dataesr/scientific-paragraphs-categorization huggingface dataset:
(2009)  modified by  Page et al. (2012)  for quantification against internal standard (apo_8 ′ carotenal) after HPLC separation on a C30-column (Develosil R C30-UG-3, Nomura chemical CO. , Seto, Japan) with a UV-visible detector (SPD-M20A; Shimadzu, Kyoto, Japan). The pericarp cell number was measured after tissue dissociation according to a method adapted from  Bünger-Kibler and Bangerth (1983). Cells were counted using a microscope equipped with a camera (QImaging, Surrey, Canada) and Qcapture Pro 6. 0 software (QImaging, Surrey, Canada;  Bertin et al. , 2002).
Data Analysis Data were analyzed using R statistical software (http://www. Rproject. org). Physiological traits, data of yield, and quality traits were analyzed by analysis of variance (the agricolae R package and aov function; De Mendiburu, 2014). Heteroscedasticity and normality tests were performed before model evaluation. Regarding field experiment, when the ANOVA F-test showed no significant difference in means between the blocks, data from block 1 and 2 were pooled. Tables  1, 2  synthesize results from univariate ANOVAs. Multiple comparison of means was performed using the Least Significant Differences (LSD) test (α = 0. 05). When heteroscedasticity was detected, we used the Kruskal-Wallis non-parametric test followed by multiple comparisons of means through a t-student test on the ranks (α = 0.
05). The Factomine R package and the plot PCA function were used to perform PCA analysis. Data from field were composed of 15 variables of fruit and puree quality and 48 observations (4 cultivars × 3 stages × 2 blocks × 2 irrigation levels). Data (means of 3 biological replicates) were centered and scaled by variables. Data from glasshouse are presented in Supplementary Figure  5. Eigen values and contribution of variables to each dimension are reported in Supplementary Tables 1, 2 for field and glasshouse, respectively. Pearson correlation was performed to investigate links between fresh fruit and puree quality traits. The coorplot package was used to draw the correlation matrix of the quality traits  (Wei et al. , 2016). The GGMselect, GeneNet, and igraph packages  (Schaefer et al. , 2013; Csardi and Nepusz, 2014; Bouvier et al.
Z-transformation The initial assessment of 180 healthy subjects from the DFNS reference database revealed that all parameters except PHS and DMA could be transformed (partly in log-space) to a standard normal distribution. 28, 31, 33 This process, called Z-transformation, normalizes all values to a mean 5 0 and an SD 5 1. Subsequently, all QST results of patients and healthy subjects were transformed in accordance with this normalization. Abnormal values are defined as values beyond the 95% confidence interval (CI). On a z-scale, this is represented by z-values ,21. 96 or. 1. 96. Paradoxical heat sensation and DMA usually do not appear in healthy subjects and therefore cannot be transferred to the same scale. Thus, PHS was transformed to a binary 0/2-variable showing absence (coded as 0) or presence (coded as 12) of pathological values.
Frequency of phenotypes across clinical entities If a new drug would be tested for efficacy in a phenotype-stratified subgroup with neuropathic pain of any single etiology, this would only be meaningful if the said phenotype appears in a relevant frequency within this etiology. To show how frequent these phenotypes are across 3 common etiologies of neuropathic pain, we applied the algorithm to patients suffering from neuropathic pain due to diabetic polyneuropathy, peripheral nerve injury, or postherpetic neuralgia from the databases of our previous studies. 5, 9, 10, 29, 36.
Frequency of phenotypes across clinical entities From the databases of the DFNS, IMI Europain, Neuropain, and the Pain in Neuropathy Study (PiNS), a total of 151 patients with painful diabetic polyneuropathy, 335 patients with painful peripheral nerve injury, and 97 patients with postherpetic neuralgia who had been part of previous analyses  5, 9, 10, 29, 36  underwent both deterministic and probabilistic phenotyping (see Table  3  for patients' basic characteristics). Frequencies of phenotypes are presented in Table  4 , and frequency and overlap between phenotypes for the full protocol and for each clinical entity are displayed in Venn and bar diagrams in Figure  3.
Discussion In a commentary to our findings of 3 distinct sensory phenotypes in patients suffering from neuropathic pain, we were asked to develop an algorithm to allocate individual patients to said phenotypes. 11 This work can be considered as a response to this request. We have developed an algorithm that enables individual allocation of patients to one or more sensory phenotypes, and, further, can separate patients from healthy subjects with a very high specificity of 94% and a sensitivity of 78%. All phenotypes are present in diabetic polyneuropathy, peripheral nerve injury, and postherpetic neuralgia in reasonable frequencies, resulting in acceptable estimated sample sizes for phenotype-stratified trials, especially if crossover-designed studies are planned. The algorithm will be implemented in the next update of QST managing software eQUISTA (distributed by StatConsult, Magdeburg, Germany), but the algorithm itself is free to use.
Acknowledgements The  EUROPAIN  project is a public-private partnership and has received support from the  Innovative Medicines Initiative Joint Undertaking  under grant agreement no. 115007 , resources for which are composed of financial contribution from the  European Union 's seventh framework programme (FP7/2007-2013) and  European Federation of Pharmaceutical Industries and Associations (EFPIA)  companies' in kind contribution. The  NEUROPAIN  project is an investigator-initiated European multicentre study with  Prof Dr Ralf Baron  as the principle investigator and 10 co-investigator sites, supported by an independent research grant from  Pfizer  Ltd. The  Pain in Neuropathy Study (PiNS)  was supported by the  Wellcome Trust , the  European Union 's  Horizon 2020 research and innovation programme  under grant agreement no.
Fast track trigger applications might benefit from this readout topology  [2]. Broadcasts of signals that are common in the detector could also be easily distributed using wireless data transmission, as a signal emitted by a single transmitter can be received by several receivers at the same time. This could be interesting for detector synchronization or to distribute a system-wide reset or trigger signal. These motivations have led to the formation of the WADAPT (Wireless Allowing Data and Power Transmission) consortium. The goal of this research group is to identify specific needs of projects that might benefit from wireless technologies with the objective of providing a common platform for research and development  [3]. Along that line, a wireless data transmission demonstrator for HEP applications is developed and shortly described in section 3.
A 60 GHz demonstrator for HEP The purpose of the demonstrator is to validate the feasibility of wireless data transmission for HEP applications. It has to fulfil several requirements: It must be operational after irradiation up to very high doses as expected for the HL-LHC. It has to provide data rates of the order of 5 Gb/s, similar to what is expected per link from the silicon detectors of the large experiments at HL-LHC. At the same time power efficiency is crucial. In order to benefit from wireless data transmission, the power consumption of the on-detector readout part should be reduced compared to commonly used readout architectures. At Heidelberg University a wireless transceiver chip is under development  [11, 12]. The prototype will be produced in the IBM SiGe BiCMOS 130 nm 8HP process.
Absorbing materials like low-density graphite foam can be used to attenuate reflections. A study was performed demonstrating that, by applying these means, two wireless links can be safely operated in parallel at a pitch of 5 cm between two highly reflective layers that are 10 cm apart, resembling a typical tracking detector scenario. A more detailed description of these studies can be found in  [14]. limit on the bit error rate of about 10 -13. At 10 Gb/s a bit error rate of 10 -14 was measured. These measurements demonstrate the enormous potential of wireless data transmission at high frequencies. However, the technology of transceivers operating above 200 GHz is still in the prototyping phase. To be fully competitive, the energy efficiency has to be improved. The link tested here consumes about 2.
We present new data and analyses revealing fundamental flaws in a critique of two recent meta-analyses of local-scale temporal biodiversity change. First, the conclusion that short-term time series lead to biased estimates of long-term change was based on two errors in the simulations used to support it. Second, the conclusion of negative relationships between temporal biodiversity change and study duration was entirely dependent on unrealistic model assumptions, the use of a subset of data, and inclusion of one outlier data point in one study. Third, the finding of a decline in local biodiversity, after eliminating post-disturbance studies, is not robust to alternative analyses on the original dataset, and is absent in a larger, updated dataset. Finally, the undebatable point -noted in both original papers -that studies in the ecological literature are geographically biased, was used to cast doubt on the conclusion that, outside of areas converted to croplands or asphalt, the distribution of biodiversity trends is centered approximately on zero.
INTRODUCTION Patterns of biodiversity over space and time represent the foundation of many ecological theories and conservation priortization schemes. Concerns have been raised  (Gonzalez et al. 2016)  about two papers that collectively analyzed >250 individual datasets on biodiversity change through time from many parts of the world  (Vellend et al. 2013 , Dornelas et al. 2014). Both of these studies found that the average magnitude of temporal change in alpha diversity across studies was indistinguishable from zero. Dornelas et al. (2014)  additionally showed significant and consistent temporal species turnover, thus highlighting marked shifts in community composition. The concerns of  Gonzalez et al. , which focus on species richness and diversity, are for the most part typical of those that could be directed at any ecological meta-analysis: different results might obtain in different places (underrepresented regions) or times (before people collected data of this nature), and it is possible to find data subsets that deviate from the overall pattern.
(3) There is no evidence in our original analyses or using a larger, updated dataset that the results were biased due to combining the effects of disturbance with post-disturbance dynamics. In fact, selectively excluding post-disturbance dynamics is itself a source of bias. (4) The ecological literature is indeed geographically biased, a fact discussed explicitly in both  Vellend et al. (2013)  and  Dornelas et al. (2014). The analysis of  Gonzalez et al. on  this issue supports the undebatable conclusion that new data (in this case from underrepresented regions) might modify conclusions from these meta-analyses, or from any other global-scale ecological meta-analysis.
SHORT-TERM TIME SERIES DO NOT PROVIDE BIASED ESTIMATES OF LONG-TERM TRENDS As we show below, a key component of the  Gonzalez et al. critique is incorrect (i. e. , not a matter of selective interpretation). Simulations of species richness (S) over 50-year periods and subsequent calculations of log ratios (log(S after /S before )) or slopes of richness on time during shorter time intervals (5, 10, 20 years) were used to argue that "Estimates of biodiversity change are systematically biased when syntheses are based on datasets composed primarily of short time series". Gonzalez et al. made two different errors, the first of which applies only to log ratios, the second of which applies to both log ratios and slopes: (i) When calculating a mean effect size for "short" windows of time, Gonzalez et al.
LOCAL BIODIVERSITY TRENDS IN STUDIES OF DIFFERENT DURATION The argument that short-term time series bias estimates of temporal biodiversity trends was used by Gonzalez et al. as a springboard to asking whether longer duration studies tend to show biodiversity declines. In this section, we address this issue for the two original studies in turn. Using the data from  Vellend et al. (2013) ,  Gonzalez et al. modeled  the log ratio of species richness at the end and start of a study (see previous section) as a function of the duration of that study, finding a statistically significant (p = 0. 04) but weak relationship (Fig. 2a ). They emphasized the conclusion that longer-duration studies tend to show richness declines, although by allowing for a non-zero intercept, their results also require explaining a nonsensical positive biodiversity trend in studies that last zero years.
With the larger data set of 212 studies (the 2013 paper had 155), there is no significant relationship between local richness change and study duration, regardless of whether one allows for a non-zero intercept (Fig. 2c, d ). The data in  Dornelas et al. (2014)  includes studies with diversity estimates for at least three time points, thus allowing the estimation of slopes of diversity vs. time, rather than only before-after log ratios. There is no significant relationship between the diversity-time slope and study duration (Fig. 3a, b ). Gonzalez et al. chose instead to calculate log ratios using the data in  Dornelas et al. (2014;  see Dataset S1 in that paper), thereby excluding most of the data used by Dornelas et al, and reported a significant negative relationship between log ratios and study duration (Fig.
(  2013 ) dataset (i. e. , leaving the impact-only data subset). Unlike the primary analyses in  Vellend et al. (2013) , Gonzalez et al. analyzed raw log ratios rather than those expressed as change per decade, and they weighted studies by the square root of sample size. The result was a distribution of effect sizes with an upper confidence limit that was slightly negative (1% loss). However, this significant result depends on the combined influence of the two analysis modifications introduced by Gonzalez et al. Using the original unweighted analyses and only the data reported in  Vellend et al. (2013) , the 95% credible intervals for the impact-only data subset selected by Gonzalez et al. overlap zero both for raw log ratios, if only slightly  ([-0. 095, 0. 012]  or between 9% loss and 1% gain), and for change expressed per decade ([-0.
061, 0. 031]). More importantly, using the larger, updated dataset, the upper credible intervals are well above zero (raw log ratio: [-0. 053, 0. 034]; per decade: [-0. 038, 0. 032]), allowing us to reject the conclusion that eliminating "recovery" studies reveals an average decline in local plant biodiversity. More generally, we disagree with the argument that disturbances (anthropogenic or otherwise) are generally expected to cause a decline in local biodiversity. Empirical studies testing the effect of disturbance on species diversity find a wide range of results, including no effects, positive effects, negative effects, and hump-shaped relationships  (Mackey and Currie 2001; Supp and Ernest 2014; Newbold et al. 2015; Vellend et al. 2017). Even logging -implicitly emphasized in Gonzalez et al.
(2014). Relationships were modeled assuming either a non-zero y-intercept (a,c,e) or a zero y-intercept (b,e,f), using either slopes (a,b) or log ratios (cf) to express temporal biodiversity change, and either including one outlier (a-d, "Original dataset") or not (e,f, "Modified dataset"). Lines represent the estimated effect size with credible intervals. See Appendix S1 for statistical methods and Metadata S1 for all data and computer code.
Acknowledgements We thank the patients who have participated and their families. NEAT022 trial was supported by  NEAT-ID Foundation , a not for profit private foundation to promote research and education projects in the HIV field. NEAT022 trial was also supported by  SSAT  and  ViiV Healthcare. We thank the NEAT022 study participants and their partners, families, care givers, and the staff of all the centres taking part in the study. We also thank the  European AIDS Treatment Group  for collaboration. Spanish centres and Spanish investigators were partially supported by the project  RD12/0017  integrated in the Plan Nacional IþDþi and cofunded by  ISCIII -Subdireccion General de Evaluacion and European Regional Development Fund (ERDF). We also acknowledge the Spanish HIV HGM biobank supported by the  Spanish Instituto de Salud Carlos III , which is integrated in the Spanish AIDS research network (RIS).
Conclusions Using a TPC demonstrator we were able to take polarised and unpolarised γ-ray data between 1. 7 MeV and 74. 3 MeV. The data analysis compared with our simulation shown that the performance of such detector (angular resolution: 8. 8 • @11. 7 MeV, polarisation asymmetry measurement capability: 13%@11. 7 MeV) have validated up to 20 MeV. The design of a larger detector is on study and is an inevitable path to the space-flight experiment long-awaited by the astrophysical community in order to fill up the gap between 1 MeV, and 100 MeV. This work was funded by the French National Research Agency (ANR-13-BS05-0002) and was performed by using NewSUBARU-GACKO (Gamma Collaboration Hutch of Konan University).
Patients Patients enrolled in the study had to be HIV/HCV genotype 1-coinfected adults, with a screening level of HCV RNA of at least 1000 IU/mL. Other inclusion criteria were: virological failure or premature discontinuation for intolerance of a previous triple therapy with pegylated (peg)-interferon/ribavirin and a first-generation PI; previous anti-HCV treatment interrupted for at least 3 months; stable antiretroviral treatment (combination of permitted molecules: tenofovir, emtricitabine, lamivudine, efavirenz, raltegravir, rilpivirine and enfuvirtide; as insufficient data on drug-drug interactions with LDV were available at this time, dolutegravir was not allowed) for at least 4 weeks before screening; HIV RNA < 50 HIV-1 RNA copies/mL for at least 3 months; CD4 count of > 100 cells/lL and CD4 percentage > 15%.
Statistical analysis Qualitative data are presented as number and percentage and quantitative data are presented as median and IQR or mean and standard deviation (SD), as appropriate. The primary efficacy endpoint was the SVR12 rate measured at week 36 for cirrhotic patients and week 24 for noncirrhotic patients. For analysis of the main endpoint, an intention-to-treat (ITT) analysis was performed with a test comparing an observed to a theoretical proportion using a unilateral formulation with a type I error of 5%. The rate of SVR12 below which the therapy would be considered inefficient was fixed at 50% (null hypothesis). The rate of SVR12 above which the therapy would bring a real benefit was fixed at 70% (alternative hypothesis). To guarantee 95% power to detect such a 20% benefit, and considering the statistical hypotheses, a minimum of 39 (out of 64) participants with an SVR12 were required to conclude that the SVR rate was > 50%.
This study was a nonrandomized, double-arm, open-label phase II pilot trial conducted between August 2014 and December 2015 in 30 French centres. The sponsor [Inserm-Agence Nationale de Recherche sur le Sida et les H epatites Virales (ANRS)] designed and conducted the study, and collected, monitored and analysed the data. Study drugs were kindly provided by Gilead Sciences, Inc. (Foster City, CA, USA). The study protocol (EUDRACT number 2013-002607-33) was approved by Ethical Committee CPP-Sud-M editerran ee-1 (Marseille, France) and by the French Regulatory Authority Agence nationale de s ecurit e du m edicament et des produits de sant e (ANSM) (Saint-Denis, France). It was registered in the ClinicalTrials. gov database under reference NCT02125500. The study was conducted in compliance with the Declaration of Helsinki and Good Clinical Practice Guidelines and regulatory requirements.
Acknowledgements Financial support was provided by the  Agence Nationale de Recherche sur le Sida et les H epatites Virales (ANRS) , with the participation of  Gilead Sciences. Dr Hugues Aumaitre ,  Dr Laurence G erard ,  Pr Michel Dupon ,  Pr. Laurent Alric ,  Dr Claudine Duvivier ,  Dr Elina Teicher ,  Dr Thierry Allegre ,  Dr Alain Lafeuillade ,  Dr Pierre de Truchis ,  Dr Vincent Jeantils ,  Dr Julie Chas ,  Dr Dominique Batisse ,  Pr Dominique Salmon-Ceron ,  Dr David Zucman ,  Dr Emmanuel Mortier ,  Dr St ephanie Dominguez ,  Dr Marc-Antoine Valantin ,  Dr Anne Simon ,  Pr. Pierre-Marie Girard ,  Pr. Jean-Michel Molina ,  Pr. Philippe Morlat ,  Pr. Didier Neau ,  Dr Djamila Makhloufi ,  Dr Isabelle Poizot-Martin ,  Pr. Yazdan Yazdanpanah ,  Pr. Jacques Reynes ,  Pr. Franc ßois Raffi ,  Dr Philippe Perr  e,  Dr Alissa Naqvi ,  Pr.
In this study, loss estimation models were developed for reasonably accurate assessment of economic and human losses from seismic events in the Mediterranean region, based on damage assessment at an urban scale. Data were compiled from existing worldwide databases, and completed with earthquake information from regional studies. Economic data were converted to a single common currency unit (2015 USD value) and the wealth of the areas affected by 65 earthquakes of the region from 1900 to 2015 was assessed. Reduced-form models were used to determine economic and human losses, with earthquake magnitude and intensity as hazard-related variables, and gross domestic product of the affected area and the affected population as exposure-related variables. Damage to buildings was also used as a hazard-related variable to predict economic and human losses.
But recent studies have shown the efficiency of data miningbased vulnerability assessment methods on a large scale that use elementary building characteristics, such as the number of floors and the roof shape. These structural features can easily be assessed by remote sensing or are contained in national databases  (Riedel et al. 2014 (Riedel et al. , 2015)). There are relatively few studies on human casualties in the literature. Samardjieva and Oike (1992) ,  Samardjieva and Badal (2002), and Badal et al. (2005)  analyzed and tested a model to estimate the (log value of the) number of casualties as a function of the magnitude and population density for earthquakes worldwide. Nichols and Beavers (2003)  established a simple equation between fatalities and magnitude for U. S. earthquakes. Jaiswal and Wald (2010)  developed a specific empirical model for estimation of earthquake fatalities based on the shaking intensity and the number of people exposed to earthquakes recorded between 1973 and 2007.
Moreover, in our dataset, many moderate earthquakes-for example, the 1960 Agadir earthquake-caused significant economic and human losses associated with severe building damage. Consequently, earthquake characteristics-for example, magnitude, intensity-alone are not sufficient to derive earthquake loss models. Moreover, to help earthquake crisis management and the prediction of the recovery period, the number of people who lost their homes is another key piece of information. The main objective of this study is to consider damage as a hazard-related parameter and to show improvement in the accuracy of loss estimation obtained in empirical models by using this parameter instead of intensity and/or magnitude alone as hazard-related parameters. We developed empirical models for economic losses and human losses (home loss, injuries, fatalities) from earthquakes as applied to the Mediterranean region through the compilation of data from different databases and countries.