dataesr/scientific-paragraphs-categorization huggingface dataset:
Assessment of Sampling Bias Prior to further analyses, we checked for obvious bias in our data attributable to differential sampling effort. If some areas or taxa are systematically more heavily sampled than others this may lead to bias. An underlying problem in quantifying this bias is that overall sample effort of botanists over the last two hundred years is unknown, as is the 'true' distribution of narrow endemic species, and the degree to which there is an interaction effect between sampling practices and diversity. However, in New Caledonia, we do have the ability to assess this issue for a large proportion of the dataset, as almost a third of the records come from a single collector whose sampling effort is documented. To determine if the number of NES was correlated with sampling effort, we assessed whether the collection effort of H.
9% (n = 1867) at the Zurich herbarium (a total of 27% of the records in the current study). In terms of species, 75. 2% of the species in New Caledonia were collected at least once by MacKee and deposited at the Noumea herbarium. We conclude from this, that although some degree of recording bias in data of this type must be inevitable, that there is at least no markedly obvious bias in terms of sample effort and the recorded distribution of NES, for the component of our data in which this can reasonably be tested. Further exploration of the likelihood of some areas being under-recorded is undertaken in the species distribution model component of this paper.
Analysis of NES from Observed Distributional Data We divided NES into three groups according to their range: narrow endemic species restricted to one location (NES 1 = sample positions are separated by less than 10 km), narrow endemics of two locations (NES 2 = not more than two sample positions separated by more than 10 km) and narrow endemic species restricted to 3 different locations (NES 3 = not more than three sample positions separated by more than 10 km). We recorded the number of species, which fell into each of these three categories. This definition of NES focuses on distributional discontinuities. There are a small number of cases where NES have a continuous distribution of records across a large area which become classified as a single site on account of the absence of a distributional discontinuity.
The IUCN status was recorded for each NES using the IUCN Red List website (www. iucnredlist. org). For the IUCN categories that were relevant to the dataset (extinct (EX), critically endangered (CR), endangered (EN), vulnerable (VU), lower risk (LR), near threatened (NT) and least concern (LC), we recorded which criterion was used for making the assessment (A -Declining populations, B -Geographic range size, and fragmentation, decline or fluctuations, C -Small population size, and fragmentation, D -Very small population or very restricted distribution, E -Quantitative analysis of extinction risk). We also recorded the number of species, which do not have an IUCN classification, and whether this was because they were 'data deficient' or whether they had simply not been assessed. To relate IUCN status to distributional data we calculated the Area of Occupancy (AOO) and the Extent of Occurrence (EOO) of each NES, according to IUCN guidelines  [12]  using Mapinfo 10.
5 when precise distributional data was available. These data allow us to compare distributional trends between NES with and without IUCN status. For each of the higher risk IUCN categories (CR, EN and VU) we correlated species with IUCN assessments against those NES 1, 2, & 3 species, which lack an IUCN assessment. This was to establish whether there are groups of unassessed species, which show similar distributional patterns to known high-risk species. This was undertaken with a Pearson's correlation coefficient. To identify areas with high density of NES, we applied a grid of cells measuring 4 km 2 (262 km) on the map of New Caledonia. For each type of narrow endemism (NES 1, NES 2, NES 3) the number of species occurring in a cell was identified. In this study, although we map all NES, we define a HPNE as where a total number.
For precipitation we interpolated average mean monthly rainfall (over a period of 10 years from 1991-2000) from 121 weather stations using two different methods: (1) trivariate thinplate splines based on x and y coordinates and elevation, implemented in GRASS GIS  [28] ; and (2) AURELHY (Analyse Utilisant le RELief pour l'HYdrometeorologie) -a method designed by Meteo France  [29]  currently under development as a package for R  [30]. The AURELHY method consists of 2 steps: first, the topography of the landscape is described at a range of scales by means of a Principal Component Analysis. Second, the weather station data are modelled using their x and y coordinates and the principal components as explanatory variables in a linear model whereby the residuals are krigged. We used the 90 m resolution digital elevation data for topography and derived the interpolation at a resolution of 1 km.
Due to the linear nature of the model some overshoots were produced, and the predictions were truncated at 4900 mm in 0. 2% of the cells. A jack-knifing procedure showed that the AURELHY method provided a closer fit to the weather station data than the thin-plate spline interpolation, and our AUREHLY interpolation also captured regional precipitation patterns better than the WorldClim and the downscaled Tropical Rainfall Measuring Mission 2B31 (atmo-spheric rainfall remotely sensed between 1997-2006;  [31] ) datasets. Consequently the AURELHY rainfall layer was used in all subsequent analyses. For geology we used a detailed and precise vector map (1:50 000)  [32]  and assigned the original geology categories to 4 different groups that are thought to be of importance for plant distributions: ultramafic, volcano-sedimentary, limestone, and riverbanks/anthropic formations.
Grouping Species Based on Environmental Similarity As not all of these species have precise locality information suitable for linking to environmental data, we subsampled the 552 species from which we have precise record data and used these in the current study. For each species, we calculated the overall mean of environmental variables from the locations at which a species occurred (altitude, slope, total annual rainfall and mean annual temperature; all layers were resampled to a 1-km resolution as even though topography data exist at a finer resolution the specimen localities often do not). For the categorical variable geology we used the mode. Based on this information we derived a distance matrix between all species using Gower's distance (symmetric distance measure designed to work with mixed data).
00001; fit regularization parameter = 1; default prevalence = 0. 5), and we set aside 25% of the data as test data. Obviously these test data are only semi-independent as some will have been collected during the same sampling trips as the model training data and might therefore be subject to the same biases. Consequently, the test AUC values are likely to be somewhat higher than when based on a fully independent test dataset, and need to be interpreted with appropriate caution. The input data consisted of 86, 290 and 516 presence records for groups 1, 2 and 3 respectively (three ecologically distinct groups detected in the preceding analyses and described in the results section), and we modelled at a 1 km resolution.
Diversity of NES and their IUCN Classification In the current paper we present the first formal floristic-scale assessment of the extent of localised endemism in New Caledonian plants. We show that a high proportion of the flora can be classified as 'Narrow Endemic Species' based on the available data. In total 22% of the assessed species in the flora have a maximum of three distributional points (defined as a record, or a cluster of records separated by. 10 km from other records). It is of course possible that further survey work will identify additional populations of these species and we fully accept that some of these species will turn out to be more common. Nevertheless, this represents the best currently available dataset on the distribution of narrow endemic species in the New Caledonian flora.
Acknowledgments The authors would like to thank  P. P. Lowry ,  M. Callmander ,  B. Suprin ,  V. Hequet ,  G. Dagostini ,  L. Barrabe ,  F. Rigault ,  J. Tinel ,  P. Gailhbaud ,  T. Le Borgne  and  F. Bart  for providing distributional data on NES. We are grateful to  X. Arcas ,  P. Birnbaum ,  J. Munzinger ,  J. Piquet ,  A. Ollivier  and  I. Wulff  for their help in the accomplishment of this paper.
Study design and patient selection The current multicenter retrospective study is based on 7 consecutive patients with lupus enteritis, referred to two tertiary care centers (department of internal medicine, Hôpital Pitié-Salpêtrière and department of clinical immunology, Hôpital Saint Louis, Paris, France) between January 1990 and December 2011. These patients were identified from computerized databases, which records the main diagnoses for each patient, and from an electronic review of all medical records. The databases were searched for the International Classification of Diseases 10th Revision (ICD-10) code for SLE (code M32) as well as for "Lupus enteritis". Additional cases were contributed after hand-search of patient files. The medical records of all patients identified were reviewed by 2 physicians (PJ and LA) to ensure that patients fulfilled the revised ACR criteria for SLE  [9, 10]  and had clinical and radiological evidence for small bowel involvement (Small bowel wall edema, abnormal bowel-wall enhancement [double halo or target sign], dilatation of bowel lumen and mesenteric abnormalities such as engorgement of mesenteric vessels, increased number of visible vessels [comb's sign], and increased attenuation of mesenteric fat).
Study setting and population All consecutive patients over 18 admitted to our 24-bed medical ICU between July 2006 and March 2008 after a successfully resuscitated CA were eligible. We retrospectively reviewed all medical records and data from our prospectively acquired ICU database, in which all CA survivors' characteristics are registered according to the Utstein style  [15]. The following data were extracted prospectively for each patient: demographic data, clinical parameters, cause of CA, "no-flow" and "low-flow" period (respectively, time from collapse to basic life support and time from basic life support to return of spontaneous circulation), initial rhythm, Simplified Acute Physiology Score II (SAPS II) and Sequential Organ Failure Assessment (SOFA) scores, hypothermia management, biological parameters and ICU mortality.
Statistical analysis Continuous variables were expressed as medians (with interquartile range) and qualitative variables were reported as count and proportions, unless specified otherwise. Statistical analysis compared ICU survivors and non-survivor patients with nonparametric tests, as appropriate: continuous variables with the Wilcoxon rank sum test; categorical variables with the χ 2 test. Statistical significance was defined as P < 0. 05. Correlation was performed with the Spearman test. Receiver-operated characteristic (ROC) curves were performed to assess the ability of TRX concentrations to predict ICU death. Results were expressed with area under the curve (AUC) and 95% confidence interval. Analyses were performed with Stata 7. 0 software (StataCorp. , College Station, TX, USA).
In the present study, we confirm the limited interest of AOPP measurement and thiol content determination in acute setting, at least after CA, despite their place in experimental conditions  [5, 6]. On the contrary, high CRP and PCT levels are consistent with previous findings suggesting the overwhelming impact of systemic inflammation  [4]. Overall, these results enhance the value of TRX after CA. Despite the large number of post-CA patients enrolled, some limitations of this study need consideration. First, we considered a retrospective single-institution cohort. However, all analyzed data were prospectively collected, and medical management was homogeneous. Second, this is a merely observational study, allowing only association rather than causation conclusions. Third, the utility of TRX to predict death was evaluated as a single parameter, without combination of TRX level to clinical or biological data that may have improved the overall prognosis value.
Fourth, measurement of other oxidative stress parameters, like MIF or manganese superoxide dismutase (MnSOD), or main inflammatory cytokines would have provided further mechanistic tracks. Similarly, determination of the TRX interacting protein (TXNIP) could give valuable data to help explain the TRX kinetics after CA. However, TXNIP is an intracellular protein that cannot be routinely measured in serum samples. Fifth, CA has the particularity of exhibiting two main causes of ICU death, that is, shock and neurological damage with subsequent care withdrawal. As inflammation and oxidative stress are expected to be greater in case of shock, subanalysis focused on this last group. Finally, the vast majority of patients underwent therapeutic hypothermia. If the relationship between inflammatory cytokines or biomarkers and effect of therapeutic hypothermia is controversial  [4, 25] , the influence of hypothermia on TRX is as yet, unknown.
Between 16 and 20 weeks of age, there was a significant increase in serum CPK activity in Lmna H222P/H222P mice treated with DMSO; however, CPK activity did not significantly increase in the mice that received selumetinib and at 20 weeks it was significantly lower than in those that received DMSO (Figure  6C ). Mean serum AST activity was also significantly lower in the selumetinib-treated mice compared to the placebotreated mice at 20 weeks of age (data not shown). To determine if selumetinib improved skeletal muscle function in Lmna H222P/H222P mice, we evaluated limb grip strength. At 20 weeks of age, mean grip strength was significantly greater in selumetinib-treated Lmna H222P/H222P mice than in DMSO-treated mice (Figure  6D ). Hence, selumetinib improved skeletal muscle dystrophic pathology and improved function in Lmna H222P/H222P mice.
Acknowledgements We thank  Dr. Gisèle Bonne  ( Institut de Myologie ) for providing Lmna  H222P/H222P  mice and  Dr. Michio Hirano  ( Columbia University ) for assistance with grip strength testing. This work was supported by grants from the  National Institutes of Health/National Institute of Arthritis and Musculoskeletal and Skin Diseases  ( AR048997 ) and  Los Angeles Thoracic and Cardiovascular Foundation  to HJW and a grant from the  Association Franaçise Contre les Myopathies  to AM.
Data collection A semistructured questionnaire uploaded in the AfyaData app  ( Karimuribo et al. , 2017 )  on smartphones was used to collect data from adult individuals ( ≥18 years old). The data collected were those related to sociodemographic characteristics and to animal and nonanimal behavioral risk practices (based on a recall period of 12 months before the interview date). Animal data collected included species, age, sex, fever on sampling day, and history of abortion in the 12 months before sampling. For the animals sampled from slaughtering facilities, the information about history of abortion was limited to only the period in which the stock owner had kept the animal, i. e. , between purchase and slaughtering time. In the absence of exact birth records, the animal's age was identified through a combination of information provided by the farmer's recall and information obtained from the dentition technique  ( Herzog et al.
Data analysis The epidemiological data were downloaded from the server, matched with laboratory results in Microsoft Excel based on individual identification numbers, and imported into STATA 13. 1 (Stat-aCorp, College Station, TX, USA) for analysis. The potential behavioral risk practices and demographic variables were analyzed for their association with RVFV seropositivity using a multivariable logistic regression model. A binary variable describing RVFV status was created and given a value of "1" (positive) when anti-RVFV IgG was detected and "o" (negative) when it was not detected. The Spearman rank-order correlation was run to assess the relation-ship and direction of the association between RVFV seropositivity in animals and humans  ( Zar, 1972 ; Altman, 1990 ). The discriminatory ability of the final model (the discriminatory accuracy of the combination of risk factors) was assessed using receiver operating characteristic (ROC) curves and quantified using the area under the curve (AUC)  ( Pepe, 2003 ; Royston and Altman, 2010 ).
Acknowledgments We are grateful to the  President's Office, Regional Administration and Local Government (PO-RALG)  and to Regional and District authorities in Dar  es Salaam , Pwani, and  Mwanza  for their support and permission to carry out this study in the study districts. We thank  Martin Anditi ,  Christina James ,  Feliciana Rawille ,  Martin Zuakuu ,  Wema Sudi , and  Hamisi Bucheye  for their support in data and sample collection in the study districts. Mariam Makange  and  Gaspary Mwanyika  are thanked for their laboratory technical assistance.
Acknowledgements We are grateful to the patients and the healthy volunteers for their blood donations. We thank  Mike Waring  and  Adam Chicoine  for excellent assistance with flow cytometry through the  Harvard University Center for AIDS Research (HU CFAR) Immunology Core at the Ragon Institute. This work was supported in part by research funding from  MGH/ECOR (Physician Scientist Development Award to MMA and Fund for Medical Discovery  to SJ) and the  NIH NIAID  ( KO8 AI074405  to MMA,  R01 AI067031  and  P01 AI074415  to MA). This research was further supported by the  Terry and Susan Ragon Foundation  and the  Harvard University Center for AIDS Research (CFAR) , an  NIH  funded program ( P30 AI060354 ) which is supported by the following  NIH  Co-Funding and Participating Institutes and Centers:  NIAID ,  NCI ,  NICHD ,  NHLBI ,  NIDA ,  NIMH ,  NIA ,  FIC , and  OAR.
Searches and Study Selection Extensive literature searches were performed between August 2011 and March 2012 (see Data in online-only Data Supplement); this included searching the gray literature. Two review authors scrutinized the searches of Cochrane Stroke Group, CENTRAL, CCDAN and the trials registers, and applied inclusion criteria. One review author scrutinized the other searches and applied inclusion criteria. We included all randomized controlled trials in patients with a clinical diagnosis of stroke, in which SSRIs were given within the first year of stroke, for any clinical indication. The control arm included usual care or a placebo. Any drug classified as an SSRI (for example fluvoxamine, fluoxetine, sertraline, citalopram, escitalopram, and paroxetine) given at any dose, by any mode of delivery, was included.
Topical Reviews Selective Serotonin Reuptake Inhibitors for Stroke Recovery A Systematic Review and Meta-analysis Gillian E. Mead, MD, FRCP; Cheng-Fang Hsieh, MD; Rebecca Lee, MRCP; Mansur Kutlubaev, MD; Anne Claxton, BSc; Graeme J. Hankey, MD, FRACP; Maree Hackett PhD for meta-analysis, 1 of which reported data separately for depressed and nondepressed people, 7 therefore we considered this as 2 separate trials. Thus, there were 52 trials randomizing 4059 patients to SSRI or control; 28 used fluoxetine, 2,7-32 7 used sertraline,  [33] [34] [35] [36] [37] [38] [39]  10 used paroxetine,  [40] [41] [42] [43] [44] [45] [46] [47] [48] [49]  5 used citalopram,  [50] [51] [52] [53] [54]  1 used escitalopram,  55  and 1 used either sertraline or fluoxetine. 56 The usual care arm was not the same in all the trials because of geographical variation in the management of stroke.
Subject Characteristics Of the 52 trials included in the meta-analysis, the mean age of patients ranged from 55 years  24  to 77 years. 35 Most excluded those with dementia or communication difficulties. The mean time since stroke was 0 to 3 months in 36 trials,  2, 7, 9, 10, [12] [13] [14] [15] [16] 18, [20] [21] [22] [23] [24] [25] [27] [28] [29] [31] [32] [33] 35, 38, 39, 41, 43, 44, 46, [48] [49] [50] [51] [52] 55  3 to 6 months in 4 trials,  11, 37, 47, 54  6 to 9 months in 2 trials,  36, 53  and not reported in 10 trials. 8, 17, 19, 26, 30, 34, 42, 45, 56  In 16 trials, a diagnosi  of depression (however made) was not one of the inclusion criteria. 2, 7, 8, 11, 14, 18, 20, 24, 27, 31, 32, 34, 38, 46, 50, 55  In the other 36 trials, participants had to have depression to be included. Five ongoin  trials were identified (see Data in the online-only Data Supplement).
Discussion This is the most comprehensive and up-to-date systematic review of SSRIs in stroke. It includes 52 completed trials in which 4059 patients were recruited. We had not expected to identify such a large number of trials, based on our knowledge of previous reviews. Much of the literature is from China. At the end of treatment, patients allocated an SSRI were less likely to be dependent, disabled, neurologically impaired, depressed, or anxious. The favorable effects of SSRIs on disability, dichotomous depression scores, and neurological deficit scores were greater in participants who were depressed at randomization, but this may have been confounded by study quality-we noted that trials recruiting people with depression tended to have a higher risk of bias than the trials recruiting patients without depression at onset (data available on request).
The alternative to performing a meta-analysis would have been to perform a narrative review of the 52 studies, but would have been less easy for the reader to interpret. Furthermore, it would have also been susceptible to subjective interpretation of the data by the review authors. The results of the meta-analysis are in broad agreement with older systematic reviews  [57] [58] [59] [60] [61] [62] [63] [64]  that all had much narrower research questions. 8] [59] [60] [61] [62]  Two previous Cochrane reviews of interventions (including SSRIs) to treat and prevent depression after stroke  63, 64  excluded studies that provided usual care without a placebo as the comparator and reported smaller effects than in the current review. The included trials mostly recruited patients aged between 60 and 70 years who were able to consent for themselves.
Performing a simultaneous fit across different jet and b-tag multiplicity bins, including regions dominated by background events, constrains the background contributions, resulting in a more precise measurement of the tt production cross section. The W + jets, Z + jets, and single-top-quark background processes are initially normalized to the expected event yields according to their theoretical cross sections. The QCD and photon + jets normalizations are evaluated as described above individually in each N jet and N tag subsample, for both channels. These background normalizations are the initial values that enter the profile likelihood fit. The cross section measurement is performed by fitting to the data to obtain corrections to these initial values. The W + jets backgrounds are split into W + b jets, W + c jets, and W + light-flavor (LF) sub-samples, with all three components free in the fit.