dataesr/scientific-paragraphs-categorization huggingface dataset:
This research was initiated as part of the ADAM interdisciplinary project on Distributed Architectures and  Multiple Multimedia Applications  supported by a grant from the  French ANR (Agency for National Research ). It has been further supported by the  Framework Programme   FP7-ICT-2013-10  of the  European Commission  through the project  P2Pvalue  (grant no. :  610961 ) and the  EINS network of the European Commission on Internet Science. Earlier versions of the typology were presented at the following conferences: the 2012  SCRIPT Conference, Law and Transformation, Arts and Humanities Research Council Centre for Studies in Intellectual Property and Technology Law at the University of Edinburgh ; the 2013  EINS First Conference on Internet Science in Brussels ; the 2013 Fourth Annual Transforming Audiences Conference, University of Westminster, London; and the Cost Action 2015 workshop on Digital Labour in Brussels.
Methods TB diagnostic yield (the proportion of presumptive TB cases with a laboratory diagnosis of TB) was assessed using a non-randomised stepped-wedge design as sites transitioned to the Xpert 1 based algorithm. We identified the full sequence of sputum tests recorded in the electronic laboratory database for presumptive TB cases from 60 primary health sites during seven one-month time-points, six months apart. Differences in TB yield and temporal trends were estimated using a binomial regression model.
Samples were initially tested at a central national health laboratory. Three sub-district laboratories were established in 2013 to perform Xpert tests and smear microscopy. Results for all samples were entered into a networked electronic laboratory database. Definitions. A presumptive TB case was defined as an individual with pre-treatment sputum samples submitted for diagnostic purposes and a TB case as an individual with one or more smears positive and / or culture positive for MTB and / or MTB detected on Xpert. Low MDR-TB risk was defined as four weeks previous TB treatment and high MDR-TB risk as >four weeks previous TB treatment, from congregate settings or with a known MDR-TB contact. A PHC site consisted of municipal and provincial health facilities linked to their satellite and mobile facilities and to each other if within a single geographic location (to account for shared diagnostic services).
Study design, population and timeframes We used a non-randomised stepped-wedge study design to assess TB yield in five groups of PHC sites over seven one-month time-points (T1 to T7) as they changed from using the smear/culture-based algorithm to the Xpert-based algorithm (Fig  2 ). We included 60 PHC sites (five of the eight sub-districts in Cape Town) that had implementation of either the smear/culture or the Xpert-based algorithm in October to December 2011. The three sub-districts transitioning in this quarter were excluded to avoid overlap between algorithms within the time-point. Other time-points were selected to also avoid overlap. The study population included all presumptive TB cases with sputum test results recorded in the electronic laboratory database in Cape Town for the selected facilities and time periods.
Data sources and management The National Health Laboratory Services provided TB test data from the electronic laboratory database for 2010-2013. Data included patient demographic information, MDR-TB risk category, test type (smear, culture, or Xpert) and result. Data for each time-point and the preceding and following months (i. e. a quarter's data) were imported into MS-SQL. The full sequence of tests for individuals at T1 to T7 were derived by identifying all tests done in that month as well as in the preceding and following months using a combination of name, surname, clinic folder number and age or birth date.
Demonstration studies done on a moderate scale and at sites selected for good performance tend to over-estimate test efficacy  [17]  and may not reliably reflect findings in routine practice. Many of the studies reporting a significant increase in sensitivity for Xpert compared to smears "diagnostic equipoise" i. e. that sensitivity was not lower with Xpert than with two smears. We assessed the proportion of low and high MDR-TB risk cases with either two smears or an Xpert test at each time point. Both showed increases over time, indicating that although adherence to the Xpert-based algorithm was sub-optimal, cases were better-off than previously for initial testing. As HIV data for presumptive TB cases were not available for all time-points, we could not assess adherence to the second step of the algorithm (i.
Study populations ProtecT (Prostate testing for cancer and Treatment) study. The association of genetic variants with IGF levels was evaluated in the control arm of a case-control study nested within ProtecT, a UK multicentre study to identify localised prostate cancer and evaluate its management in a randomly allocated controlled trial. 5 All men without evidence of prostate cancer were eligible for selection as controls; that is, men with a prostate specific antigen (PSA) test < 3 ng/mL, or men with a raised PSA ( 3 ng/mL) combined with at least one negative diagnostic biopsy. Of the 2,766 controls who underwent measures of IGFs in ProtecT  5  , 700 men also had genome-wide genotype data available (mean age 6 SD: 62. 1 6 5. 0 years). Blood samples for IGF measurement in ProtecT were drawn at the time of the PSA test, frozen at 2808C within 36 hr, then transferred on dry ice for assay.
and phased What's new? Circulating insulin-like growth factors (IGF) and their binding proteins have been associated with prostate cancer risk in observational epidemiological studies but it is not clear whether there is a causal relationship with disease. To address this question, the authors used Mendelian randomization, a method that uses genetic variants as proxies for measured exposures. Their results implicate the IGF pathway in general in prostate cancer development but specific biomarkers remain to be determined. haplotype data from European (CEU) individuals (HapMap release 22, Phase II NCBI B36, dbSNP 126) based on 514,432 autosomal SNPs. After imputation, all SNPs with indication of poor imputation quality (r 2 hat < 0. 3) were eliminated. The working dataset consisted of 2,927 individuals (1,136 cases, 1,791 controls) of European ancestry.
ALSPAC (Avon Longitudinal Study of Parents and Children). We used ALSPAC to replicate ProtecT findings. ALSPAC is a population-based prospective cohort study of children and their parents. The study is described in detail elsewhere  [15] [16] [17]  (http://www. bristol. ac. uk/alspac/). Measurement of circulating IGF-I, IGF-II and IGFBP-3 in plasma or serum was carried out as in ProtecT. IGFBP-2 was not measured. The intra-and inter-assay coefficients of variation (CV) were 7. 0 and 14. 3% for IGF-I, 7. 9 and 18. 6% for IGF-II, and 6. 1 and 8. 7% for IGFBP-3. 18 enome-wide association study (GWAS) data for the children were generated by Sample Logistics and Genotyping Facilities at the Wellcome Trust Sanger Institute (Cambridge, UK) and the Laboratory Corporation of America (Burlington, NC, USA) with support from 23andMe (Mountain View, CA, USA) using the Illumina HumanHap550 quad chip.
The mothers were genotyped at CNG using the Illumina Human660W quad array. All individuals of non-European ancestry, ambiguous sex, extreme heterozygosity, cryptic relatedness (IBD > 0. 1 in children, > 0. 125 in mothers), high missingness (> 3% in children, > 5% in mothers) and insufficient sample replication (IBD < 0. 8) were removed. SNPs with genotyping rate < 95%, MAF < 1%, or out of Hardy-Weinberg equilibrium (p < 5 3 10 27 in children, p < 1 3 10 26 in mothers) were excluded. Genotypic data was subsequently phased with ShapeIT v2. r644,  19  and imputed using IMPUTE v2. 2. 2  20  and phased haplotype data from the 1000 Genomes reference panel (phase 1, version 3), based on 465,740 SNPs. The cleaned dataset consisted of 8,237 children and 8,196 mothers. Up to 400 pregnant women (mean 6 SD age at delivery: 28.
ac. uk/). Serum IGF-I levels were measured using an electrochemiluminescent immunoassay on an IDS ISYS analyser. The inter-and intra-assay C vs. were < 14%. No measurements of IGF-II, IGFBP-2 or IGFBP-3 were available. In total, 10,480 samples were genotyped on the Illumina HumanCoreExome chip (v1. 0) at the Wellcome Trust Sanger Institute. Data QC was performed at the sample-level using the following filters: call rate < 98%, autosomal heterozygosity outliers (> 3 SD), gender mismatches, duplicates as established by IBD analysis (PI_HAT > 0. 9), ethnic outliers. Var We investigated associations of published IGF-related genetic variants, evaluated as instruments in ProtecT and replicated in ALSPAC and/or UKHLS, with prostate cancer risk, progression and mortality in men from 25 studies contributing to the international PRACTICAL consortium 22 (http://practical.
ccge. medschl. cam. ac. uk). Seventeen studies were from Europe, six from North America and two from Australia, and comprised population samples of predominantly European ancestry  22  (Table  1 ). Data on cancer stage, grade and method of diagnosis were collected by each study using a variety of methods. We categorised cancers as localised (T1 or T2 on TNM staging, or if not available, "localised" on SEER staging) or advanced (T3 or T4, or "regional" or "distant" on SEER staging). Genotyping of PRACTICAL samples was carried out using an Illumina Custom Infinium genotyping array (iCOGS), designed for the Collaborative Oncological Gene-Environment Study (COGS) (http://www. cogseu. org/) and.
Cancer Epidemiology Bonilla et al. consisting of 211,155 SNPs. 22 This array was devised to evaluate associations of genetic variants with breast, ovarian and prostate cancer (85,278 were specifically chosen for their potential relevance to prostate cancer). A total of 201,598 SNPs passed QC for the European ancestry samples. 22 Imputation of 17 million SNPs/indels using the 1000 Genomes Project (version 3, March 2012 release) as a reference panel was performed with the program IMPUTE v. 2. 20 Polymorphisms with quality information scores of (r 2 ) > 0. 3 and MAF > 0. 5 were taken forward for analysis. 23 Overall there were 22,992 prostate cancer cases and 22,936 controls with genotype data available. All studies have the relevant Institutional Review Board approval in each country in accordance with the Declaration of Helsinki.
Statistical analysis All SNPs were examined for deviation from Hardy-Weinberg equilibrium using the hwsnp function in the statistical package Stata. Linear and logistic regression were used as appropriate to investigate the effect of SNPs on IGF-I, IGF-II, IGFBP-2, IGFBP-3, PSA and potential confounders. For the validated SNPs we ran meta-analyses across all PRACTICAL studies to evaluate between-study heterogeneity in the association with prostate cancer risk, Gleason grade (low: <7 vs. high: 7) and stage (localised vs. advanced). We computed pooled ORs assuming a fixed-effects model when there was no evidence of heterogeneity (p > 0. 05), otherwise we used a random-effects model. Logistic regression with robust standard errors, to account for within-study clustering, was performed to test for associations of all polymorphisms across the IGFBP1/IGFBP3 region and SNPs in other chromosomal regions with the above prostate cancer outcomes.
Linkage disequilibrium (LD) between pairs of variants in the IGFBP-1/IGFBP-3 gene region was calculated with the program LDlink using data for the GBR population (English and Scottish) in Phase 3 of the 1,000 Genomes Project. 26 r 2 values obtained with LDlink were then used to create an LD plot of the region with the R package LDheatmap (http:// www. R-project. org). Functional consequences of genetic polymorphisms were predicted using SNPnexus (http://www. snpnexus. org/). Survival analysis. Amongst men with prostate cancer, we estimated associations of the validated SNPs with long-term (15-year) survival, examining all-cause and prostate cancerspecific mortality using Cox proportional hazards regression with date at diagnosis as the start date and date at death or final follow-up time-point as the exit date, with robust standard errors to account for within-study clustering.
Acknowledgements The authors thank the tremendous contribution of all members of the  Pro-tecT study research group , and especially the following who were involved in this research ( Prasad Bollina ,  Sue Bonnington ,  Lynn Bradshaw ,  James Catto ,  Debbie Cooper ,  Michael Davis ,  Liz Down ,  Andrew Doble ,  Alan Doherty ,  Garrett Durkan ,  Emma Elliott ,  David Gillatt ,  Pippa Herbert ,  Peter Holding ,  Joanne Howson ,  Mandy Jones ,  Roger Kockelbergh ,  Howard Kynaston ,  Teresa Lennon ,  Norma Lyons ,  Hing Leung ,  Malcolm Mason ,  Hilary Moody ,  Philip Powell ,  Alan Paul ,  Stephen Prescott ,  Derek Rosario ,  Patricia O'Sullivan ,  Pauline Thompson ,  Sarah Tidball ). They thank  Gemma Marsden  and  Luke Marsden , who processed the blood samples at the biorepository, and  Rajeev Kumar , data manager.
Data and simulated samples The data presented in this paper were recorded by the ATLAS detector during the 2012 LHC run with proton-proton collisions at a centre-of-mass energy of 8 TeV, and correspond to an integrated luminosity of 20. 3 fb -1. For the H → γ γ and H → Z Z * → 4 channels, the data collected in 2011 at a centre-of-mass energy of 7 TeV corresponding to an integrated luminosity of 4. 5 fb -1 , are also used. Data quality requirements are applied to reject events recorded when the relevant detector components were not operating correctly. More than 90 % of the recorded luminosity is used in these studies. The trigger requirements used to collect the data analysed in this paper are the same as those described in previous publications  [18] [19] [20]. They are only briefly recalled in the following sections.
5. 2). The spin-2 samples are generated using the MadGraph5_aMC@NLO  [16]  program with LO accuracy for zero, one, and two additional partons, and with subsequent matching of the matrix-element calculation with a model of the parton shower, underlying event and hadronisation, using Pythia 6  [30]. In the H → Z Z * → 4 analysis the signal samples representing the production and decay of Higgs bosons with spin-0 and different parities are generated as follows. The SM Higgs boson production via gluon fusion at the mass m H = 125. 5 GeV is simulated using the Powheg-Box generator. For the non-SM signals, the decays of the generated Higgs bosons are simulated, according to the Higgs boson parity assumptions, using the JHU  [9, 10]  MC generator at leading order (LO). The spin-2 samples are generated using the MadGraph5_aMC@NLO MC generator, as for the H → γ γ analysis.
For the H → Z Z * → 4 analysis, the MC production is only performed for one set of tensor couplings: g 1 = 1, g 2 = 1 + i, g 4 = 1 + i. All other configurations of couplings are obtained by reweighting this sample at generator level. The ratios of the corresponding squares of ME values calculated at LO are used as weights. To calculate these ME values, the JHUGenME  [10]  program is used. In the H → W W * → eνμν analysis, only one MC sample is generated, using MadGraph5_aMC@NLO with parameters κ SM = 1, κ AW W = 2, κ H W W = 2, cos(α) = 0. 3, and all other samples are obtained from it by reweighting the events on the basis of the ME amplitudes. In all the analyses presented in this paper, the mass of the Higgs boson is fixed to 125. 4 GeV  [12].
Asimov datasets  [31, 32]  created from independently generated Monte Carlo samples with κHV V /κ SM and ( κAV V /κ SM ) • tan α equal to 0, ±2, ±4, ±6, ±8 and ±10 were injected into the analysis procedure. The tests were repeated for samples corresponding to 1 and 100 times the LHC Run-I integrated luminosity. In all cases the fitted values of coupling constants were found to be in agreement with the injected values within statistical uncertainties. The results of the matrix-element-observable fit were validated and cross-checked using a nine-dimensional matrixelement method (9D fit). The method implements a multivariate per-event extended likelihood that is sensitive to both the κHV V /κ SM and ( κAV V /κ SM ) • tan α mixing parameters and is based on nine experimental observables. The probability model is constructed with separate components for signal, the SM Z Z * background and the reducible background.
Definitions of terms and data collection Perioperative myocardial infarction (i. e. , within 7 days after intervention) was defined as a creatine kinase-MB !5 times the upper limit of normal, with new Q waves in 2 contiguous leads on the postoperative electrocardiogram or the new development of bundle branch block. After this postoperative period, myocardial infarction was defined as the presence of new pathologic Q waves or the new development of left bundle branch block with increased cardiac marker levels (i. e. , creatine kinase-MB ! 3 times the upper limit of normal). Cardiovascular death included death resulting from an acute myocardial infarction, hospitalization for heart failure, stroke, pulmonary embolism or digestive ischemia. Patients did not undergo systematic control angiography during the follow-up period.
Statistical analysis Baseline characteristics of the cases and control patients were described and compared. Continuous variables are presented as mean ± SD or median and interquartile range, and compared using a Student t-test or Wilcoxon test, according to conditions of application. Categorical data are presented as numbers and percentages, and compared using a χ 2 or Fisher's exact test, according to conditions of application. We assessed the prognostic value of low LVEF, considering it as a time dependent variable, by the use of hazard ratios (HR). Crude and adjusted HR for all-cause and cardiovascular mortality were calculated using Cox semiparametric regression models. Baseline characteristics associated with all-cause or cardiovascular mortality were entered into a stepwise logistic regression model with the significance level to enter set at 0.
Results Baseline characteristics and perioperative course data are described in Tables  1  and 2. During the median [interquartile range] 65  months follow-up, a total of 166 died, of whom 78 in the case group and 88 in the control group. Among these deaths, 107 had a cardiovascular cause, 51 in the case group and 56 in the control group. At a six-month follow-up, a LVEF value was obtained in 93% (n = 383) and 80% of cases (n = 165) in the normal-and low-LVEF groups, respectively. Compared with the preoperative LVEF estimation, the sixmonth LVEF value in the low-LVEF group was improved in 59% of cases (n = 97) and unchanged or worsened in 41% of cases (n = 68), and a LVEF of >45% was achieved in 38% of cases (n = 63). A decrease in mitral regurgitation at 6 months was observed in only 50% cases.
Isolation and characterization of adipose tissue-derived mesenchymal stem cells Human fat tissue samples were obtained through abdominal dermolipectomy from patients undergoing plastic surgery after written informed consent signed by the patients. The study was specifically approved by the French Ministry of Research and Innovation and the Personal data Protection ethics Committee (CPP) of Languedoc-Roussillon (approval AC-2010-1200). In accordance with French law on the protection of personal data, the medical records were anonymized prior to use. Subcutaneous abdominal fat was digested with 250 U/mL of collagenase type II at 37°C for 1h. After centrifugation (300 g, 10 min), the stromal vascular fraction (SVF) was collected, filtered successively through a 100 μm, 70 μm and 40 μm porous filter (Cell Strainer, BD Biosciences, le Pont-de-Claix, France).
Acknowledgments Work in the laboratory Inserm U1183 was supported by the  Inserm Institute , the  University of Montpellier  and funding from the  Centre Hospitalier Régional Universitaire of Montpellier. We are grateful to  Christelle Reynes  ( IGF, Montpellier University ) for advice on biostatistical analyses. We also thank the  Agence Nationale pour la Recherche  for support of the national infrastructure: " ECELLFRANCE: Development of a national adult mesenchymal stem cell based therapy platform"  ( ANR-11-INSB-005 ). Thanks to the " Réseau des Animaleries de Montpellier " animal facility and the  "Réseau d'Histologie Expérimentale de Montpellier " histology facility for processing our animal tissues.
Work in the laboratory Inserm U1183 was supported by the  Inserm Institute , the  University of Montpellier  and funding from the  Centre Hospitalier Régional Universitaire of Montpellier. We also thank the  Agence Nationale pour la Recherche  for support of the national infrastructure: " ECELLFRANCE: Development of a national adult mesenchymal stem cell based therapy platform"  ( ANR-11-INSB-005 ). The funders had no role in study design, data collection and analysis, decision to publish, or preparation of the manuscript.
Analysis of GFP + TEC-derived cells within grafts recovered after 4 weeks revealed a strong contribution of GFP + cells in both cortical and medullary epithelial compartments of all grafts analyzed (Figures  S1B  and S1C ; n = 7). The GFP + cells within the grafts had proliferated, as demonstrated by Ki-67 staining (Figure  S1C ). Collectively, these data validated this assay for detection of thymic epithelial progenitor/stem cell (TEP/SC) activity.
For grafts seeded with PLET1 À mTEC lo , one small Summary of distribution of GFP + cells in grafts seeded with the input test cell populations and test cell numbers noted. Grafts were sectioned at 8 mm and every section was collected. Each section was then screened for the presence of GFP + cells. When foci of GFP + cells were observed, the sections were stained for mTEC and cTEC markers to determine the identity of the GFP + cells. GFP + foci were tracked through sections, with notes taken on the number of sections spanned and the position of foci in the graft relative to other GFP + foci. n is shown for each population tested and represents the number of independent grafts analyzed. Statistical data show mean ± SD for proportion of GFP + clusters in cortical and medullary regions across all grafts analyzed for a given condition.
, 2015; Mayer et al. , 2015). However, to date, prospective isolation and functional testing has narrowed down the adult TEC progenitor compartment only to a broad population comprising MHCII lo cTECs, that comprises $20% of the total adult TEC population and contains both cTEC-and mTEC-generating activities  (Wong et al. , 2014). We have investigated the differentiation potential of six defined subpopulations of adult TECs using a validated transplantation-based assay of TEC potency. Our data show that a previously unidentified EpCAM + UEA1 À Ly-51 + PLET1 + population, that comprises <1% of total TECs, can efficiently generate both cTECs and mTECs at limiting dilution, suggesting that this population contains a common TEPC. No other TEC subpopulation tested contained this activity. Within EpCAM + UEA1 À Ly-51 + PLET1 + TECs, the MHCII hi fraction could efficiently generate cTECs and mTECs, including clones spanning cortex, medulla, and CMJ, again consistent with the presence of a common TEPC, while EpCAM + MHCII lo/neg UEA1 À Ly-51 + PLET1 + TECs generated only cTECs.
Images show immunohistochemical analyses of a graft seeded with 1,000 UEA1 À Ly-51 + PLET1 + GFP + input TECs analyzed 9 months after grafting. Images show representative data from one of two independent grafts. See also Table  S2. many other lineages  (Snippert et al. , 2010; Buczacki et al. , 2013). Overall, in identifying an adult TEP/SC comprising <0. 5% of total TECs, our data represent an enrichment for progenitor/stem cell activity of >30-fold over the previous report  (Wong et al. , 2014). Furthermore, our immunohistochemical analyses indicate that this population is located at the CMJ and consistent with this, CMJlocated PLET1 + TECs were present in grafts derived from EpCAM + PLET1 + Ly-51 + UEA1 À TECs. Our data thus provide the physical location of a phenotypically defined bipotent adult TEC progenitor cell population.