dataesr/scientific-paragraphs-categorization huggingface dataset:
In the subsequent phase 3 study (NCT014339768), adults with chronic ITP were randomized to 20 mg of avatrombopag (32 subjects) or placebo (17 subjects) for 24 weeks; the median cumulative number of weeks with a platelet response. 50 3 10 9 /L was 12. 4 for avatrombopag and 0. 0 for placebo (P ,. 0001). 160 Platelet responses were rapid (66% by day 8) and were maintained for. 12 months in an extension study. Efficacy in children has not yet been established. Dose changes are made by increasing the 20-mg starting dose to 40 mg/d or reducing dose frequency. Rituximab. Long-term data for rituximab have become available since 2010 (supplemental Table  3 ). A meta-analysis of 5 trials revealed that complete responses were more likely with rituximab than with standard of care; thus, rituximab can improve platelet count responses at 6 months in ITP patients.
205 Other immunosuppressive agents, including vinca alkaloids and MMF, have been associated with teratogenicity and should not be used. 229, 230 O-RAS AND RHTPO. TPO-RAs may be considered in exceptional circumstances, ideally only in the third trimester near delivery, although tolerability and toxic effects on the fetus have not been rigorously evaluated. An unpublished safety database (Amgen, Thousand Oaks, CA) exists for. 25 women treated with romiplostim during pregnancy without untoward effects. All available published data are from 11 available case reports, and all show a platelet response. [119] [120] [121] 141, [231] [232] [233] [234] [235] [236] [237]  Romiplostim and eltrombopag may cross the placenta and could cause potential harm to the fetus. 119, 120, 227  In the 11 cases reviewed here, there were no safety concerns.
[119] [120] [121] 141, [231] [232] [233] [234] [235] [236] [237]  TPO-RAs are not recommended except in very serious cases; if used, romiplostim may be preferred given its fewer off-target effects (lack of hepatic toxicity and iron chelation) and limited safety database. rhTPO is likely to cross the placenta and was effective in increasing the platelet count in 23 of 31 pregnant women who were unresponsive to steroids without untoward effects on the fetus. 227 hTPO is only available in China. SPLENECTOMY. Splenectomy is required in very rare cases. It is best performed in the second trimester, by laparoscopy if possible and with vaccination according to national guidelines. VENOUS THROMBOEMBOLISM. In addition to the known prothrombotic state of pregnancy, some women may have risk factors (eg, anticardiolipin antibody syndrome) requiring venous thromboembolism prophylaxis (evidence level IV).
277 Limited toxicity (asthenia, vomiting, and skin effects in 1 patient each) was observed. Doses were not specified. FOSTAMATINIB. Fostamatinib is not licensed for the treatment of children with ITP and there are no data on its use in children. As for splenectomy in adults, longterm antibiotic prophylaxis and repeat vaccinations should be given according to national guidelines. The risk of infection in later life should be considered for polysaccharide-encapsulated organisms (eg, pneumococcus), as well as, for example, if the child later works in a school or as a veterinarian or travels to areas with endemic and/or epidemic Babesia, malaria, or dengue. The patient should be advised of the risks of postsplenectomy sepsis and thrombosis (and their appropriate management), to obtain yearly blood counts, and to keep vaccinations up to date.
1. Changes in agricultural practices have reshaped agricultural landscapes and triggered a drastic decline in spatial and temporal heterogeneity leading to changes in habitat quality and food resources for birds. However, the precise relationships between landscape changes, spatial and temporal heterogeneity, and habitat preferences in response to those changes remain poorly known. 2. We investigated patterns of habitat selection and causes for changes over the years 1997-2017 for the gray partridge Perdix perdix, an iconic farmland species which has experienced a severe decline since the 1950s. Using a long-term  dataset collected over 435 km 2 , we modeled relationships between annual land-cover maps and partridge sightings over 17 5-year moving windows, assessing the effects of landscape metrics, the strength of the relationships, and the dynamics of habitat suitability.
| Partridge data In total, there were 989 sightings of gray partridges in three distinct datasets covering 1997 to 2017 (Figure  1e ). The largest dataset (90% of total data) comprised opportunistic sightings either when ornithologists working in the LTSER were studying other bird species  (Bretagnolle et al. , 2018b)  and saw partridges, or during the landcover surveys carried out twice a year systematically over the entire study site. Each year, opportunistic sightings were reported daily between late March and late July (see Figure  1f ). In this dataset, the observation effort is not standardized and reporting is not systematic. In the second dataset, with 5% of total sightings, observation pressure is standardized as this dataset collates the results from systematic bird point counts since 1995 (see Brodier, Augiron, Cornulier, & Bretagnolle, 2014 for methods).
The number of point counts has increased from 160  (1995) (1996) (1997) (1998) (1999) (2000) (2001) (2002) (2003) (2004) (2005) (2006) (2007) (2008)  to about 450 since 2009, spread over the whole of the LTSER  (Bretagnolle et al. , 2018a). Count duration was either 5 or 10 min (see  Brodier et al. , 2014)  between 7:00 a. m. and 11:00 a. m. , without call playback. The third dataset, with 5% of total sightings, collates the results of standardized call playback counts specifically for both gray partridges and red-legged partridges Alectoris rufa L. , in March-April 2016 (140 sites) and 2017 (275 sites), either in the morning (from 7:00 a. m. to 11:00 a. m. ) or evening (from 5:00 p. m. to 9:00 p. m. ). Count duration was 10 min. In all three datasets, the location of the birds, the date and time, and the number of birds were recorded.
(2013) , resulting in nine uncorrelated explanatory variables for the dataset with 5-year moving windows (Supporting Information Table  S2 ); however, the rape density was excluded for modeling the dataset aggregated over the whole period (Supporting Information Appendix S2). Secondly, to se- Supporting Information Appendix S3 gives the methods and numbers of pseudo-absences). Generalized linear models (GLM, binomial response variable, logit link) were selected for the rest of the analyses as they showed a high true skill statistic (TSS = sensitivity + specificity-1;  Allouche, Tsoar, & Kadmon, 2006)  and high sensitivity (Supporting Information Appendix S3). For each period, the models were calibrated on a random subset of 70% of the presence/ pseudo-absence data and then cross-validated using the remaining 30%: This cross-validation was performed three times  (Thuiller, Lafourcade, & Araujo, 2009).
| RE SULTS First, we investigated the key landscape metrics explaining the gray partridge occurrence over the study period , and their effect on gray partridge occurrence. Of the nine candidate explanatory variables, cereal density, distance to woodlands, and distance to the nearest building were the metrics with the highest contributions when fitting to the aggregated dataset for 1997-2017 (contributions >0. 1, Figure  2 ; Supporting Information Table  S3 ). Most response curves were quadratic: Moderate values of cereal density (25%-65%) and distance to woodlands (300-1,310 m) were preferred by partridges while extreme values were avoided (Figure  3 ). Partridges also preferred being more than 125 m from a building (Figure  3 ). For the contiguous 5-year periods, most of the metrics had similar contributions to those for the whole dataset (Figure  2 ; compare black and gray boxes), though the contribution of the distance to buildings was below the 0.
lect the modeling technique (Elith, Ferrier, Huettmann, & Leathwick, 2005) , we used version 2. 0 BIOMOD multimodel platform (Thuiller, Lafourcade, Engler, & Araújo, 2009)  implemented in R. This can (a) use heterogeneous data from different counting methods (Farashi & Shariati, 2017; Jackson, Gergel, & Martin, 2015)  and (b) compare the most frequently used modeling techniques (Barbet-Massin et al. , 2012; Monnet, Hingrat, & Jiguet, 2015 ; see Supporting Information Appendix S3). As these modeling techniques require presence and absence data but as our dataset did not contain true absences (the absence of a sighting does not imply the absence of partridges), we generated 100 replicates of 1,000 random pseudo-absences to obtain reliable confidence intervals (Barbet-Massin et al. , 2012;.
Location of the Long-Term Socio-Ecological Research (LTSER) platform "Zone Atelier Plaine & Val de Sèvre" in France. Black dots inside the LTSER represent gray partridge sightings recorded from 1997 to 2017. (b)Density of gray partridge opportunistic sightings (see Methods) from 1997 to 2017. (c)Density of cereals (time varying) and woodland (stable) from 1997 to 2017. Solid lines represent significant trends from generalized least squares (GLS) model with autoregressive moving average (see TableS1for statistical details). (d) Average size of crop patches from 1997 to 2017 on the LTSER. Solid lines represent significant trends from GLS models with autoregressive moving average. (e) Gray partridge sightings each year between 1997 and 2017 for each dataset (black: opportunistic sightings; dark gray: point counts; light gray: point counts with call playback).
Acknowledgements Support for RMBY and PLR provided by  UK STFC  Grants  ST/F003145/1 ,  ST/I001948/1 , and  ST/K00106X/1. Support for PLR provided by  UK STFC  Grant  ST/N00082X/1. Part of this work was completed during a visit to the  Kavli Institute for Theoretical Physics at UC Santa Barbara. This work was supported in part by the  US National Science Foundation  under Grant No. NSF   PHY-1125915. The authors would like to thank  Jeremy Yates  for his patience while we ran the simulations, and two anonymous reviewers whose extensive comments improved the paper significantly. Several figures used David Fanning's  Coyote IDL Program Library. This work used the DiRAC Data Centric system at Durham University, operated by the  Institute for Computational Cosmology  on behalf of the  STFC DiRAC HPC Facility  (www.
dirac. ac. uk). This equipment was funded by  BIS National E-infrastructure capital  grant  ST/K00042X/1 ,  STFC  capital grant  ST/K00087X/1 ,  DiRAC Operations  grant  ST/K003267/1 , and  Durham University. DiRAC is part of the  UK National E-Infrastructure. The authors acknowledge the use of the  University of Oxford Advanced Research Computing (ARC)  facility in carrying out this work. http://dx. doi. org/10. 5281/zenodo. 22558.
Molecular procedures For each isolate, DNA was extracted from 50 mg of fresh fungal material scraped from a fiveday culture grown on PDA medium in the dark at 18°C. DNA was extracted in an automated system (Biorobot 3000, Qiagen), with the appropriate DNeasy kit (Qiagen). All 794 isolates were genotyped for 11 single sequence repeat (SSR) markers  (St2, St3A, St3C, St4, St5, St6, St7, St9, St10, St11 and St13; Gautier et al. , 2014)  in two multiplex PCRs, ensuring that amplicons with similar magnitudes of sizes bore distinct fluorochromes. Automatic allele recognition and annotation, based on binning analysis, were performed, to ensure the homogeneous naming of the alleles at each locus (Beckmann Coulter CEQ 8000 software;  Gautier et al. , 2014). In total, 585 isolates were successfully genotyped with the 11 SSRs.
The remaining isolates (27% of the total dataset) provided no amplicons for only one or two markers, even spread between the SSRs; they were treated as missing values in subsequent analysis. Because the proportion of identical genotypes (clones) in the subpopulations was very low, genetic analyses were performed without clone correction. Mating type was determined by PCR for 739 isolates  (Waalwijk et al. , 2002). Eight per cent of the isolates were not genotyped for mating type due to technical failures or the contamination of PCR assays.
, 2005)  with STRUCTURE HARVESTER  (Earl & von Holdt, 2012; Belkhir et al. , 1996 Belkhir et al. , -2004)). Three STRUCTURE analyses were carried out independently, to investigate three factors thought likely to affect the structure of the dataset: 1) collection time, during one of three or four epidemic periods during the three cropping seasons, 2) isolate origin (plot with or without debris), and 3) spore type giving rise to the isolate (ascospore or pycnidiospore). A first set of analyses was carried out to assess, for each cropping season separately  (2009-2010, 2010-2011, 2011-2012) , the effect of epidemic period on the structure of the pycnidiospore-derived subpopulations collected from the plot with debris (P02-P03-P04, P09-P10-P11-P12, and P17-P18-P19-P21, respectively). An analysis grouping together the three cropping season datasets plus the data for P29  (2012) (2013) , was also carried out to assess the effect of year.
A second set of analyses was carried out to assess, for each cropping season separately (and then overall), the effect of isolate origin (plot with or without debris) on the structure of pycnidiosporederived subpopulations (P02-P03-P04-P05-P06-P07, P09-P10-P11-P12-P13-P14-P15-P16, and P17-P18-P19-P21-P23-P24-P25-P27, respectively). Finally, we also analyzed the structuring effect of spore type giving rise to the isolate on the complete dataset (combining P01-P08-P28).
Population partitioning Bar plots of the three independent STRUCTURE analyses carried out to evaluate population subdivision with the admixture model and without prior assumptions about time (year and epidemic period), isolate origin (plot with or without debris), and spore type giving rise to the isolate (ascospore or pycnidiospore), are provided in Fig. 2. The three cropping seasons  (2009-2010, 2010-2011, 2011-2012)  and the adjacent epidemic periods (end of the 2008-2009 cropping season, P01; start of the 2012-2013 cropping season, P29) are presented together. In the first set of analyses, pycnidiospore-derived subpopulations from the D+ plot were analyzed for each cropping system separately (Table  S1 ) and then for all cropping seasons together (Table  S1 ; Fig. 2A ). The rate of change in the log probability of data between successive K values (∆K) had a mode at K = 3 for the 2009-2010 season, at K = 2 for the 2010-2011 and 2011-2012 cropping seasons, and at K=7 for the overall dataset.
Regardless of the epidemic period considered in each cropping season (year), it was difficult to assign individuals to any one of these K clusters (none of the individuals were assigned to one cluster with a posterior probability > 0. 5). Instead, the posterior probability was split equally between the seven clusters (0. 1426 < overall mean P < 0. 1431), suggesting that neither the epidemic period nor the cropping season had any major effect on the structure of pycnidiospore-derived subpopulations. In the second set of analyses, pycnidiospore-derived subpopulations from the D+ and D0 plots were analyzed for each cropping season separately and then for all cropping seasons together ( Table  S1 ; Fig. 2B ). Modes were observed at K = 3 for the 2009-2010 season and at K = 2 for the 2010-2011, and 2011-2012 cropping seasons and for the overall dataset.
The posterior probability of individuals was equally distributed between the two clusters (0. 498 < overall mean P < 0. 502). Only 0. 15 % of the individuals were assigned to one cluster with a posterior probability greater than 0. 6, whatever the epidemic period, again suggesting that isolate origin (D+ or D0 plots) had little effect on the structure of the dataset. Finally, after the addition of ascospore-derived subpopulations to the previous dataset in the third STRUCTURE analysis (Table  S1 ; Fig. 2C ), a mode at K = 3 and evenly distributed posterior probabilities (0. 332 < overall mean P < 0. 334) were again observed. We can therefore conclude that spore type, giving rise to the isolate, had no measurable effect on population structure. As no population structure was detected without prior assumptions, we explored the dataset with the a priori subpopulations defined in Fig.
Acknowledgments The authors thank  Angélique Gautier  ( INRA BIOGER ) and  Christophe Montagnier  ( INRA Experimental Unit, Thiverval-Grignon ) for technical assistance. This study was supported by a grant from the  European Union Seventh Framework Program  (Grant Agreement no. 261752 ,  PLANTFOODSEC  project) and a grant from the  European Union Horizon   Framework 2020 Program  (Grant Agreement no. 634179 ,  EMPHASIS  project).
ACKNOWLEDGEMENT This study was supported by the "Région Nouvelle Aquitaine", FEDER Funds from Plan Loire IV and Agence de l'Eau Loire Bretagne. 4. Macronutrient concentrations (mg kg -1 DW) in wood and branches of the 14 poplar genotypes from the Pierrelaye site. Data represent the mean (± SE), resulting from samples taken at three different heights as described in the material and method section (see also Fig. S1 ). Levels of significance are indicated by asterisks : ***p < 0. 001; **p < 0. 01. Different letters represent significant differences for element content between the genotypes (one-way ANOVA, Tukey-Kramer HSD test). ns : not significant. Table  5. Trace element concentrations (mg kg -1 DW) in wood of the 14 poplar genotypes from the Pierrelaye site. Data represent the mean (± SE), resulting from samples taken at three different heights as described in the material and method section (see also Fig.
The remaining sixteen communications included a multitude of different complaints made by Canadian citizens regarding the alleged violation of, for instance, the principle of non-discrimination or the right to a fair trial. These citizen-lodged complaints were excluded from review. Through further analysis, we found that fourteen communications were discontinued, 30  three were duplicates, and two were against another country and were erroneously filed in the UN database as complaints against Canada. Consequently, these nineteen communications were excluded from further review. This left us with a dataset of sixty-two communications in total. We analyzed all sixty-two communications using a table we created that organized the communications according to the following criteria: the treaty the complaint was made under; the articles allegedly violated; legal representation; and admissibility.
These remedies include: (1) appeals to the Federal Court for judicial review; (2) humanitarian and compassionate (H&C) applications; (3) pre-removal risk assessments; and (4) administrative deferrals of removal. However, our analysis of committee decisions reveals issues with the effectiveness of these remedies. Almost all of the cases in our dataset concerned alleged rights violations engendered by deportation orders. The complainants are either asylum seekers whose refugee claims were refused by the IRB, and their family members, or undocumented migrants or permanent residents who have been found inadmissible for "serious criminality. " In all of these cases, non-citizens who receive a negative decision from an immigration officer or the IRB 48 can apply to the Federal Court for judicial review of this decision.
Canada's Non-Compliance with the UN Committee Decisions This brings us to a discussion of Canada's compliance with the UN committee decisions. The cases we have examined point to two issues in this regard. The first one pertains to interim measures whereby the HRC and the CAT call upon a state party to suspend any action while the communication is being considered by either committee. Canadian authorities have not always complied with the interim measures asking Canada not to deport a complainant to their country of origin. 117 An extension of the non-refoulement principle, interim measures are essential to the effectiveness of the UN individual complaints mechanisms. We identified three cases in our dataset where Canada did not act in accordance with an interim measure, sending the complainant to possible persecution or another life-threatening situation.
40  However, this number may increase, as communications analyzed during the October-November 2018 sessions of the CAT and the HRC have not been published on the UN Treaty Search Database as of February 2019. https://www. cambridge. org/core/terms. https://doi. org/10. 1017/cyl. 2019. 13 Downloaded from https://www. cambridge. org/core, IP address: 51. 91. 132. 236, on subject to the Cambridge Core terms of use, available at.
Based on indirect experimental evidence and different lines of argumentation, some predict TER increases with increasing temperature  (10, 13)  while others predict the opposite  (12, 14). This absence of consensus clearly hinders our ability to predict the consequences of climate changes on matter and energy flows within ecosystems  (5). Here, we address this issue by combining a TER model and a unique dataset of Daphnia magna TERC:P measured along a wide range of temperatures. Our model differs from previous approaches  (15)  in that all processes involved in the determination of the TER are hypothesized to be temperature-sensitive (Fig. 1 ). We assume that certain physiological processes such as ingestion rate and assimilation efficiency respond to temperature with typical Thermal Performance Curves (TPC)  (16, 17)  while others such as metabolic rate display an exponential response.
Both elements are only mildly radioactive, and their handling allows to obtain heat capacity data of the highest quality and accuracy with the instrument employed, and radiation-induced disorder is negligible. Secondly, we measured the low-temperature heat capacity of uranium dioxide doped with the highly radioactive Pu isotope (half life 87. 7 years) as a function of aging time to assess the effect of alpha decay and the concomitant lattice disorder in the fluorite structure.
Each individual peak has been been integrated using Origin Software (version 8. 1) and the results are shown in Table  1  and compared to similar analysis done for literature data of UO 2. The expected entropy of the triplet Γ 5 ground state of the  3 H 4 multiplet of U 4+ in pure UO 2 is R ln(3) = 9. 1 J • K -1 • mol -1. Huntzicker and Westrum 3 reported 8. 4 J • K -1 • mol -1 for the total anomaly and ~6. 3 J • K -1 • mol -1 at the Néel temperature. The latter becomes 7. 24 J • K -1 • mol -1 using the same analysis as for our measurements. The entropy values associated to the magnetic transition of (U 0. 95 Th 0. 05 )O 2 , (U 0. 91 Th 0. 09 )O 2 and (U 0. 88 Th 0. 12 )O 2 at the Néel temperature were found to be 4. 64, 4. 26 and 3. 74 J • K -1 • mol -1 , respectively (corresponding to 4.
This is close to the value 42% found by Comly  11  and about 45% Th by Hinatsu and Fujino 12 by magnetic susceptibility measurements. (U, 238 Pu)O 2 : effect of radiation damage. The results of the series low-temperature heat capacity measurements of the (U, 238 Pu)O 2 sample in the temperature interval from 5 to 50 K for different storage times after annealing of the sample are presented in Fig. 5 , together with the low-temperature heat capacity data of the end-members. and Magnani et al. 5. The excess heat capacity peak of the measured sample has been derived as a function of the accumulated α dose up to 277 days. Using again the Origin Software (version 8. 1), the peak evaluation has been performed and the related thermodynamic data are given in Table  2. Applying the estimated heat capacity of UO 2 based on the    same corrections made by Huntzicker and Westrum 3 the excess entropies at the Néel temperature of the magnetic transition are found to decrease with time (Table  2 ).