in conducting this experiment. We would like to thank  Rudie Koopmanschap  for his assistance with the cortisol analyses,  Sietse-Jan Koopmans  for his valuable advice on the study and the personnel from the  experimental farms of Carus (Wageningen)  and  VIC (Sterksel)  for taking care of the animals and their technical assistance. We are also very thankful to  Dr. Claudia Kalbe  for her invaluable help in interpreting data on piglet performance. This research has been supported by a post-doctoral study grant from the  Fyssen Foundation  awarded to  Dr. C. Clouard. Scientific interactions with the  INRA DID'IT Metaprogramme made possible  to share insights on the methodology and results interpretation.
Meta-analysis A literature search and screening process were conducted using the PubMed, CAB and Google Scholar search engines to create a dataset of papers with the key words "passive immunity", "IgG", "immunoglobulins", "colostrum management", "colostrum", and "calf", separately or in combination. Additional papers referenced by at least 1 of the papers identified in the search were also included. To be included in the dataset, the papers must have examined the risks of various disorders (mortality, all diseases and production changes in calves with or without FPT) and have been peer-reviewed. No other inclusion criteria were used. Exclusion criteria were (i) papers with no quantification of the risk of diseases in case of FPT, since they cannot included in the meta-regression, (ii) definition of PFT that did not fit with the retained one, since they cannot be included in the related co-variables (Table  1 ) and (iii) outcomes that are not mortality, diarrhoea, BRD or average daily gain (ADG), since other outcome only gather one or two data.
Papers published through June 2014 were included (Fig  1). Fifteen papers evaluating the association between FPT and the above-mentioned outcomes in calves were identified. The main reasons of exclusion were no quantification of risk and no clear definition or quantification of FPT (Fig  1 ). Most of the papers studied several outcomes, and 68 different models published in the literature were included in the present study. A template for data extraction was drafted that included the numbers of calves and herds studied; the breed (beef or dairy); the statistical method used (logistic regression, Poisson regression, or raw data with contingency table); the expression of risk (relative risk  [RR]  or odd ratio [OR]); the prevalence of FPT; the metabolite used to diagnose FPT (total proteins (TP) or Ig); the threshold of TP or Ig used to diagnose FTP; the nature (univariate [U] or multivariate [M]) of the reported model; the prevalence of the outcome or of the mean value if relevant; the duration of the period of outcome inclusion; the value of the risk and its 95% confidence interval (95% CI); standard error (SE) or standard deviation (SD); and all the covariates included within the given model.
No distinction was made between total Ig and IgG, and only the designation IgG was used. For diseases, the method of diagnosis (in particular, whether by a veterinarian or a farmer) was also reported in the dataset. Calves were sampled after 24 hour of life, so as to wait the final immunoglobulin absorption. Sampling occurred in a narrow window of time, such as from 24 to 36 hour of age, in particular when TP where measured. When sampling occurred in a large window of time, up to 1 week old, Ig and not TP were sampled in most of case. This was consequently not further considered. All the studied publications had obtained their samples prior to the onset of disease. The threshold used to diagnose FPT, the presence of a gap between the thresholds used to define calves with or without FPT, the duration of observation of the outcomes, and the diseases analysed in cases of multiple diseases were transformed into categorical moderators for the meta-analysis (FPT DIAG1 ; FPT DIAG2 ; FPT GAP ; FPT THRES ; OUT TIME and OUT MORBIDITY ) according to the rules noted in Table  1.
Moderator Classes Definition 1 extra moderator was created (designated "Group") for each outcome by grouping the different models in the same paper into different random classes. A meta-analysis was conducted on the extracted outcomes using the Metafor package of R (version 3. 0. 2; R Foundation for Statistical Computing, Vienna, Austria). The meta-analysis was performed as previously described  [12] , and the different steps were recently outlined  [13]. Briefly, fixed-effects and random-effects models were first conducted for each outcome to estimate the log-effect size, its 95% CI and its statistical significance. The inconsistency of results among trials was quantified using both Cochran's Q test and the I 2 statistic  [14]. An I 2 value greater than 50% was considered to indicate substantial heterogeneity.
C MORBIDITY ¼ ( C MORBIDITY Ã Sev FPT for calves with FPT C MORBIDITY for calves without FPT ð14Þ C MORBIDITY ¼ ( C MORBIDITY Ã Sev PRESSURE when Prev HERD FPT ! 40% C MORBIDITY when Prev HERD FPT < 40% ð15Þ Eqs 12, 13, 14 and 15 apply additively in Eq 5. The economic model was run using Scilab open source software (www. scilab. org) with 10,000 iterations, and 95% prediction intervals (PI) and 95% CIs were calculated. The PI makes it possible to predict the situation for the next farm with 95% probability, whereas the CI indicates the situation in 95 of 100 farms visited. The RR i , the prevalence and the unit costs C i used in the assessment are presented in Table  2  and explained in detail. Most of the input parameters were included as a law of distribution (normal or log-normal) and not as a point estimate.
Four scenarios were proposed that combined different input parameters. The baseline scenario was the most probable. The alternative scenario included more contributors and different calibrations. The low and high scenarios used minimum and maximum values for calibration, respectively. For RR i , the alternative and high scenarios included raw data from the literature review before any correction, whereas the baseline and low scenarios included the results of the above meta-analysis. Prev COMPi_NoFPT was calculated using Eq 6 and applied to studies of the meta-analysis (S1-S3 Tables). Lack of data on omphalitis and septicaemia led to a definition in line with the entity "morbidity" of the meta-analysis. Prev Mortality_NoFPT was defined based on Eq 6, and data from studies included in the meta-analysis led to a value considered low by the authors (3.
Meta-regression The association between FPT and mortality was reported in 28 models from 10 publications (S2 Table ). The mean (SD) risk of mortality associated with FPT was 4. 87  (7. 62 ). The heterogeneity of the dataset was high (I 2 = 68% [95% CI = 46-79] and Q statistics χ 2 = 80, df = 27, P < 0. 001). The intercept of the log-effect size in the mixed-effects model with no moderator was 0. 88 (SE = 0. 020, P<0. 001), which corresponded to an effect size of 2. 41 (1. 62-3. 58). Including the moderators FTP DIAG1 , FTP DIAG2 and FTP GAP reduced the heterogeneity by 23, 29 and 22%, respectively (Table  3 ). All other moderators were not significant (P>0. 05), and their inclusion in the meta-regression did not decrease the heterogeneity. Similarly, no multivariate meta-regression allowed us to reduce the heterogeneity.
Replacing FTP DIAG1 with FTP DIAG2 in the meta-regression led to an OR of mortality in the case of FPT of 1. 92 (data not shown) instead of 2. 12 (Table  3 ). The sensitivity analysis showed no outlier for the metaregression without a moderator, but doubt exists regarding model #4 (S2 Table) when the moderator FTP DIAG2 was considered based on DIFFTS and cov and τ 2 statistics. Excluding these  The association between FPT and BRD was reported in 12 models from 5 publications (S3 Table ). The mean (SD) risk of BRD associated with FPT was 2. 24 (1. 20). The heterogeneity of the dataset was high (I 2 = 72% [55-82] and Q statistics χ 2 = 61, df = 11, P < 0. 001). The intercept of the log-effect size in the mixed-effects model with no moderator was 0. 82 (SE = 0. 17, P<0. 001), which corresponded to an effect size of 2.
27 (1. 62-3. 17 The association between FPT and diarrhoea was reported in 5 models from 2 publications (S1 Table ). The mean (SD) risk of BRD associated with FPT was 1. 71 (0. 70). The heterogeneity of the dataset was high (I 2 = 80%). The intercept of the log-effect size in the mixed-effects model with no moderator was 0. 56 (SE = 0. 32, P = 0. 07), which corresponded to an effect size of 1. 76 (0. 94-3. 27). None of the moderators were significant. Excluding the data from model #2 as suggested by the sensitivity analysis led to a risk (95% CI) of diarrhoea in calves with FPT of 1. 51 (1. 05-2. 17). Four publications (8 models) provided data on the risk of at least one disorder (except mortality) or at least one treatment in cases of FPT (S3 Table ). The mean (SD) risk associated with FPT was 2.
72 (1. 71). I 2 was 51%. The intercept of the log-effect size in the mixed-effects model with no moderator was 0. 58 (SE = 0. 17, P<0. 05), which corresponded to an effect size of 1. 80 (1. 28-2. 52). None of the moderators were significant. The sensitivity analysis did not reveal any outliers. When combining data related to the last 3 outcomes, 25 studies from 10 publications were available (Table  3 ). The mean (SD) risk of BRD associated with FPT was 2. 34 (1. 31). The heterogeneity of the dataset was high (I 2 = 73% [50-88] and Q statistics χ 2 = 107, df = 24, P < 0. 001). The intercept of the log-effect size in the mixed-effects model with no moderator was 0. 64 (SE = 0. 08, P<0. 001), which corresponded to an effect size of 1. 91 (1. 63-2. 24). The moderators FTP DIAG1 and OUT MORBIDITY were significant, but their inclusion did not reduce the heterogeneity and led to slight changes in the log-effect size (Table  2 ).
This is in accordance with the results of the meta-regression for mortality and morbidity showing the significance of the moderator FTP DIAG and the decrease in heterogeneity when this moderator was applied (Table  3 ). The adjustment of the mean risk by the threshold used to diagnose FPT allowed us to prevent an overestimated mean risk (2. 12 instead of 2. 41) in the present work (Table  2 ). Similarly, the risk of mortality for FPT comparing two extreme populations for IgG or PT is overestimated (FPT GAP , Table  2 ). The duration of observation of mortality was not reported as a significant bias of the risk value in the present work, even though the longer the study, the higher the expected mortality. The absence of any significant moderator for BRD, diarrhoea, or at least one other disorder is linked to the low statistical power permitted by the size of the dataset.
These changes may be high. For example, the OR for mortality decreased from 4. 87 (raw mean) to 2. 41 (mixed-effects meta-regression without moderator, Table  3 ). The final meta-regression (and relative adjusted risk) that was retained was judged on the reduction of the heterogeneity relative to the regression without moderators. Only the mixed-effects models were reported in the present work, in accordance with the structure of the dataset (several lines of the dataset were from the same publication). The funnel plots did not show any publication bias. They were not asymmetric, suggesting no studies with a reduced likelihood of inclusion within the meta-analysis because they were not published due to small or non-significant results. A double-step stochastic method for easy application of the results in the field In the cattle industry, many decisions about animal health are made without considering economics.
Horses Between 2012 and 2014, 96 Arabian and Arabian cross endurance horses participated in the study. The horses were aged between 4 and 6 and came from across France. Nineteen of the tested horses were excluded definitively from the final analysis, due to either poor-quality cardiac recordings (n = 18) or failure to complete the test (lameness, n = 1). For the remaining 77 horses, we excluded six poor-quality cardiac recordings at rest, 23 poorquality recordings made during the first 15 min of exercise, and 16 poor-quality recordings made during the recovery period. The characteristics of the remaining 77 horses are presented in Table  1  by age and by test site. Information on each horse's training and racing profile (such as the number of races, the total distance covered during races and the weekly duration of training) was obtained from the French Equestrian Federation's database  (FFE compet)  or a questionnaire completed by the owner.
Statistics All data were found to be normally distributed. The effects of age, exercise duration and test conditions were analyzed separately. In order to determine the effect of age on HRV variables at rest and during exercise (T: 4-vs. 5-vs. 6-year-olds; 2T: 5-vs. 6-year-olds) a one-way analysis of variance (ANOVA) was performed. The effect of test conditions on HRV variables at rest and during exercise and recovery was assessed in a one-way ANOVA. The effect of exercise duration on HRV variables was evaluated separately in 5-(T vs. 2T) and 6-year-olds (T vs. 2T vs. 3T) using a one-way repeated-measures ANOVA. Spearman's rank order correlation was also calculated. All analyses were performed with Sigma Stat software (version 3. 5, 2007, Systat Software Inc. San Jose, CA, USA). The threshold for statistical significance was set to p < 0.
DISCUSSION By analyzing a large dataset of HR recordings (in 77 young endurance horses performing exercise tests at four different sites), we generated new information on the effects of age, test conditions, and exercise duration on HRV in endurance horses. Our main finding was that HRV amplitude decreased as the age and exercise duration increased. Furthermore, HRV analysis during recovery might be of value for selecting the fittest young endurance horses (i. e. , those with the fastest cardiac recovery within 20 min of arrival at a vet gate). Blocks of 15-min RR recordings at rest and during exercise and recovery were analyzed with methods for quantifying HRV in a nonlinear analysis (the Poincaré plot) and in the timefrequency domain (computation of the SPWVD). The major advantages of the Poincaré plot are its ability to provide an overall preview of HRV and its relative insensitivity to the RR interval recording quality  (Kamen et al.
Methods A review of the available literature was accomplished by searching electronic databases including MEDLINE and PubMed, using the following key words: 'non tuberculous mycobacteria', 'nontuberculous mycobacteria', ' Mycobacterium avium complex', 'M. abscessus' , 'M. kansasii' , 'M. xenopi' , and 'M. malmoense'. In addition, reference lists from review articles and guidelines were handsearched for relevant articles.