dataesr/scientific-paragraphs-categorization huggingface dataset:
As such, it can be easily observed that the current distribution of the S. suis serotype 2 STs greatly varies throughout the world, though data have only been available for a little over a decade, from only a few countries and mostly only for the serotype 2. Nevertheless, there is an important issue regarding the serotypes attributed to the different strains typed by MLST. Many of the studies, including most of the strains from the S. suis MLST Database, do not specify the method used for serotyping. In other cases, such as the study of King et al. ,  51  the authors did not necessarily confirm the serotypes of the strains used. As explained above, the use of PCR for serotyping of serotypes 2 and 14 strains is an inappropriate method, especially for clinical pig cases, even though cases serotyped using this    method were still considered for this part of the review.
Serotype 5, 16 and 24 human cases of infection were identified as ST181, ST106 and ST221, respectively. 133, 146 nterestingly, no data have yet been published on the possible presence of these three newly identified STs for human isolates in diseased pigs. a Serotypes were identified by co-agglutination (or an equivalent method using reference antisera) or by PCR. However, it is impossible to distinguish between serotypes 1 and 14 and serotypes 1/2 and 2 by PCR, so serotypes remain to be confirmed. b  Gottschalk M, unpublished data (2014). Up-to-date distribution of Streptococcus suis G Goyette-Desjardins et al.
Though the MLST data currently available for S. suis strains isolated from both diseased pigs and human cases of infection come from different countries throughout the world, there remains a long way to go before a complete picture of the current situation of S. suis can be obtained. Yet, this picture is necessary as it could hint at both dominating and emerging STs and could possibly help identify epidemic strains quickly and categorize the zoonotic potential of specific STs. With the dissemination of information about STs across the internet, particularly thanks to the S. suis MLST Database, it would be important to set up a complete and reliable serotype identification of the different strains added. Although this database is a great tool that facilitates the sharing of knowledge, the lack of many details pertaining to each strain such as the serotype, method used for serotyping, health status of the host, and host itself, limit the full potential of the STs as was demonstrated in this review where many strains could not be included.
ACKNOWLEDGMENTS The research carried out on Streptococcus suis in our laboratories is funded by the  Natural Sciences and Engineering Research Council of Canada  (grant # 154280  to  Marcelo Gottschalk  and  342150-07  to  Mariela Segura ) and by  China-Canada Joint Health Research Initiative  financed by  Canadian Institutes of Health Research  and the  National Natural Science Foundation of China  ( 812111251 ). This publication made use of the  Multilocus Sequence Typing Website  (http://www. mlst. net)at  Imperial College London  developed by  David Aanensen  and funded by the  Welcome Trust.
USE OF THE PHOSPHO-ERK ASSAY TO MONITOR EGFR-MEDIATED ERK1/2 ACTIVATION First, we quantified the total ERK1/2 levels in various cell lines using our HTRF-based assay and as shown in Figure  2A. Positive and specific HTRF signals reflecting the total ERK1/2 were measured and the signal varied with the cell line used. Then, we assessed the kinetics of ERK1/2 phosphorylation mediated by EGFR endogenously expressed in HEK293 cells. As shown in Figure  2B , stimulation with 100 nM EGF showed a maximal level of ERK1/2 phosphorylation at 2-5 min of stimulation, which had largely disappeared after 10 min. The transient EGF-induced ERK1/2 activation is consistent with many previous studies using different cell lines  (34) (35) (36). Consequently, all the data on EGFR presented below have been generated at 5 min of stimulation with EGF.
, 2012; Sun et al. , 2003)  and some candidate genes have been proposed such as PLAGL1  (Abdollahi et al. , 2003) , GRM1, SOD2  (Shridhar et al. , 1999) , SASH1  (Zeller et al. , 2003)  or Parkin  (Denison et al. , 2003). However, few studies so far have aimed to define specific DNA copy number markers that may have clinical relevance in predicting outcome in ovarian cancer  (Baumbusch et al. , 2013; Bruchim et al. , 2009; Engler et al. , 2012; Wang et al. , 2012; Yamamoto et al. , 2009). Most studies that have focused on the assessment of specific alterations have been limited by the absence of independent copy number replication datasets  (Bruchim et al. , 2009; Yamamoto et al. , 2009). Other studies including independent validation series have been mainly focused on describing general features (ie.
genomic instability or LOH profiles) which are more difficult to implement in the clinic than distinct individual changes  (Baumbusch et al. , 2013; Wang et al. , 2012). In our study, we used a series of familial and sporadic HGSOCs, whose DNA copy number profiles had been previously characterized  (Kamieniak et al. , 2013). We performed a DNA copy number-based unsupervised clustering that revealed two main groups of tumors of distinct immunohistopathological features and clinical outcome. We further identified a single region of deletion at 6q24. 2e26 that differentiated both clusters and that was subsequently found to be associated with longer survival. We used different independent datasets to validate the prognostic value of this deletion at the DNA copy number (N ¼ 447) and at the expression level (N ¼ 799) and finally propose some candidate genes that may potentially explain the association.
Definition of regions differentiating DNA copy number-based clusters Called CGH data for 42 HGSOCs, preprocessed as described previously  (Kamieniak et al. , 2013)  were subjected to unsupervised hierarchical clustering using total linkage and overall similarity algorithms implemented in WECCA (Weighted Clustering of Called aCGH Data) R package  (Van Wieringen et al. , 2008). The cut-off of two clusters was chosen by plotting a ratio of the within-cluster similarity and the betweencluster similarity (a statistics that evaluates the effect of a new cluster split). Two was shown to be the optimal cut-off and optimal number of clusters. The DNA copy number alterations that best distinguished the generated clusters were defined in Nexus Copy Number v5. 1 (BioDiscovery, Inc; El Segundo, CA) using "comparisons" option, minimal frequency difference between both clusters of 35% and Fisher Exact Test with significant p-value lower than 0.
Validation at the global gene expression level The association of the 6q24. 2eq26 deletion with survival was validated at the gene expression level using KM-plotter  (Gyorffy et al. , 2012)  with the JetSet probset. The KM-plotter is an online tool that allows the assessment of the prognostic value of the expression levels of microarray-quantified genes in ovarian cancer patients. The current database is set up using gene expression data and survival information of 1436 ovarian cancer patients downloaded from Gene Expression Omnibus and The Cancer Genome Atlas (10 different datasets). The association with overall survival was assessed in 799 HGSOCs (grade 3) and in 675 high-FIGO HGSOCs (FIGO III and IV, grade 3). The mean expression of all the genes from the 6q24. 2eq26 region with available data in the KM-plotter (53 out of 85 with validated RefSeq, as for May 2014) was used and the data was dichotomized at the automatically selected, best fitted cut-off into higher and lower expressing groups.
Definition of candidate genes at 6q24eq26 In order to propose individual candidate genes that might explain the observed association with patient survival, we first identified those genes in the 6q24e26 region whose loss had an impact on expression. To this end we used 232 tumors from TCGA ovarian study for whose copy number and expression data were available and assessed a total of 81 genes (out of 85 in the region with validated RefSeq) localized at 6q24. 2eq26 for whom RNAseq data (RPKM) was available. The Wilcoxon rank test was used to compare the expression values of each gene between tumors with normal copy number status and tumors with genomic loss at each locus. Significant genes were then evaluated for potential association between their expression levels and patient survival. Before running survival analysis, RPKM values were normalized, by subtracting the mean of all samples and dividing by the standard deviation, in order to allow a direct comparison of hazard ratios between genes.
3. 4. Validation of the prognostic value of the 6q24e26 deletion in the TCGA ovarian cancer series We aimed to further validate the prognostic value of the 6q24. 2eq26 deletion in a larger homogeneous group of 411 HGSOCs from TCGA ovarian cancer study (TCGA, 2011). Normalized log 2 ratios from 1 M Agilent Sure Print Human Microarray platform were downloaded from TCGA website (https://tcga-data. nci. nih. gov/tcga/dataAccessMatrix. htm) and subjected to the segmentation and calling algorithms using CGHcall R package and following the pipeline applied for our discovery series  (Kamieniak et al. , 2013). The previously defined region of loss at 6q24. 2eq26 (as described in Section 2. 4) was also called as "deleted" in more than 35% of TCGA series. Tumors were considered deletion-positive if at least 90% of the defined region (6q24.
Acknowledgments This study was financially supported by the  Fondo de Investigaci on Sanitaria (FIS) ,  Instituto de Salud Carlos III  (grants  PS09/ 01094  and  PI12/01319 ). MJG is recipient of a research contract from the  Instituto de Salud Carlos III of the Ministerio Español de Sanidad y Consumo  ( Miguel Servet tipo II Program ,  CPII 13/ 00047 ). MMK and KI have financial support from the  Fundaci on La Caixa. DR and KI are supported by grant  BIO2012-40205. LP-A, JCC and JP are recipients of financial support from  Red Tem atica de Investigaci on Cooperativa en C ancer (RTICC )  (grants  RD12/ 0036/0028 ,  RD12/0036/0037  and  RD12/0036/0064 ). We want to thank to  Mar ıa del Carmen Gonz alez-Neira  and  Victoria Fern  andez (from the  CNIO Human Cancer Genetics group)  and to the  CNIO Tumor Bank Network , in particular to  Mar ıa Jes us Artiga , for exceptional assistance in the retrieval of clinical data.
Mouse 1 Mouse 2 Figure  3  Effect on the viability of CD45R/B220 -CD138 þ plasmocytes of P140 peptide administrated in different media to MRL/ lpr lupus mice. Two 12-week-old female mice received the peptide intraperitoneally once per day during four consecutive days (100 mg/mouse). At day 5, the blood was collected, peripheral white cells were purified on a ficoll gradient, and purified cells were labeled with antimouse CD45R/B220-fluorescein isothiocyanate (FITC) and CD138-allophycocyanin (APC), both from BD Biosciences. Cells were analyzed using a FACSCalibur TM apparatus and the CELLQuest TM software (BD Biosciences). It is the case that trehalose, which is a known mTOR-independent activator of the macroautophagy pathway,  3, 21  further activates a mechanism that is already deregulated towards an excess of activity.
Material This study has been carried out on a subset of the ICP Forests (International Co-operative Programme on Monitoring and Assessment of Air Pollution Effects on Forests) / EU Forest Focus permanent intensive monitoring (Level II) plots that had been selected by countries participating in the ForestBIO-TA (Forest Biodiversity Test phase Assessments) project. Both surveys included 91 (dead wood) and 89 (stand structure, a subset of the dead wood plots) plots in 11 countries. According to ICP Forests (2010) one of the selection criteria for Level II plots was to have typical forests, and thus the plots are mainly located in managed forests. However, in a few cases stands recently abandoned or even unmanaged for some decades (e. g. , von Lührte & Seidling 1993) have been included. Data on dead wood and stand structure were collected according to the ForestBIOTA field protocol  (ForestBIOTA 2005).
2. Each plot was assigned to the forest type classification scheme of the BEAR project  (Barbati et al. 2007 , see also  Fischer et al. 2009) , which is almost congruent with the EUNIS classification  (EEA 2006). The pHCaCl2 of the organic and upper mineral soil layers and the annual wet throughfall deposition of sulphur and nitrogen had been determined for these plots within the regular ICP Forests surveys according to national protocols meeting the requirements described in the ICP Forests Manual (ICP Forests 2010). Values had been taken from the ICP Forests database available in 2006 and referring approximately to the time span from Tab. 1 -List of attributes surveyed in the field for dead wood and stand structure assessment. (*): References: Finland, Laasasenaho 1982; Greece,  Apatsidis & Sifakis 1999; Italy, Castellani et al.
Evaluations All dead wood related parameters were log-transformed to approximate normal distributions. To give answers to the questions, which of the parameter from both domains are most distinguishing within this geographically broad-scaled dataset, all dead wood parameters and, separately, all stand structural parameters (and indexes) were correlated with each other (PROC CORR, SAS 9. 2). As the domain-related correlation analyses with high numbers of parameters and comparatively low numbers of cases revealed a complex dependency structure, multiple re-gression or respective mixed models seemed not to be suitable approaches. Therefore, principal component analysis (PCA, PROC FACTOR, SAS 9. 2) was applied as an appropriate means to combine information from the dead wood and the stand structure survey (e.
If categorical parameters were used as predictors, main effects models were applied (PROC GLM, SAS 9. 2). All p-values were adjusted post-hoc for multi-comparisons with the Bonferroni-Holm procedure. The selection of the parameter was based on both availability and hypothetical importance of processes in forest ecosystems. Other available parameters, like longitude or discrete N-compounds, had tentatively been tested. However, they have been excluded from this family of bivariate analyses in order to avoid overload of the Bonferroni-Holm procedure.
6). This does not necessarily mean that such a relationship does not exist, however, the dataset was too small to support models with nested dependency structures or interaction effects between country and management intensity. For instance,  Fridman & Walheim (2000)  calculated 7 m 3 ha -1 for managed and 30 m 3 ha -1 for unmanaged forests in Sweden. As already mentioned, the ICP Forests manual recommends that intensive monitoring plots should be located in managed forests (ICP Forests 2010), thus the generally low quantity of dead wood is not surprising. Mean dead wood volumes are in the order of magnitude known from national forest in-ventories in Europe, even if harmonization efforts between dead wood inventories of different states have not yet been finalized  (Rondeux et al. 2012). For Switzerland 19 m 3 ha -1 were estimated on average  (Böhl & Brändli 2007) , for Italy 9 m 3 ha -1  (Pignatti et al.
Therefore the number of tree species in combination with evenness may give a rather comprehensive picture of both qualitative and quantitative aspects of tree diversity  (Liu 1995 , Magurran 2004 ). However, if an indicator integrating both species richness and species abundance is aspired, Shannon's or Simpson's index has to be used. The latter gives higher importance to the more abundant species  (Tóthmérész 1998). Mean mingling is given less priority. In our dataset it was closely correlated with the other diversity measures. As it refers to small-scale mingling of species, it might describe spatially differentiated mixtures of tree species at respective scales more efficiently. Basal area or the closely related volume of living trees per area, along with stem density, are doubtlessly important features of forest stands and should by all means be reported.
Integrated evaluations The joint evaluation of stand structure and dead wood was based on the assumption that both are closely linked at each site by common historic and recent forest management practices and natural processes, especially climatic events. Such mostly instantaneous interventions might be superimposed by soil conditions and unintentional anthropogenic factors like air pollution impacts. PCA is an efficient method to reduce the dimensionality of a multidimensional response structure composed in this case of different features of dead wood and the horizontal stand structure (Jolliffe 2002). Reductions of dimensionality are especially necessary for datasets with a comparatively low numbers of cases and large plenty of parameters, typical for networks composed of case studies like the Level II network of ICP Forests (De  Vries et al.
Against an intuitive or a purely hypothesis-guided pre-selection of considered parameters, which may have advantages of its own  (Klap et al. 2000) , this approach makes it highly probable that all important statistical relationships are considered, which is especially important for studies with a more exploratory character. The dataset presented originates from a huge geographic area, covering macro-ecological gradients across Europe and a wide range of management practices between countries, which both inevitably influences the relationships between forest ecosystem components. Therefore, categorical parameters like country or forest type reveal relationships to more than one PCA axes. Especially the amount of dead wood and most of its fractions, which differentiated according to the PCA the plots best, seems largely be influenced by country-specific forest legislations and management practices, like the above mentioned removal of woody debris in the Mediterranean countries.
95). This might be due to a lack of unmanaged forests, and due to often purely descriptive approaches  (Linder et al. 1997 , 1998 , Schulze et al. 2009 ). The positive relationship between the number of tree layers and PCA axis 3 may refer to the same complex of stand features, as forest stands more differentiated in tree height should also reveal a higher differentiation of tree diameters. Latitude and altitude are proxies for the major climatic drivers, e. g. , temperature and precipitation. Country does also not vary independently from climatic conditions due to their geographically fixed positions. Therefore, relationships with both variables partly include influences of climatic factors. Any statistical separation of respective climatic and management effects needs either much larger or more systematic datasets.
This can only be achieved by inventories based on probabilistic sampling schemes  (Winter et al. 2008 , Rondeux et al. 2012)  possibly combined with remote sensing techniques  (Stümer & Köhl 2005 , Corona 2010 ). Thus, the results described above apply primarily for the evaluated set of plots. However, considering the ForestBIOTA dataset as "found" data  (Overton et al. 1993) , which can be taken as a random sample with respect to the questions asked, the detected relationships might be valid beyond this mere sample. This study may be also of particular importance as a reference to which different national methodologies can be linked for harmonization or bridge-building between different national forest inventories  (Winter et al. 2008 , Rondeux et al. 2012 ).
Acknowledgments We dedicate this work to  Matthias Dobbertin  ( †: 31 st October 2012) remembering him and the stimulating discussions on this and related subjects. This work was supported by national experts, who contributed to method adaptation, were responsible for field work, and stimulated the evaluations. The teams were led by  Václav Buriánek  ( Czech Republic ),  Lars Vesterdal ,  Annemarie Bastrup-Birk  and  Jacob Heilmann-Clausen (Denmark) ,  Maija Salemaa  and  Tiina Tonteri  (Finland),  Ulrich Matthes ,  Joachim Block ,  Marcus Schmidt ,  Henning Meesenburg ,  Gerhard Raben ,  Claus Schimming ,  Angela Steinmayer , and  Lutz Genssler  (Germany),  George Baloutsos and Evangelia Daskalakou (Greece), Bruno Petriccione (Italy) ,  Gerard Grimberg  and  Han van Dobben  (The Netherlands),  Jozef Vladovic ,  Roman Longauer , and  Jozef Istona  ( Slovak Republic ),  Gerardo Sanchez Peña , and  Mayte Minaya (Spain) ,  Norbert Kräuchi ,  Franziska Heinrich  and  Nadine Hilker  (Switzerland), and  Igor Buksha (Ukraine).
Additional measurements For the determination of O 3 eddy covariance fluxes, required for the application of the MBR method, a closed-path fastresponse O 3 detector (Enviscope GmbH, Germany) was employed together with a sonic anemometer (CSAT3, Campbell Scientific Inc. , USA) at z ref = 3 m a. g. l. (see  Moravek et al. (2014)  for details on eddy covariance setup and O 3 calibration). In addition, CO 2 and latent heat fluxes were determined with an open-path CO 2 /H 2 O analyser (LI-7500A, LI-COR, USA). All turbulent fluxes were calculated using the eddy covariance software TK3. 1  (Mauder and Foken, 2011) , applying state-of-the-art corrections methods as listed in  Foken et al. (2012). Additionally, the O 3 flux was corrected for high-frequency loss of the 2. 5 m long inlet tube  (Moravek et al.
Acknowledgements. This project was funded by the  Max Planck Society. J. -C. Mayer  and  D. Plake  contributed to the setup and operation of additional measurements at the field site. L. Smoydzin  applied the WRF model for the estimation of the boundary layer height. We thank  C. Ammann  and  D. Plake  for discussion on the flux divergence and chemical correction methods. The service charges for this open access publication have been covered by the  Max Planck Society. Edited by:  J. Roberts.
Background: French general practitioners (GPs) were enrolled in a new payment system in January 2012. As part of a national agreement with the French National Ministry of Health, GPs were asked to decrease the proportion of patients who continued their benzodiazepine treatment 12 weeks after its initiation and to decrease the proportion of patients older than 65 who were prescribed long half-life benzodiazepines. In return, GPs could expect an extra payment of up to 490 euros per year. This study reports the evolution of the corresponding prescribing practices of French GPs during that period regarding patients who were prescribed a benzodiazepine for the first time. Methods: The national healthcare system's administrative database was used to report the longitudinal follow-up of two historical cohorts of French patients from the Pays de la Loire area.
Design, setting, and patients This study used the national health care system's administrative database to report the longitudinal follow-up of two historical cohorts. Access to anonymized data was provided by the National Healthcare Insurance services who participated to the study, after permission of the Healthcare Insurance authorities. All eligible patients lived on the French West Coast in the Pays de la Loire geographic area (3,571,495 inhabitants), were older than 16 years and were affiliated with one of the 1,350 GPs who practised in the geographic area at the beginning of the study (April 1, 2011). The "2011 cohort" included all patients who had been prescribed a benzodiazepine from April 1 to June 30, 2011, and had not taken any benzodiazepines during the preceding 4 months. The "2012 cohort" included all patients who had been prescribed a benzodiazepine from April 1 to June 30, 2012, and had not taken any benzodiazepines during the preceding 4 months.