dataesr/scientific-paragraphs-categorization huggingface dataset:
Design This paper draws on project implementation records and semi-structured interviews. The project has been implemented in stages over 5 years, and is ongoing, as shown in Fig. 1. All of the trials in the project were carried out in Phnom Penh. Details on each of the trials are included under the relevant sub-headings. The trials were registered at ClinicalTrials. Gov (LNS-CAMBIN-FANTS, NCT02257437; LNS-CAMB-INFANTS-EFF, NCT02257762; FLNS_SAM, NCT02907424).
RUSF effectiveness trialchildren A six-month prospective, cluster randomised, non-blinded controlled trial with a 1:1 allocation ratio was conducted with 485 healthy, non-severely acutely malnourished children aged 6 to 11 months. The aim was to establish the novel RUSF's superiority to CSB++, MNP, and a control group. Twenty-eight sites were randomly allocated to one of the four arms. Data collection and food distribution were conducted monthly until endline. The main outcome was anthropometric status, and secondary outcomes were children's body composition, biochemical status, and cognitive development. In addition, long-term acceptability was assessed. The RUSF was not as effective as expected. All groups continued to experience growth faltering, although the RUSF group faltered at a lower rate.
Statistical analyses For the quantification of CK8 + cell clusters, one cryosection out of two through the entire soft palate was analyzed in Neurog1 and Neurog2 KO embryos, wild types serving as controls. This raw number was used as an estimate of the number of CK8 + cell clusters in the palate, without multiplying by two and performing the Abercrombie correction. Indeed, all the clusters we counted in the wild type spanned the whole height of the epithelium and/or had a visible pore, de facto excluding objects that would be tangentially sectioned. This amounts to a systematic loss of 'caps', an important source of errors in the Abercrombie correction  (Hedreen, 1998). Our method makes it next to impossible that clusters (with an average width of 28 mm (E18. 5) and 38 mm (E20. 5), see Source data 1) would be counted more than once on alternate 20 mm sections.
On the other hand, a few clusters were probably lost, if they were centered on one of the uncounted section. The accuracy of our method is verified by the fact that we found the same number of clusters at E20. 5 and E18. 5 despite the 36% increase in average diameter, and also that we found (not shown) the same number of clusters on alternate 30 mm sections (as opposed to 20 mm). For each genotype, all CK8 + clusters (if there were less than 10), or up to ten (if there were more) were imaged, for the wild type in 2 to 3 sections around the midline of the soft palate (where taste buds were the densest), for the mutants, in one out of two sections throughout the palate. Confocal imaging was performed on a Leica SP5 microscope or a LSM 880 Airyscan Zeiss microscope. The surface occupied by each CK8 + cluster was automatically outlined and measured with the Fiji software, and the mean cross-sectional area was calculated for each genotype.
To calculate the entire surface occupied by CK8 + cell clusters on alternate sections of one half of soft palate, the mean value of the calculated surfaces of CK8 + cell clusters was multiplied by the raw, uncorrected number of CK8+ clusters. Statistical analysis was performed using unpaired two-tailed t-test. Results are expressed as mean ± SD. All graphs were performed with GraphPad Prism software. For the reason that we did not count 'caps' of wild type CK8+ clusters (see above), the decrease in total area is likely underestimated.
Acknowledgements We thank the  Imaging Facility of IBENS , which is supported by grants from  Federation pour la Recherche sur le Cerveau, Region Ile-de-France DIM  NeRF 2009 and 2011 and  France-BioImaging. We wish to thank the animal facility of IBENS,  C Goridis  for helpful comments on the manuscript and all the members of the  Brunet laboratory  for discussions. This study was supported by the  CNRS , the  E ´cole Normale Superieure, INSERM ,  ANR  award  17-CE160006-01  (to J-FB),  FRM  award  DEQ2000326472  (to J-FB), the ' Investissements d'Avenir' program  of the French Government implemented by the  ANR  (referenced  ANR-10-LABX-54   MEMO LIFE  and  ANR-11-IDEX-0001-02 PSL Research University ). D-F received a fellowship from the  China Scholarship Council. GGC's lab was funded by the  Italian Telethon Foundation  grant  GGP13146.
Methods We conducted a systematic review and network meta-analysis of individual participant data (IPD). Network meta-analysis (NMA) is an approach which compares effects of multiple interventions using evidence from trials which have directly compared them as well as estimating indirect effects for interventions which have not been compared to each other in a trial  (Salanti 2012). NMA has been described as the 'next generation' of evidence synthesis because it increases the precision of effect estimates for direct comparisons and allows comparisons of interventions which may not have been directly compared  (Salanti 2012). The use of IPD further strengthens the approach leading to decreases in heterogeneity within comparisons and improving overall network consistency because it accounts for effect modification at the individual participant level and missing data can be addressed  (Debray et al.
We searched grey literature, websites, asked experts, contacted authors and screened 299 included studies from relevant systematic reviews. There were no language limits. Two independent reviewers screened titles and abstracts. Conflicts were discussed and resolved with VW or MG. All potentially eligible studies were retrieved and screened in duplicate in full text. The first authors of all eligible studies were invited to join the Trialists' Collaborative in December 2016. In case of no reply, we contacted all authors and their institutions. Data collection was closed in March 2018. Authors signed a data-sharing agreement. IPD were checked for completeness. We verified any questions with the authors. Data were prepared in an excel spreadsheet which included outcomes, covariates and study design variables.
Data analysis Two independent reviewers collected details on participants, interventions, outcomes, study design, setting and duration (Table  1 ) using a pre-tested form. The primary outcomes were weight (kilograms), height (centimetres), haemoglobin (g/L) and cognitive outcomes. We chose to use weight and height in kg and cm, respectively, because these would have the greatest precision to detect differences. To adjust for differences in the importance of a given weight and height change for different ages, sex, anaemia and nutritional status, we adjusted for these factors in our model, as well as infection intensity of each STH. We did not include harms since these have been assessed in previous reviews and are not considered serious (World Health Organization 2017). We intended to assess plasma ferritin but only six studies reported this outcome, thus we decided not to conduct these analyses which the advisory board felt could be misleading.
We used multiple imputations for missing data, based on the assumption that data are missing at random  (Debray et al. 2018; Groenwold, Moons, and Vandenbroucke 2014; Jakobsen et al. 2017)  For studies with less than 50% missing data, we used multiple imputation for baseline and endline and created five complete datasets using Proc MI in SAS9. 4/STAT (SAS Institute Inc. , Cary, NC, USA). All model estimates and standard errors were obtained by fitting the model to each of these five imputed datasets and aggregating results across them using Rubin's Rule which incorporates uncertainty due to imputation. Proc MIANALYZE in SAS 9. 4/STAT was used to obtain the overall estimates across imputed datasets. We used WHO Anthro 3. 2. 2 to calculate BMI-for-age, and heightfor-age z-scores. Anaemia was defined according to age and sex using WHO standards and corrected for altitude, if needed (World Health Organization 2011).
2016). We used the GLIMMIX procedure in SAS 9. 4/STAT (SAS Institute Inc. , Cary, NC, USA) for the GLMM NMA, considering models that account for multi-arm trials and adjust for covariates. Results are summarised as point estimates with 95% confidence intervals for the outcomes of weight (kg), height (cm), and haemoglobin (g/L). Data were analysed as intention to treat. The transitivity assumption was assessed by comparing the distribution of effect modifiers for each comparison  (Salanti 2012 ). The evidence network was designed with the advisory board, research team and trial authors based on clinical judgement and other evidence. The consistency assumption was assessed by evaluating the heterogeneity of direct comparisons (using visual inspection of forest plots and I 2 )  (Higgins and Thompson 2002)  and comparing the effect estimates for direct evidence (aggregate), direct evidence (IPD) and indirect + direct evidence (NMA).
Search and study identification We screened 14034 titles and abstracts. Three hundred and forty studies were screened in full-text. Fifty-five studies were considered eligible (Figure  1 ). Reasons for exclusion at full-text were not randomised or quasi-randomised trial (n = 262), no baseline infection intensity (n = 14) or too short (9 studies were <4 months duration) (Appendix S3). We obtained data from only two trials of deworming for schistosomiasis (2/14 studies). The advisory board decided not to analyse these since results would not be representative of the totality of the evidence. We obtained data for 79% of children  (31,945 of 40,525)  randomised to eligible studies of STH deworming (19 of 42 eligible published trials). For the 22 studies that we could not retrieve, reasons for not providing data were non-response (n = 7 studies), lost data (n = 10 studies) and administrative issues (n = 5).
Comparison of studies retrieved with those not We used aggregate data from our previous Campbell review to compare studies of deworming for STH vs. placebo for which we received data compared to those for which data were not received. The effects on weight gain were 0. 02 kg (-0. 04 to 0. 08) for studies received vs. 0. 17 kg (-0. 11, 0. 44) for those not received (interaction test for subgroup differences: p = 0. 31). Studies for which we did not receive IPD had a larger pooled effect size on weight gain with a wider confidence interval.
Thirdly, we updated previous meta-analyses by adding previously unpublished data obtained from authors for this review. These analyses reinforce the findings of population-level small effects on average for weight. This included three previously unpublished estimates for weight. This analysis replicated estimates of three prior systematic reviews of an average effect size on the weight of about 100 g and showed that this effect size was not related to the year of publication. The findings are based on trials that have an overall low risk of bias, and main effects were robust to restricting to the lowest risk of bias studies. We considered the assumptions of transitivity and consistency for network meta-analysis were plausible based on our analyses. In 2010, the prevalence of all STH infections was below 30% for all regions of the world for each STH which is somewhat lower than the average prevalence of 46% in our sample  (Pullan et al.
39 kg annually  (Kristjansson et al. 2007 ). Our systematic review cannot predict outcomes or cost-effectiveness for chemoprophylaxis where the of infection is very heavy, since less than 2% of our sample had heavy intensity infections according to WHO cut-offs. Strengths are a comprehensive search and transparent methodology described in an a priori protocol. We included an expert advisory board with expertise in nutrition, parasitology, economics and methodology and a trialists' collaborative. We were able to obtain data from 79% of children randomised to eligible deworming studies. We accounted for differences in baseline nutritional status, age, sex, anaemia and infection intensity and also assessed possible effect modification across these factors. The results were robust to a range of sensitivity analyses including complete case analysis and restricting to studies at low risk of bias.
We were able to include data from studies which may not have been designed or powered for these outcomes. A limitation of our analyses is that we were unable to retrieve 22 out of 41 eligible studies (which included 8,580 children), despite, in most cases, the goodwill of the original trialists, due to inability to locate the data, administrative barriers to sharing data or lost datasets. Another limitation is that only 2% of children in these studies had heavy intensity infections defined according to WHO cut-offs, which limited our assessment of effects across intensity, in particular for children with heavy intensity infections. These analyses across the intensity of infection are limited by measurement issues since WHO cut-offs for intensity are arbitrary and their importance may vary according to different contexts.
Acknowledgments We would like to thank  Celia Holland  for being part of our Advisory board and providing comments on interim results. We would like to thank  Bishop Beasley ,  Nilanthi de Silva ,  Rebecca Stoltzfus  and  James Tielsch  for contributing data. We would like to thank  Jessica Trawin ,  Christine Mathew , who contributed to data collection and preparation; and  Jordan Bernick  who contributed to the analysis. We would like to thank anonymous peer referees who provided constructive comments that helped us improve on an earlier version of this paper. We would like to acknowledge the  Campbell Collaboration International Development Coordinating group  which managed the peer review process for the protocol and final review. This systematic review will be co-published with the  Campbell Collaboration.
Inclusion criteria Families with at least one child aged from 3 to 10 years old, living in the northern districts of Saint Denis were eligible to the study. For families with several children matching with the age criterion, we included the child with the closest anniversary birth date (regarding the inclusion date). In addition, adult participants had to be unemployed, or benefit from social minima (Active Solidarity Income, Allocation of minimum pension) or any income-terms allowance, or have incomes below the poverty line. The poverty line threshold was defined using the French National Statistical Institute (INSEE) according to the French incomes data (that is, 1234€ per month rounded up to 1300€ for a single-parent household with at least one child aged under 14 years old, and 1777€ per month rounded to 2000€ for a couple with at least one child aged under 14 years old)  [36, 37].
Data collection Data were collected via face-to-face questionnaires administered by trained interviewers at inclusion, after 6 months and one year. Volunteer families were interviewed at community centres, or at home in order to sign the consent form, and complete the questionnaires. Questionnaires were adapted from those used in the ABENA study, which was specifically designed to be administered to disadvantaged groups  [20]. A food frequency questionnaire was used to describe the consumption of children and adults in 13 main food groups (cereal products, starches, vegetables, fruits, legumes, dairy products, meats and eggs, fish and sea-food products, fast-food and pizza, salty snacks, sweet products, and beverages). The baseline questionnaire also included information on inclusion criteria, living conditions, and food security (see details below).
We relied upon the latter data and sociodemographic characteristics to compute the EPICES score  [38]. Based on 11 questions on various socioeconomic determinants, this individual score assesses the precariousness level of subjects living conditions. It ranges from 0 (the less precarious situation) to 100 (the most precarious situation), with a threshold of 30. 17 to define precariousness, a score upper than 53. 84 reflecting a great precariousness. The baseline questionnaire had a duration of about one hour, and follow-up questionnaires about 30 min. The vouchers were electronically traceable, so we were able to know exactly the number of vouchers used for each household. After the interview, parent-child pairs were assigned in the intervention or in the control group through an algorithm of random distribution performed on a laptop.
The study was approved by the Ethics Review Committee of the National Institute of Health and Medical Research (Institut National de la Santé et de la Recherche Médicale) (Inserm) IRB00003888 under the number 15-247. The declaration to the National Commission of Data Processing and Liberties (Commission Nationale de l'Informatique et des Libertés) (CNIL) of February 26, 2015 was made under number 1838429v0. The study protocol has been registered on clinical trials website under no. NCT02461238.
Acknowledgements The authors are grateful to  Claudia Chahine  and  Cynthia Perlin , dieticians, for their precious participation to this study and the elaboration and conduction of workshops during the whole study, and to the FLAM interviewers,  Laëtitia Defoi ,  Anita Houeto ,  Anouchka Kponou ,  Lysa Tagherset  and  Marion Genest. The authors thank for their welcome and their help the community centres of  Saint-Denis  ( Pierre Sémard ,  Romain Rolland , Floréal), the Municipality health care centres ( CMS du Cygne, CMS Henri Barbusse, CMS des Moulins ) and their directors ( Dr Brahim Bouselmi ,  Jean-Marc Robinet  et  Dr. Samia Khimoun ). They also thank all the associations which have been involved in this study: APIJ association, Maison des parents, ludothèque of Allende neighbourhood and ASAFI association.
thaliana GRX C5, and commercially available E. coli TRX (Sigma-Aldrich; protein id AAA24693), were conducted through 5, 15, and 30 min of incubation of 20 µM glutathionylated AtAMY3 in the presence of 2 mM GSH with or without 5 µM GRXs, or in the presence of 0. 2 mM DTT with or without 10 µM TRX. Upon incubation, enzyme activity was measured on 5-fold diluted samples as described above. All data are expressed as percentages of activity relative to the fully reduced sample obtained by incubating glutathionylated AtAMY3 in presence of 80 mM DTT for 30 min at 25 • C.
Determination of pK a Value Unreactive cysteines are characterized by pK a values around 8. 5, whereas reactive cysteines are typically surrounded by unusual microenvironment that destabilizes the unreactive protonated form of cysteine residues through electrostatic interactions, significantly decreasing their pK a value  (Roos et al. , 2013). IAM is a strong alkylating agent able to react specifically with thiolate anion (-S -). When cysteine residues are involved in the catalytic process, the pK a of catalytic relevant cysteines can thus be determined by measuring the pH dependency of IAM inactivation. AtAMY3 was treated with IAM at different pH values and the activities were compared with those measured at the same pH but in absence of the alkylating reagent. Experimental data were fit both to a one-pK a -dependent (Supplementary Figure  S4 ) and to a two-pK a -dependent event (Figure  5 ).
To reduce potential bias caused by population stratification, we restricted the stroke dataset to individuals of European ancestry. Thus, our analyses included data from up to 404,630 noncases and 34,217 ischemic stroke cases, subtyped into cardioembolic stroke (n = 7,193), large artery stroke (n = 4,373), and small vessel stroke (n = 5,386). Stroke subtypes were classified according to the Trial of Org 10172 in Acute Stroke Treatment criteria. 21 andard protocol approvals, registrations, and patient consents Each study included in the GWAS used in the present study was approved by an institutional review board, and all participants had provided informed consent.
Statistical analysis The primary analyses were conducted using the inversevariance weighted method (hereafter referred to as standard MR analysis), which gives accurate estimates if all SNPs satisfy the instrumental variable assumptions (data available from Open Science Framework, figure e-1, osf. io/ b57sq/). 22 In sensitivity analyses, we used other MR approaches, including the following: (1) the weighted median method, which provides consistent estimates if at least 50% of the weight in the analysis comes from valid instrumental variables 22 ; (2) the heterogeneity-penalized model averaging method, which gives consistent estimates if a plurality of the instrumental variables are valid  23  ; and (3) the MR-Egger method, which can detect and adjust for pleiotropy. 22, 24 The MR-Egger analysis is disposed to regression dilution bias.
3 × 10 -4 ) for all ischemic stroke, 0. 63 (95% CI 0. 50-0. 80; p = 1. 6 × 10 -4 ) for cardioembolic stroke, 0. 60 (95% CI 0. 44-0. 82; p = 0. 001) for large artery stroke, and 0. 90 (95% CI 0. 67-1. 20; p = 0. 46) for small vessel stroke (figure  1 ). Only the association with cardioembolic stroke remained in sensitivity analyses (figure  1 ). The I 2 GX value from the MR-Egger analysis was 0. 87, indicating 13% dilution of the estimates. The MR-Egger analysis, with adjustment for dilution bias using the SIMEX method, provided imprecise estimates (data available from Open Science Framework, table e-3, osf. io/b57sq/). In this analysis, genetically predicted serum magnesium concentrations were associated with cardioembolic stroke but the CI included the null (OR 0. 66, 95% CI 0. 21-2. 10); there was no evidence of directional pleiotropy (data available from Open Science Framework, table e-3).
In contrast, directional pleiotropy was detected in the analysis of large artery stroke, and this was not explained by any single SNP (data available from Open Science Framework, table e-3). MR-PRESSO analysis identified potential outlying SNPs (at p < 0. 10), which varied for different subtypes (data available from Open Science Framework, table e-4, osf. io/ b57sq/). The association of genetically predicted serum magnesium concentration with cardioembolic stroke persisted after exclusion of the outlier in TRPM6 (OR 0. 56, 95% CI 0. 43-0. 73). The association also remained after exclusion of 2 SNPs associated with estimated glomerular filtration rate and 1 SNP associated with blood pressure and serum urate levels, but was attenuated after omitting 2 SNPs associated with atrial fibrillation (OR 0.
Serum calcium None of the calcium-associated SNPs was statistically significantly associated with ischemic stroke as a whole or any subtype (data available from Open Science Framework, table e-6 and figure e-2, osf. io/b57sq/). There were no associations between genetically predicted serum calcium concentrations and any stroke outcome in the standard MR analysis (figure  2 ). The OR of all ischemic stroke per genetically predicted 0. 5 mg/dL (about 1 SD) increase in serum calcium concentrations was 1. 03 (95% CI 0. 88-1. 21; p = 0. 68). The lack of association remained in sensitivity analyses (figure  2 ), and there was no evidence of directional pleiotropy in the MR-Egger analysis (data available from Open Science Framework, table e-7, osf. io/b57sq/). The I 2 GX value was 0. 96, indicating no significant dilution bias in the MR-Egger analysis.
25, 95% CI 1. 08-1. 45). 17 The estimates for serum calcium and cardiometabolic stroke are also similar to those for calcium supplementation and stroke from a meta-analysis of 8 randomized controlled trials (relative risk 1. 15, 95% CI 1. 00-1. 32; p = 0. 06). 4 It is unclear why genetically predicted serum calcium concentrations were not associated with large artery stroke, which like coronary heart disease is related to atherosclerosis. A possibility is that we may have overlooked an association because of low power (data available from Open Science Framework, table e-1, osf. io/b57sq/). A major strength of this MR study is that biases that can be of concern in conventional observational studies were avoided. Other important strengths are the large number of cases of ischemic stroke and that associations with ischemic stroke subtypes could be investigated.