dataesr/scientific-paragraphs-categorization huggingface dataset:
With this data, propensity scores (the probability of being an intervention village) were calculated using logistic regression analysis as has been done in previous studies  [18, 19]. The most closely matched 1001 F and control village pair was then chosen for inclusion, followed by the next most closely matched, etc. All households within chosen villages were then visited and a recruitment questionnaire administered covering a range of demographic, socio-economic and environmental variables. This questionnaire also asked about whether or not 1001 F water was the primary source of drinking water. At this first visit, respondents were asked whether or not they would be prepared to participate in the prospective study. Using just the data from 1001 F villages a propensity score model was derived for whether or not the household consistently used 1001 F water using binary logistic regression.
Data collection The first visits in the prospective part of the study were done on the 8 th December 2011. Each respondent was given a diary in which to record diarrhoea and or vomiting every day for each child. Households were then visited every two weeks for a total of 12 visits. At each visit the interviewer read the diarrhoea diary and asked about whether or not the drinking water source had changed since the previous visit and whether or not the child had been away from home. The case definition for an episode of diarrhoea was 3 or more episodes of diarrhoea in a 24 period or any number plus vomiting.
Data analysis and sample size calculation All analyses were done with SPSS version 18. To determine incidence rate ratios we used Generalized Estimating Equations for negative binomial family and log link. In order to account for repeat sampling of the same individual and possible within group correlation, person, household and village level were specified as subject variables. An autoregressive (1) working matrix was chosen because of a degree of autocorrelation between visits. As well as using propensity score matching to match people in villages without access to 1001 F to users in villages with such access, a range of potential confounding variables with tested in single predictor analyses. Those confounders that were significantly associated with diarrhoea at the p < 0. 2 level were included in a multiple predictor model.
Data Analysis We used Kolmogorov-Smirnov tests to determine whether the data were normally distributed. When the distribution did not fit a normal distribution, we used non-parametric statistics. In adult hens, all morpho-physiological measurements were analyzed using one-way ANOVA for repeated measures (treatment 6 time), and paired t-tests for intra-group comparisons. For the sensory evaluation of yolk from control and treatment eggs, we used a permutation test on the total number of panelists who detected menhaden odor for each vial. Fertility and hatching success were tested by using a Chi-square test. We used Mann-Whitney U-tests to compare the laying rate between groups, and performed oneway ANOVAs for repeated measures (treatment 6 time) to compare the mass of eggs between groups. Variables recorded in the tonic immobility test were analyzed with Mann-Whitney Utests.
The GI value of one food is calculated from 640 data points (ten subjects, eight blood samples, in duplicate, one test series for the test food and three test series for the reference food). The margin of error of , 15 % (i. e. the standard error of the mean expressed as a percentage of the mean) is considered reliable in the context of the considerable day-to-day variation in glucose tolerance in healthy individuals ( ^30 -50 %)  (5). By testing a reference food, the GI method takes into account 'betweenperson variation'. Concerning the accuracy and precision of any nutritional attribute, one cannot let perfect be the enemy of good. For example, both whole-grain and fibre claims are permitted on food labels, despite the fact that the definition and measurement of each varies among nations and is neither perfect nor precise.
The guideline specified 2TU as optimal PPD strength for tuberculin skin testing, to be read after 48-72 hours, with a cut off for significant induration of five mm in SAM children. Each peripheral health worker in Karnataka maintained a register of all SAM children identified in their area during the calendar year. SAM children thus identified were referred to NRCs with a referral slip mentioning the name of the subcenter (health care facility at villages staffed by a health care worker) and a unique identifier for each referred child. When these children reached the NRCs, this information was entered into the registers at the NRCs. NRC registers also captured data on TB evaluation status, the methods used and the outcomes of evaluation including initiation of treatment.
Data management and analysis The data were systematically collected on a standardized data collection form and 10% were re-verified for transcription errors. The data were doubly entered electronically into a predesigned and tested EpiData file (Version 3. 1, EpiData Association www. epidata. dk), validated for consistency and the data finalized. The data were analyzed using EpiData software [Version V2. 2. 2. 182] Frequencies and proportions were calculated for each variable. Differences between groups were compared using Chi square tests and their 95% confidence intervals. A P value of <0. 05 was considered statistically significant.
Acknowledgements This research was supported through an operational research course, which was jointly developed and run by the  Centre for Operational Research ,  International Union Against Tuberculosis and Lung Disease, France , The Union South-East Asia Regional Office and the  Operational Research Unit (LUXOR) ,  Medecins sans Frontieres, Brussels-Luxembourg. Additional support for running the course was provided by the  Center for International Health ,  University of Bergen, Norway.
Search Strategy PRISMA guidelines for systematic reviews and meta-analyses were used  [16]. We compared international organisations guidelines on antibiotics in SAM. The systematic review addressed the question: ''Do children with uncomplicated Severe Acute Malnutrition need antibiotics?''. Sub-questions were: clinical efficacy of AB in SAM; bacterial susceptibilities to AB in SAM; prevalence of infections in SAM according to AB given. Boolean operators and Medical Subject Headings (MESH) were used, through eight databases from January 1960 to August 2011: OVID-MEDLINE, EMBASE, COCHRANE, GLOBAL HEALTH, CINAHL, POPLINE, AFRICA-WIDE-NiPAD, LI-LACS. We searched articles in English, French, Spanish, Portuguese, and German, and translations from other languages (Chinese, Japanese, and Korean). No geographical limits were applied.
Results The review of guidelines is summarized in File S2: it showed several differences for the 2 types of first-line AB (AMX and CTX): five different dosages (CTX 4 or 5 mg/kg/d; AMX 50-100 or 70-100 mg/kg/day or three weight-classes) and two different durations (5 or 7 days). Second-line AB are also shown, and highlight guideline diversity  [5, 8, 19, 20, 21, 22]. Figure  1  shows the systematic review search flowchart and numbers of articles. From eight databases, 2772 titles were found, ranging from 15 (LILACs) to 1304 (EMBASE) excluding duplicates. After reading titles, only 326 abstracts were included. Narrowing to quantitative data on AB efficacy, AB susceptibility, and infection prevalence, 45 articles were included: three RCT, five Cochrane reviews, and 37 observational studies. The data obtained were included in Table  1  for randomised trials, Cochrane reviews, and a cohort efficacy study, and Tables  2  and 3  for observational studies on bacterial resistance and infection prevalence respectively.
Randomised trials and other effectiveness studies Two studies directly addressed our research question: a retrospective cohort-study in Malawi  [23] : it compared two therapeutic feeding programs for children 6-59 months with uncomplicated SAM (WH,23Z), one using AMX for 7 days at 60 mg/kg/day (n = 498), and the other using no AB (n = 1955). It showed worse recovery rates at four weeks in the AMX group (40% vs. 71%), and no evidence of superiority of AMX over the no-antibiotic group at 12 weeks (84% vs. 86%). However these two cohorts were located in two separate districts, and data were collected retrospectively, two major risks of bias. Levels of height and weight for age were lower in the amoxicillin group. Whilst statistical adjustments were made for minor baseline differences in age and anthropometry, unknown confounding or selection bias cannot be excluded, and caution is thus needed interpreting this data.
Bacterial resistance studies Data on AB in-vitro susceptibility were analysed to infer possible in-vivo effectiveness. Although these resistance studies are hospitalbased, the vast majority of infections are community-acquired e. g. 1590 community-vs 212 hospital-acquired pathogens in one study in Kenya  [32] : thus the analysed studies reflected the situation of complicated-SAM children admitted from the community. Our meta-analysis of observational data from 12 AB resistance studies using bacterial cultures (blood, urine, cerebrospinal fluid) in SAM  [33] [34] [35] [36] [37] [38] [39] [40] , and not only SAM children  [35, 41, 42, 43] , is shown.
Grading evidence With regards to methodology, relevance and generalisability, following GRADE scoring framework  [18] , nine studies were considered high grade but not relevant to our question  [27, 28, 29, 31]. One RCT  [24]  available online on the FANTA site, but still not in a peer-reviewed journal qualified as high GRADE: relevant, high-quality in design (ClinicalTrials NCT010000298), high power and low bias for the main outcome, but possibly included some complicated-and HIV-infected SAM children. One RCT  [25]  was relevant for SAM but medium grade due to unblinded randomisation, and lack of separation of uncomplicated SAM. The 37 observational studies were considered as moderate, low, or very low quality. Six were considered as moderate showing strong associations  [33, 34, 35, 41, 46] , low risk of confounding and bias, clear multivariate analysis of confounders.
We would need to test outpatients with uncomplicated SAM to answer our question. The other limitations of our systematic review were mainly due to the lack of robust studies to address the research question. GRADE evaluations showed insufficient relevance to the research question. Most RCTs had strong internal validity, but lacked generalisability. Observational studies were closer to the research question, but presented heterogeneity and inconsistencies Therefore we minimised the risk of bias of the review as much as possible: ''Selection bias'' was reduced by subgrouping similar studies. Source bias and database biases were minimized by using a wide range of databases and grey literature sources: LSHTM library, CAPGAN congress, non peer review or unpublished randomized trials  [24]  books and nutrition guidelines.
Data were obtained in scan mode with a mass range of m/z 50-550 atomic mass unit (amu). Identifi cation of FAMEs was based upon retention times (R t ) obtained for methyl ester of authentic standards, when available. The National Institute of Standards and Technology (NIST) database (version 2. 01) was sometimes used to identify unknown FAs. For positional analysis of carbon double bonds, FAMEs were converted to 4,4-dimethyloxazoline (DMOX) derivatives  ( 29, 30 )  and analyzed by GC-MS in electron impact ionization mode. The column, gas vector, and detector were similar to those used for the analysis of FAMEs (see above). To preconcentrate the minor FA be a mammalian membrane-bound front-end desaturase. However, no function has been attributed to the FADS3 protein yet  ( 15 ). In the absence of any described functional role, investigations have been focused on the tissue distribution of the FADS3 mRNA transcripts in different species.
Plasmid constructs for expressing rat FADS1, FADS2, and FADS3 in COS-7 cells The plasmids constructed for expressing rat FADS2 (pCMV/ FADS2) and FADS3 (pCMV/FADS3) in mammalian cells have been described  ( 11, 24 ). A plasmid coding for rat FADS1 was also constructed. Briefl y, from the published ( 25 ) rat FADS1 cDNA sequence (GenBank accession number AF320509), oligonucleotide primers were designed to PCR amplify the entire coding sequence from the rat liver marathon-ready cDNA library (BD Biosciences). The forward primer (5 ′ -gact cc atg g ctcccgacccggtgcaga-3 ′ ) included the translation start codon (in bold) and Nco I restriction site (underlined). The reverse primer (5 ′ -ctag gcggccg c ta ttggtgaaggtaagcatccagccaga-3 ′ ) contained the translation stop codon (in bold) and Not I restriction site (underlined).
It shows that only the cells expressing FADS2 were able to use RA as a substrate. Indeed, detectable amounts of an octadecatrienoic acid (18:3) with FAME molecular ion at m/z 292 amu and R t = 29. 1 min were produced when compared with control cells. Although not fully characterized in the present study, this FA was identifi ed as the cis 6, cis 9, trans 11-18:3, the known ⌬ 6-desaturated product of RA  (56) (57) (58) , by using the NIST database. Representative example of n = 3 distinct experiments. suggested that the unknown ⌬ 11,13-CLA was not the cis 11, cis 13 or the trans 11, trans 13 isomer. However, because the difference in R t between the cis 11, trans 13 standard isomer and the unknown ⌬ 11,13-CLA was small (only 0. 2 min), time (R t = 28. 0 min). Knowing that the four geometrical isomers of a CLA elute with a specifi c order ( cis -trans , transcis , cis -cis , and trans -trans ) when separated on a capillary column with a polar stationary phase  ( 32 ) , this result fi rst complete conclusive identifi cation, these observations strongly suggested that this FA was likely the trans 11, cis 13-CLA corresponding to a cis -⌬ 13-desaturated product from TVA in cells expressing FADS3.
2  ), was specifi cally detected ( Fig. 7  ). The presence of another small 18:2 (0. 4 ± 0. 1% of FA, n = 5), with R t = 22. 1 min and FAME molecular ion at m/z 294 amu or m/z 295 amu (incubation with 12 C-TVA or 13 C-TVA), similar to the ⌬ 6-desaturated product ( cis6 , trans 11-18:2) Fig. 6. GC-MS profi le of cellular FAMEs extracted from COS-7 cells expressing rat FADS1, FADS2, or FADS3 and incubated with cis -vaccenate. Control COS-7 cells and COS-7 cells transiently expressing the recombinant FADS proteins were cultivated for 24 h with albumin-bound cis -vaccenate (200 µM). The identity of each important FA is indicated above its respective peak. It shows that only the cells expressing FADS1, but not FADS2 or FADS3, were able to desaturate CVA into an octadecadienoic acid (R t = 22. 4 min), which was identifi ed as the cis 5, cis 11-18:2 by the NIST database.
The authors gratefully acknowledge  O. Loreau  ( CEA Saclay and Cadarache, France ) for the supply of [ 1-13 C]-TVA  and the  Transqual  project ( ANR-05-PNRA-No. 5. E. 24 ). The authors thank  P. Toral  for the supply of the lipid extracts from goat tissues and  F. Boissel  for her able technical assistance. The authors also thank  S. D'Andréa , who was at the beginning of this long story, and thank  J. Wilson-Giron  for her English revision of the manuscript.
Copper Deficiency Leads to Anemia in Mice To determine the extent to which copper deficiency affects duodenal iron homeostasis, an accepted nutritional paradigm of copper deprivation was used. Mice at birth were subjected to a copper deficiency (CuD) regime, weaned at P24 and maintained on their respective diets until P45 (Figure  1A ), to ensure that the mice were severely copper deficient. The copper deficient mice at P45 were anemic with decreased hemoglobin (Hb), hematocrit (Hct) and red blood cell count (RBC) (Figure  1B ), along with elevated serum erythropoietin levels, weight loss and cardiomegaly (data not shown). Iron and copper content was measured in the proximal duodenum, plasma and liver using atomic absorption spectrometry. Both copper and iron levels were significantly decreased in plasma from CuD mice (Figure  1C ), including transferrin saturation (Figure  S1A ).
Acknowledgments We thank  Andrija Matak  for assistance with hypoxyprobe immunohistochemistry,  Jean-Christophe Deschemin  ( Institut Cochin ) for experimental assistance and mouse management and  Madeleine Flynn  for graphical design (QIMR). We also thank  Patarabutr Masaratana ,  Robert J Simpson ,  Andrew T McKie  ( King's College London );  Gregory J Anderson  and members of the  QIMR Iron Metabolism Group ;  Bruno Galy  (EMBL) and  Karin Finberg  ( Duke University ) for fruitful scientific discussions.
All participants gave their written informed consent. Baseline data collection included socio-demographic and lifestyle characteristics including dietary habits, symptoms and complaints, main chronic conditions, neuropsychological testing, a physical examination, and blood sampling. Fatty acid analyses were performed at baseline in 1,518 participants from Bordeaux who accepted a blood draw. We excluded 108 subjects with missing data for certain dietary variables and 254 for other missing covariates, including APOE genotype. We further excluded 21 participants who were diagnosed as demented at baseline following the standardized procedure described elsewhere  ( 20 ). Thus the main study sample included 1,135 subjects.
The COGINUT project was carried out with the fi nancial support of the ANR-Agence Nationale de la Recherche-The French National Research Agency under the Programme National de Recherche en Alimentation et nutrition humaine, Grant ANR-06-PNRA-005. C. S. was on a grant from the  Fondation Plan Alzheimer. The  Canada Research Chairs ,  FRQS , and  NSERC  fi nanced S. C. C. 's contribution to this paper. P. B-G. has received funding for travel or speaker honoraria from  Lesieur, Bausch & Lomb, Aprifel, Danone Institute, the Jean Mayer Human Nutrition Research Center on Aging, Tufts University, Groupe Lipides et Nutrition ; has received consultancy fees from  Vifor Pharma ; and receives research support from  Danone ,  Institut Carnot LISA , and  Groupe Lipides et Nutrition. C. S. , S. L. , B. B.
Determination of cholecystokinin secretion using ELISA. A CCK (26-33, non sulphated) fluorescent ELISA immunoassay kit (Phoenix Pharmaceutical Inc. , Burlingame, CA, USA) was used according to manufacturer's instructions. CCK content was determined using 25 μL of supernatant. Fluorescence (excitation 337 nm, emission 460 nm) was read using a FLUOstar Omega multi-mode microplate reader and analysed with the FLUOstar Omega software (BMG LABTECH GmbH, Ortenberg, Germany). Concentration of CCK in samples was determined by extrapolation to a standard curve. Data were expressed as a percentage of the control value (cells with media only). STC-1 cells do not produce detectable level of gastrin, so the detected signal was attributed to CCK  (McLaughlin et al. , 1998). Experiments were performed in triplicate.
ACKNOWLEDGEMENTS S. Le Maux  is currently supported by a  Teagasc Walsh Fellowship  and the  Department of Agriculture, Fisheries and Food (FIRM  project  08/RD/TMFRC/650 ). We also acknowledge funding from  IRCSET-Ulysses Travel Grant. The authors would like to express their gratitude to  Marie-Noelle Madec  for her assistance with confocal imaging.
ACKNOWLEDGMENTS This work was supported by funds from  European Community 's  Health Seventh Framework Programme  under grant agreement No. 259867 , and  Thierry Latran Foundation  to  Jean-Philippe Loeffler ; and " Association pour la Recherche sur la Sclérose Latérale Amyotrophique et autres Maladies du Motoneurone " to  Jose-Luis Gonzalez de Aguilar. Ghulam Hussain  is supported by the  Higher Education Commission of the Pakistani government and "Association pour la Recherche et le Développement de Moyens de Lutte contre les Maladies Neurodégénératives" (ARE-MANE). Florent Schmitt is granted by  "Association Française contre les Myopathies " and  AREMANE. Jose-Luis Gonzalez de Aguilar  is recipient of a  "Chaire d'Exellence INSERM/Université de Strasbourg. ".
Methods This study was embedded within Euro-Peristat, which developed a list of valid and reliable indicators for monitoring and evaluating perinatal health in the European Union (EU)  [18]. Twenty-five EU member states and Norway participated. Detailed information on the design and methods is available elsewhere  [5, 19, 20]. National population-based data for each indicator for the year 2004 were requested in aggregated form from members of the Euro-Peristat Scientific Committee. If national data were not available, population-based regional data could be provided instead. The Euro-Peristat core indicator list includes fetal and neonatal mortality. The fetal mortality rate is defined as the number of deaths before or during birth in a given year per 1000 live and stillbirths in the same year.
The neonatal mortality rate is defined as the number of deaths at 0 to 27 days after live birth in a given year per 1000 live births in the same year. Euro-Peristat collects data on births and deaths at or after 22 weeks of gestation, regardless of birth weight. Aggregated data on the number of live births, fetal and neonatal deaths by each week of gestation and by birth weight intervals of 500 grams were collected. These data were used to calculate fetal and neonatal mortality rates for births and deaths weighing 1000 grams and over and for those born at or after 28 completed weeks. Twenty-seven countries were able to provide data to calculate fetal mortality rates with birth weight and gestational age thresholds. Germany, Hungary, Ireland and Italy did not have data on neonatal deaths by birth weight or gestational age.
Some countries could only provide data on some regions (Valencia in Spain, Brussels and Flanders in Belgium). Data from France came from a one-week national perinatal survey in October 2003, vital registration, and neonatal death certificates. Data on neonatal deaths from England and Wales related to 2005 and data from Italy were for 2003. Table  S1  presents additional information about the data sources. These constraints reflect the diversity of sources for perinatal health data in Europe  [5].
Statistical Analysis We calculated fetal and neonatal mortality rates with 95% confidential intervals, using both birth weight and gestational age thresholds. We also computed differences between rates with confidence intervals to test whether these were significant within countries. To test whether there was a systematic difference between countries in rates based on birth weight versus gestational age we used the non-parametric Wilcoxon signed rank test. In addition, we used the Wilcoxon rank sum test to assess whether rates based on a 28-week threshold minus a 1000-gram threshold differed significantly across countries. Finally, we tested the correlation between these two rates using the Spearman rank test to assess how these affected country rankings. These statistical tests were repeated on recalculated rates after imputation of missing observations, as described above, to ensure that the addition of these data would not change our results.
Conclusions In the European countries included in our analysis, fetal mortality rates calculated using a threshold of 28 weeks of gestation were higher than those based on birth weight cut-offs of 1000 grams, probably due in part to the role of intra-uterine growth restriction in antepartum fetal deaths. Assessing the health burden of third trimester fetal deaths using a cut-off based on gestational age provides valuable additional information. Comparisons based on this cut-off are possible in countries where a clinical estimate of gestational age is recorded in routine data sources and where women have access to early antenatal care and dating ultrasound as is the case in European and other highincome countries.