dataesr/scientific-paragraphs-categorization huggingface dataset:
It depicts linear regression scatter plots of the total 137 Cs deposition based on individual measurements of each country (REM database) in descending order, in terms of the best linear fitting (1 : 1 dependence), as well as the respective calculated biases from the comparison with the observations. Given the large heterogeneity of the samples and the 50 % uncertainty of the emissions, the model results are in very good agreement with observations. The best performance was achieved for 14 countries (ex-Yugoslavia, Spain, Hungary, Italy, Finland, Sweden, Greece, UK, Netherlands, Switzerland, Belgium, Germany, Norway and France) with correlation coefficients between 0. 4 and 0. 9, whereas the estimated average bias was -0. 81 ± 0. 15 for the RG19L run and the Z19L run and lower for the 39 level run (-0.
25 ± 0. 91). This seems very convenient if compared with other model assessments, which have showed biases around 1. 3 for the total deposition of 137 Cs (e. g. Brandt et al. , 2002). Some discrepancies were observed in countries near the Chernobyl site (Romania, Poland, and ex-Czechoslovakia) and, also underestimations in Denmark, Ireland and Austria, while raw data from Ukraine, Belarus and Russia were unavailable from the public database. An important issue that should be stated here, regarding the data of 137 Cs deposition from the REM database, is the fact that these data refer to total deposition of 137 Cs over Europe, which means that the respective deposition from global atmospheric weapon testing, as well as other regional releases (e. g. Sellafield in Great Britain, Mayak in Urals, local releases from fuel fabrication, etc.
Comparison with depositional observations reported by European countries of 137 Cs from the Chernobyl accident in 1986. The model is based on a combination of the aerosol module INCA, the general circulation model LMDz and the global vegetation model ORCHIDEE. The conclusions are based on comparisons with measurements both from the REM database and from the Chernobyl Atlas. Simulations of the Chernobyl accident showed that comprehensive tracer models are powerful tools for estimating the activity concentrations and depositions after accidental scenarios. According to the comparison between model and observations, the most sufficient results were obtained when the highest horizontal resolution of the model was used (Z19L run). Specifically, this model version managed to predict the radioactive contamination in most of the regions alike to Atlas.
46-0. 65). Expected lack of dependence was estimated when surface emissions were assumed. The model is able to simulate and predict the development of the specific activity fields with high efficiency, although rarely underestimated. This is expected taking into account the uncertainties of the source term, the deposition processes and the heterogeneity in the samples. However, statistical tests applied to the respective datasets proved a likely dependence. A general conclusion is that the high resolved grid gives results that track closely the observations, especially in the first days of the emissions. This imposes the essential usage of modelling applications as tools for the decision makers, given that the first days of a nuclear accident are very important for life, in terms of addressing the appropriate evacuation criteria for the radiation protection of the population.
There is a critical need for open data policy after accidental releases. It is a pity that no data from all European countries are present in the public section of the REM database. The paper shows the importance of knowing the emission height of the source in such studies and how much it affects the dispersion and deposition of 137 Cs. However, only speculations can be made about the real altitude where 137 Cs was injected in the atmosphere and therefore, an uncertainty of 50 % is always used in the case of Chernobyl. Nowadays, the existence of several modelling tools is able to predict the overall details of the emission after a NPP accident (e. g. using inverse modelling,  Stohl et al. , 2012). Knowing the exact core inventory by the official authorities or the real emissions during the first days, these dispersion models are able to predict the fate of the radioactive fallout.
Comparison of the total cumulative deposition of 137 Cs from the Chernobyl accident between the REM database and the different model versions used (RG19L(S), RG19L, RG39L and Z19L). Spearman Rank Order and Kendall Tau correlations (R 2 ) between the datasets (N = 4266) for 95 % confidence level (p < 0. 05).
Acknowledgements. The authors would like to appreciate the funding source of the project ( GIS Climat-Environnement-Société , http://www. gisclimat. fr/projet/radioclimfire). This work was granted access to the HPC resources of [CCRT/TGCC/CINES/IDRIS] under the allocation 2012-t2012012201 made by  GENCI  ( Grand Equipement National de Calcul Intensif ). Acknowledgements are also owed to  Patrick Brockman  who initiated us in the secrets of Ferret Analysis Scripting Toolbox and to the  National Oceanic and Atmospheric Administration -Pacific Marine Environmental Laboratory (NOAA-PMEL ) which created and shared the tool in public. The authors would like to acknowledge the  Alfred Wegener Institute for Polar and Marine Research (AWI)  and the  Unidata community , as well, for releasing in public the free software Ocean Data View (ODV) and Integrated Data Viewer (IDV), respectively.
Measurements General characteristics and dietary behaviour assessments. A pre-piloted self-administered questionnaire was used to determine socio-demographic and dietary behaviour such as daily breakfast consumption, snacking practices, eating-out practices, and consumption of carbonated and sweetened beverages. This questionnaire, comprised some structured questions and an unstructured open-ended response, where the participants were asked to provide detailed information about their dietary behaviour such as the frequency and place of consumption of common foods and beverages. Frequency of daily breakfast consumption was assessed using an open-ended questionnaire response format, in which participants were asked to report the actual frequency of their usual breakfast consumption. Data were recorded as a continuous variable in the raw form and were then categorized into two breakfast groups namely, frequent breakfast eaters as $5 times/week and infrequent breakfast eaters as less than 5 times a week.
Objective: Although surgical-site infection (SSI) rates are advocated as a major evaluation criterion, the reproducibility of SSI diagnosis is unknown. We assessed agreement in diagnosing SSI among specialists involved in SSI surveillance in Europe. Methods: Twelve case-vignettes based on suspected SSI were submitted to 100 infection-control physicians (ICPs) and 86 surgeons in 10 European countries. Each participant scored eight randomly-assigned case-vignettes on a secure online relational database. The intra-class correlation coefficient (ICC) was used to assess agreement for SSI diagnosis on a 7-point Likert scale and the kappa coefficient to assess agreement for SSI depth on a three-point scale. Results: Intra-specialty agreement for SSI diagnosis ranged across countries and specialties from 0.
Development of Case-vignettes Case-vignettes allow an assessment of the same cases by ICPs and surgeons involved in diagnosing SSI. We used blinded random assignment of the case-vignettes to ICPs and surgeons to assess agreement regarding SSI diagnosis and depth. In addition, we determined whether providing SSI definitions influenced SSI diagnosis and depth assessment. The case-vignettes were built from SSI surveillance data collected in six surgical units in four French university hospitals. Surgical procedures were selected based on the following criteria: i) preferentially clean or clean-contaminated surgical procedure; ii) presence of a skin incision allowing standard wound surveillance and SSI diagnosis, iii) surgical procedure usually requiring at least 1 week of in-hospital post-operative surveillance, and iv) sufficiently high SSI incidence to ensure the collection of a large number of suspected cases within a short period.
Study Design and Data The 20 vignettes were assigned at random to allow assessments of agreement among (i) participants in the same speciality in the same country; (ii) participants in the same speciality in different countries; and (iii) participants in different specialities in the same country. Each of the 20 vignettes was to be scored four times by different ICPs and surgeons in all 10 countries. Then, four ICPs and four surgeons taken at random in each country read the SSI definitions and repeated the scoring of one vignette. Scores were assigned using a seven-point Likert scale ranging from ''SSI certainly absent'' (score 1) to ''SSI certainly present'' (score 7)  [14]. When the score was between 4 and 7, the participant scored SSI depth on a 3-point scale (1, superficial SSI; 2, depth unclear; and 3, deep or organ/space-related SSI).
We simplified the depth assessment by classifying deep and organ/space-related SSIs in the same group, as both SSI categories have the same severe consequences in terms of mortality, morbidity, and hospital stay prolongation. A secure online relational database was established for data collection. Each participant had a personal login and password  [15]. Patient data were presented chronologically, and the scores assigned before reading the SSI definition could not be changed. Before scoring the vignettes, each participant provided the following information: age, gender, type of hospital, and time working in the current job.
Statistical Analysis We estimated the number of vignettes and participants needed to assess agreement within specialties based on the precision of the intra-class correlation coefficient (ICC)  [16]  and on feasibility considerations (number of participants available in each specialty, maximal time needed for scoring). With 20 vignettes each scored four times and an expected ICC of about 0. 60, half the exact 95 per cent confidence interval (95%CI), i. e. , precision, would be 0. 29. Data were described as mean+/ 2SD, median (interquartile range), or percentage. Agreement was assessed before and after reading the SSI definition without distinguishing specialties or countries. To evaluate intra-and inter-specialty agreements for SSI diagnosis based on 1-7 Likert scale scores, we computed the ICC with the 95%CIs.
Finally, a study based on the same methodology as the one reported here assessed agreement among a large number of healthcare workers in France  [11]. Agreement regarding SSI diagnosis and depth varied across specialties and across individuals within each specialty. Reading the SSI definition produced small improvements in agreement about SSI diagnosis and depth. Our study further supports the existence of considerable uncertainty regarding SSI detection at the European level. Our results are probably reliable, as we placed the participants in unbiased conditions by asking them to score the same casevignettes through an Internet database. This method ensured that the participants were not influenced by factors such as perceived SSI risk in a particular unit or patient. Considering such factors would probably have increased disagreement among participants.
Our results further support the need for a multidisciplinary approach to SSI surveillance  [27]. Our data from 10 European countries consistently showed differences in agreement in each country, suggesting that our results may be also relevant to other countries. Our study has several limitations. First, the vignettes were scored for the presence or absence of SSI by each participant working alone. SSI is often a difficult diagnosis that is typically made after discussion among surgeons and ICPs. Thus, SSI surveillance aims not only to obtain accurate SSI rates, but also to enhance teamwork between surgical and infection-control teams in order to ensure the implementation of effective preventive strategies. Our results indicate that surveillance should not be performed by individuals in a single specialty  [27].
Second, the vignettes were scored via an online database. The vignettes were built from real cases, and the diagnosis of SSI may have been easier for surgeons or ICPs who had had direct contact with the patients. Third, the study was not designed to assess the accuracy of SSI diagnosis. Instead, we focused on agreement among ICPs and surgeons. We were therefore unable to determine which participants made the correct diagnosis, as illustrated in Table  1  showing SSI diagnosis score differences of up to 6 points between Table  1. Distribution of scores assigned by infection control physicians and surgeons before reading the definitions of surgical site infections, on a 7-point Likert scale, in each of the ten European countries. two participants from the same specialty. Fourth, we selected cases of suspected SSI to assess agreement about the diagnosis of SSI.
Acknowledgments We thank  P Nataf ,  MD, Bichat-Claude Bernard University Hospital, Paris; Philippe Despins, MD, University Hospital, Nantes; Jean-Pierre Marmuse, MD, Bichat-Claude Bernard University Hospital; Philippe Massin, MD, Bichat-Claude Bernard University Hospital; Baptiste Roux, PharmD, Fast4 Company, Paris , and all the European 186 participants who scored the vignettes:  Finland Infection control physicians. Outi Lyytikainen ,  Helsinki University Central Hospital ;  Bodil Eriksen-Neuman , Vaasa central hospital;  Kaisa Huotari , Peijas hospital;  Maija Liisa Rummukainen , Jyva ¨skyla ¨central hospital; Veli-Jukka Anttila ,  Helsinki University Central Hospital ;  Peter Klemets , Porvoo hospital;  Pekka Suomalainen ,  South Carelia Central Hospital ;  Nabil Karah ,  University Hospital of North -Norway ;  Arvola Pertti ,  Tampere University Hospital ;  Kirsi Skogberg ,  Jorvi hospital.
Their estimates support the conclusion that STW schemes had an economically important impact, with the largest impacts of STW on employment in Germany and Japan among the countries considered. However, the positive impact of STW was limited to workers with permanent contracts, thereby further increasing labour market segmentation between workers in regular jobs and workers in temporary jobs. The estimated jobs impact was smaller than the potential number of jobs saved as implied by the full-time equivalent number of participants in STW. This suggests that STW schemes end up supporting some jobs that would have been maintained in the absence of the subsidy. Subsequent studies by Boeri and Brucker (2011) and Cahuc and Carcillo  (2011), who make use of largely the same dataset but employed somewhat different estimation strategies, reached similar conclusions  4.
This paper also analyses the quantitative effects of STW programmes on employment and hours by exploiting the country and time variation in STW take-up rates. It provides an update on the initial assessment by Hijzen and Venn (2010) by extending the time and country coverage of the dataset from 2004 Q1-2009 Q3 to 2004 Q1-2010 Q4  and from 19 to 23 countries 5. The use of more recent data is important as it allows analysing the impact of STW not just during the crisis, but also during the early phase of the recovery. The econometric analysis takes account of differences in institutional settings across countries that might affect the relationship between labour market outcomes and output and addresses the potential endogeneity of STW take-up with respect to labour market conditions using the age of the programme as an instrument.
Data For the purposes of this paper, the dataset constructed by Hijzen and Venn (2010) was extended to cover a larger number of countries and to include a larger part of the recent crisis period. This increased the number of countries covered from 19 to 23 and the data period from 2004 Q1-2009Q4 to 2004 Q1-2010 Q4. The use of five additional quarters over a period in which the use of STW was relatively high is considered to be an important contribution of this paper. Of the 23 countries that are included in the analysis, 13 countries operated a scheme already before the start of the global financial crisis, five introduced a new scheme in response to the crisis and five countries never operated a STW scheme. Since national data on the use of STW are not readily comparable across countries significant efforts were made to enhance cross-country comparability in terms of the kind of programmes that are covered and the way take-up is recorded across countries.
Take-up is measured in this paper as the number of participants across months as a share of dependent employment. Note that it would have been preferable to measure take-up in terms of full-time equivalents (FTE). However, this requires information on the reduction in working time among STW participants which is not available for the majority of countries. For further details on the construction of this dataset, see Hijzen and Venn (2010) and the Additional file 1. Table  3  provides an overview of the use of STW before, during and after the crisis for countries for which comparable data are available 8. At the onset of the crisis in the fourth quarter of 2007, take-up of STW was negligible in most countries. An important exception is Belgium where take-up amounted to about 3% of dependent employment.
Take-up increased most strongly in response to the crisis in Belgium, where take-up rose by more than one percentage point for each percent decline in GDP. The responsiveness of take-up was also strong in countries such as Germany and Japan, where it increased by more than half a percentage point for each percent decline in GDP. By contrast, responsiveness was very low in Denmark, Ireland, Poland and Portugal. During the initial recovery, the responsiveness of take-up to improvements in GDP was similar to that during the crisis in most countries. This suggests that concerns about the difficulty of scaling down STW schemes in the recovery may not be very important. The core database on labour market outcomes and output is derived from Eurostat's Quarterly National Accounts, the European Labour Force Survey and various national sources for the non-European countries.
Acknowledgements This paper was prepared as part of a broader OECD project on «The role of policies for labour market resilience». The views expressed in this paper are those of the authors and cannot be attributed to the OECD or its member states. The authors would like to thank  Mark Keese ,  Paul Swaim ,  Danielle Venn  and  Josef Zweimuller  and one anonymous referee for helpful comments and suggestions. Financial support from the  European Commission  under grant number  VS2010/0617 -SI576449  is gratefully acknowledged. All remaining errors are our own. For any queries in relation to this paper the authors can be contacted on the following email addresses: alexander. hijzen@oecd. org[ Alexander Hijzen ] and sebastien. martin@oecd. org[ Sebastien Martin ]. The Responsible Editor is:  Juan Francisco Jimeno.
Formally, the main effects model is written as where, for fetus i at age t, y it is the observed number of counts of a specific touch type (upper, side, lower or perioral) s it is the accumulated length of coded scan(usually 600 seconds), u i is the individual fetus random effect, which is assumed to be normally distributed with mean 0 and variance σ 2 , and β 0 and β 1 are unknown parameters representing the intercept and age slope. The model accounts both for the skewness of the data and also allows for individual variability between fetuses in their propensity to the event, accounting for the base activity level of each fetus. The mixed effects modelling was carried out using the glmer function in the lme4 package of the statistical package R. (Bates and Maechler, 2011)   We primarily focused on two sequences: mouth movement before touch of the face, and mouth movement after touch of the face, which were types 2 and 4 in the coding scheme, and which corresponded to the anticipatory and reactive oral behaviour.
For each of these two sequences we used a binomial logistic mixed effects analysis using glmer  (Pinheiro and Bates, 2000)  to assess developmental change in the relative proportion of these sequences by gestational age, taking the response variable to be the number of face touch events F it following a specific sequence out of the total number of face touch events N it in the accumulated scan period. Using the previous notation, the main effects binomial logistic mixed effects model can be written as where p it is the proportion of touch-mouth movements of a specific type out of all touchmouth movements for fetus i at age t. As in the earlier analysis, we took account of the repeated measures nature of the data by including an individual random effects term for the fetus. As before, significance was assessed by likelihood ratio tests.
5406 Å). The samples were gently crushed before being introduced into 0. 5 mm inner diameter Lindemann glass capillaries. The high resolution XRPD patterns were indexed using the peak picking option of the software package of Materials Studio Modeling 5. 5. 7 Potential solutions for unit cell parameters and the space group were found using the X cell algorithm. 8 The unit cell dimensions were refined using a Pawley profile fitting procedure. 9 This procedure was also followed to determine unit cell parameters as a function of temperature. For structure determination, the Dreiding force field procedure 10 was used to get the distances, angles, and torsions in the molecule. Specific Volumes in the Liquid State. The specific volumes as a function of temperature were determined using a method that has already been described.
NMR measurements 1 H and 13 C NMR spectra were recorded on an Avance 300, Avance 400 and Avance 600 spectrometers (Bruker) operating at 300. 13, 400. Lab nitrogen generator (Germany) and Finnigan X Calibur 1. 3 software for data collecting and processing. Acetonitrile (Panreac, HPLC gradient grade) was used as the mobile phase. Dilute MeCN solutions of compounds 1b, 2b, 3a and 5 were studied in the presence of 1 eq. of HClO 4. The isocratic elution was maintained at a flow rate of 50 mL • min À1 without column. The effluent from LC was passed directly into the electrospray ion source without a split. Positive electrospray ionization was achieved using a ionization voltage at 3 kV with temperature at 200 C. Isotope patterns were calculated with Molecular Weight Calculator, Version 6. 37 (Matthew Monroe).
Equilibrium constant determination Complex formation of dyes 1b, 2b and 3a with Ca 2þ in aceto nitrile at 20 AE 1 C was studied by spectrofluorometric (for 1b, 2b and 3a) and spectrophotometric (for 3a) titration  [43, 44]. The ratio of dye to Ca 2þ was varied by adding aliquots of a solution of Ca(ClO 4 ) 2 of known concentration to a solution of ligands 1b, 2b or 3a of known concentration. The fluorescence (absorption) spec trum of each solution was recorded, and the stability constants of the complexes were determined using the SPECFIT/32 program (Spectrum Software Associates, West Marlborough, MA). The following equilibria were considered in the fitting (equations (  1 ) and (  2 ), L 1b, 2b or 3a; M Ca 2þ ): In doing so, it was found that the experimental data corresponded to the theoretical ones if only the equation (  2 ) was taken into account and the formation of the complexes with composition of 2:1 was not observed.