dataesr/scientific-paragraphs-categorization huggingface dataset:
Tat can also induce OS by downregulation of the antioxidant mechanisms. Here we have shown that Tat induced mitochondrial ROS generation and a simultaneous decrease in cellular GSH/GSSG ratio in B-cells. Our data are in agreement with the previous study where Tat was found to reduce the GSH/GSSG ratio in Fig. 4. Tat induces glutathione deficiency in B-cells. (A) Intracellular glutathione (GSH) staining using monochlorobimane (MClB). B-cells treated with 250 ng/ml of Tat for 1 h and untreated controls were stained with MClB and the amount of emitted fluorescence was analyzed by FACS. The experiment was carried out on blood from two different healthy donors. Total glutathione (GSH) (B) and GSH/GSSG ratio (C) determined using the GSH/GSSG Ratio Detection Assay Kit. B-cells treated with 250 ng/ml of Tat for 1 h and untreated controls were processed following the manufacturer's instructions.
The experiment was carried out on blood from two different healthy donors. (D) Analysis of expression of genes encoding for heavy and light subunits of γ-GCS in B-cells treated with Tat. The transcriptional levels were measured by RT-qPCR after 1 h of treatment with 250 ng/ml Tat. Data represent means ± SEM from three different donors and are expressed as fold change normalized vs. GAPDH and compared to untreated controls. (E) Transcriptional activity of Tat is essential for ROS production. Bcells were treated with either 250 ng/ml Tat or its transcriptionally inactive form TatC22. ROS production was measured by DHE staining 6 h after treatment. The Y axis shows ROS production calculated from a minimum of 1000 cells examined for each condition, compared to the untreated B-cells, for which the obtained basal value was set as 1.
The experiment was carried out on blood from two different healthy donors. (F) Transcriptional activity of Tat is essential for DNA damage induction. B-cells were treated with 250 ng/ml of either Tat or TatC22 mutant, and then fixed at 6 h post-treatment and immunostained for γH2AX. The number γH2AX-positive cells was calculated from a minimum of 300 cells was examined per time point, compared to B-cells as negative control, for which the obtained basal value was set as 1. The experiment was carried out on blood from three different healthy donors. All data are expressed as the mean ± SEM. The statistical analyses were carried out by the one-way ANOVA test. The statistical significance was calculated vs. untreated controls; ***: p value < 0. 001; **: 0. 001 < p < 0. 01; * 0. 01 < p < 0. 5.
Acknowledgements This research was supported by grants from  ANRS ,  INSERM (ENVIBURKITT) ,  La Ligue Contre le Cancer  ( M27231 ) to YSV, the  Russian Foundation for Basic Research  ( 16-54-16001 ) to EVS and the  Russian Scientific Foundation  ( 14-24-00107 ) to BC. TT was a recipient of the  ANRS  fellowship. The funders had no role in study design, data collection and analysis, decision to publish, or preparation of the manuscript.
E. S. S. Collaboration et al. 2016). The data were analysed using the Model Analysis (de Naurois & Rolland 2009), in which shower images of all triggered telescopes are compared to a pre-calculated model by means of a log-likelihood optimisation. This advanced reconstruction technique improves the sensitivity with respect to more conventional ones. The "Standard cuts" of the Model Analysis were adopted. This set of cuts includes a minimum charge in the shower images of 60 photoelectrons resulting, for this dataset, in an energy threshold of 290 GeV. This set of cuts was used for signal extraction as well as for the spectral and morphological analyses. The results presented in this paper have been cross-checked with the analysis methods described in  Ohm et al. (2009)  and  Parsons & Hinton (2014)  and yield compatible results.
Fermi-LAT observations and data reduction The LAT detects γ-ray photons by conversion into electronpositron pairs in the energy range between 20 MeV to higher than 300 GeV, as described by  Atwood et al. (2009). It contains a high-resolution converter/tracker (for direction measurement of the incident γ rays), a CsI(Tl) crystal calorimeter (for energy measurement), and an anti-coincidence detector to identify the background of charged particles. The following analysis was performed using 5 years of Fermi-LAT data collected primarily in survey mode (2008 August 04 -2013 August 04). This new dataset consists of Pass 7 LAT data that have been reprocessed using updated calibration constants for the detector systems  (Bregeon et al. 2013). The primary differences with re-spect to the Pass7_V6 instrument response functions (IRFs) are the correction of a slight (1% per year) expected degradation in the calorimeter light yield and significant improvement of the position from the calorimeter reconstruction, which in turn significantly improves the LAT point-spread function above 5 GeV.
Acknowledgements. The support of the  Namibian authorities and of the University of Namibia  in facilitating the construction and operation of H. E. S. S. is gratefully acknowledged, as is the support by the  German Ministry for Education and Research (BMBF) , the  Max Planck Society , the  German Research Foundation (DFG) , the  French Ministry for Research , the  CNRS-IN2P3  and the  Astroparticle Interdisciplinary Programme of the CNRS , the  U. K. Science and Technology Facilities Council (STFC) , the  IPNP of the Charles University , the  Czech Science Foundation , the  Polish Ministry of Science and Higher Education , the  South African Department of Science and Technology and National Research Foundation , the  University of Namibia , the  Innsbruck University , the  Austrian Science Fund (FWF) , and the  Autrian Federal Ministry for Science, Research and Economy , and by the  University of Adelaide  and the  Australian Research Council.
A. Wallenberg Foundation , the  Swedish Research Council  and the  Swedish National Space Board in Sweden. Additional support for science analysis during the operations phase is gratefully acknowledged from the  Istituto Nazionale di Astrofisica in Italy  and the  Centre National d'Études Spatiales in France. This publication makes use of molecular line data from the  Boston University-FCRAO Galactic Ring Survey (GRS). The GRS is a joint project of  Boston University and Five College Radio Astronomy Observatory , funded by the  National Science Foundation  under grants  AST-9800334 ,  AST-0098562 ,  AST-0100793 ,  AST-0228993 & AST-0507657.
Statistical analysis All statistical analyses were performed using GraphPad Prism 6 (GraphPad Software, USA). Data are presented as mean ± standard error of the mean (SEM). p < 0. 05 (calculated using a one-way ANOVA with Tukey's post hoc test) was considered statistically significant. For ANOVA tests with triplicate technical repeat in two independent lines at least, the different levels of significance are denoted by different symbols in the figures.
Acknowledgements This work was supported by the  Council of Agriculture, Executive Yuan, Taiwan  (grant numbers  106AS-2. 2. 2-AD-U1 (Z) and  107AS-2. 2. 2-AD-U1  awarded to SCW) and by  ANR, France  (grant number  CRB-ANIM-ANR-11-INBS-0003  awarded to BP). We thank  Mr. Nobutomo Tsuruzoe ,  Dr. Masato Horikawa ,  Ms. Chun-Yun Gu , and Ms. Charlene Hung  ( Nissan Chemical Industries, Ltd. ) for assistance and the  National RNAi Core Facility, Academia Sinica, Taiwan  for providing recombinant lentiviruses.
This paper presents a measurement of the polarisation of τ leptons produced in Z/γ * → ττ decays which is performed with a dataset of proton-proton collisions at √ s = 8 TeV, corresponding to an integrated luminosity of 20. 2 fb -1 recorded with the ATLAS detector at the LHC in 2012. The Z/γ * → ττ decays are reconstructed from a hadronically decaying τ lepton with a single charged particle in the final state, accompanied by a τ lepton that decays leptonically. The τ polarisation is inferred from the relative fraction of energy carried by charged and neutral hadrons in the hadronic τ decays. The polarisation is measured in a fiducial region that corresponds to the kinematic region accessible to this analysis. The τ polarisation extracted over the full phase space within the Z/γ * mass range of 66 < m Z/γ * < 116 GeV is found to be P τ = -0.
The τ decays are assumed to follow the Standard Model expectations. The first τ polarisation measurement at ATLAS was performed in W → τν decays in proton-proton collisions at the centre-of-mass energy of √ s = 7 TeV recorded in 2010  [9]. The concept to extract the polarisation from a template fit to a polarisation sensitive observable is retained from that analysis. To exploit the larger dataset collected at √ s = 8 TeV, refined experimental techniques for τ polarisation measurements at hadron colliders are utilised for the measurement presented in this paper. In particular, the impact of systematic uncertainties in the modelling of the polarisation observable for signal events and the significant backgrounds are estimated more thoroughly, because they are more important in the current measurement using a larger dataset.
23147. The τ decays were simulated using the Tauola decay library  [4]. The helicities of 2 ATLAS uses a right-handed coordinate system with its origin at the nominal interaction point (IP) in the centre of the detector and the z-axis along the beam pipe. The x-axis points from the IP to the centre of the LHC ring, and the y-axis points upwards. Cylindrical coordinates (r, φ) are used in the transverse plane, φ ∈ (-π, π] being the azimuthal angle around the z-axis. The pseudorapidity is defined in terms of the polar angle θ as η =ln tan(θ/2). Angular distance is defined as ∆R ≡ (∆η) 2 + (∆φ) 2. The ∆φ separation is defined as min(|φ 1 - τ leptons generated by the Tauola algorithm were not stored so that the helicity is reconstructed in the generated signal samples with the TauSpinner  [15]  package associated with the Tauola decay library.
The TauSpinner algorithm assigns the helicity of τ leptons randomly based on probabilities derived from the kinematic configuration of the τ decays. The signal sample is thereby split into events with left-handed τ - (and right-handed τ + ) and those with right-handed τ -(and left-handed τ + ). The TauSpinner algorithm averages over incoming parton flavours and four-momenta whereas the Tauola algorithm directly accesses the incoming partons in each event. The average over initial parton states is performed using the MRST-MCal PDF set  [16]  in this analysis. Spin correlations as expected in Z/γ * → ττ decays are assumed. The TauSpinner package was extensively tested and validated by its authors  [15, 17, 18]  and used in several measurements  [19, 20]. φ 2 |, 2π -|φ 1 -φ 2 |). For studies of systematic uncertainties, an auxiliary sample of Z/γ * → ττ events was produced using the Pythia8  [21]  event generator with the CTEQ6L1 PDF set and AU2  [22]  tune for the UE.
The uncertainty related to the choice of event generator for the signal sample is estimated with the help of two auxiliary samples produced with the Pythia8 event generator and with the Powheg event generator interfaced with the Pythia8 hadronisation and τ decay modelling (see Table  1 ). Because the latter was generated using the CT10 PDF set, it is reweighted to match the default one (CTEQ6L1) with the LHAPDF package  [47]  to avoid double-counting of possible systematic effects. These two samples are used to obtain a set of event weights relative to the default Alpgen sample before any event selection with respect to the kinematics of τ leptons and Z bosons and to the Υ spectra of various hadronic τ decay modes. The resulting uncertainties are among the leading ones in the analysis. Most of the impact arises from the uncertainties in the τ lepton pseudorapidity distributions and from the uncertainties in the Υ distributions in τ ± → h ± π 0 ν decays.
Conclusion A measurement of the τ polarisation in Z/γ * → ττ decays with one leptonic and one single-prong hadronic τ decay is performed. Sensitivity to τ polarisation is gained from the hadronic τ decay. The 20. 2 fb -1 dataset of proton-proton collisions at √ s = 8 TeV collected by the ATLAS experiment at the LHC in 2012 is utilised. The measurement is complementary to previous measurements in electron-positron collisions. In the fiducial region, the measured τ polarisation is P τ = -0. 27 ± 0. 02 (stat) ± 0. 04 (syst). It agrees with the value predicted by the Standard Model (as implemented in the Alpgen event generator interfaced with the Pythia6 and Tauola) hadronisation and τ decay modelling, which is P τ = -0. 270 ± 0. 006. The polarisation is then extracted in the mass-selected region of 66 < m Z/γ * < 116 GeV and a value of P τ = -0.
Acknowledgements We thank  CERN  for the very successful operation of the LHC, as well as the support staff from our institutions without whom ATLAS could not be operated efficiently. We acknowledge the support of  ANPCyT , Argentina; YerPhI, Armenia;  ARC, Australia ;  BMWFW  and  FWF , Austria; ANAS, Azerbaijan; SSTC, Belarus;  CNPq  and  FAPESP, Brazil ;  NSERC ,  NRC  and  CFI ,  Canada ; CERN; CONICYT, Chile; CAS,  MOST  and  NSFC, China ;  COLCIENCIAS, Colombia; MSMT CR, MPO CR  and  VSC CR, Czech Republic ;  DNRF  and  DNSRC, Denmark ;  IN2P3-CNRS ,  CEA-DRF/IRFU , France; SRNSF, Georgia; BMBF,  HGF , and  MPG, Germany ;  GSRT, Greece; RGC, Hong Kong SAR, China ;  ISF, I-CORE  and  Benoziyo Center , Israel; INFN, Italy;  MEXT  and  JSPS, Japan ; CNRST, Morocco; NWO, Netherlands; RCN, Norway;  MNiSW  and  NCN , Poland; FCT, Portugal; MNE/IFA, Romania;  MES of Russia  and  NRC KI , Russian Federation; JINR; MESTD, Serbia; MSSR, Slovakia; ARRS and MIZŠ, Slovenia; DST/NRF, South Africa; MINECO, Spain;  SRC and Wallenberg Foundation, Sweden ;  SERI ,  SNSF  and  Cantons of Bern and Geneva, Switzerland; MOST, Taiwan ;  TAEK, Turkey; STFC, United Kingdom ;  DOE  and  NSF ,  United States of America.
Methods The PICO criteria were adopted as follows: Patients: people with secondary or tertiary hyperparathyroidism in dialysis; Intervention: parathyroidectomy; Comparators: the same patients before and after the intervention; Outcome: changes in blood pressure. The eligibility criteria were broad, because we expected retrieval of a low number of papers with heterogeneous designs and definitions. Consequently, we included all published studies that dealt with blood pressure following parathyroidectomy. We included all study designs, i. e. prospective or retrospective cohort studies, case-control studies, trial-based analyses, and case reports. Reviews, whether or not systematic, were also searched to check for papers that had escaped from our wide search strategy. We searched PubMed indexed for MEDLINE, Embase and the Cochrane Review database from January 1, 2000 to December 31, 2016 using a combination of MeSH terms and keywords related to hyperparathyroidism, hypertension and dialysis, including: hyperparathyroidism, parathyroidectomy; dialysis, hemodialysis, hemofiltration, hemodiafiltration, renal replacement therapy; blood pressure, hypertension, hypotension, hypertensive and hypotensive.
We also checked the reference lists of relevant articles and of studies (Fig. 2 ). Although we did not limit our search to by language, we only considered studies published after 2000, to enable us to focus on the considerable changes in the management of the hyperparathyroidism that have occurred in the new millennium, and thereby contextualize our clinical decision in keeping with the current recommendations on hyperparathyroidism in dialysis patients. Articles identified were screened for eligibility solely on the basis of the content: papers reporting on any outcome regarding blood pressure after parathyroidectomy were selected. Data extraction was performed by AS and   Statistical analysis (meta-analysis) was performed using Metanalyst Beta 3. 13, for the outcome blood pressure.
Neurofilament-HuC/D double stained embryos were dehydrated and cleared as previously documented  [11]. Neurofilament-stained hearts were photographed using a Zeiss SteREO Lumar. V12 microscope equipped with an Axioxam HRc camera and UV lamp under a rhodopsin red filter. Neurons were traced using the software ImageJ (NIH, USA) and branch length extracted.
Statistical Analysis Total nerve branch lengths are represented as mean ± SD. Statistical analysis was performed using a one-way ANOVA followed by Dunnett's multiple comparison test between Tbx1 +/+ and Tbx1 +/-on one side and Tbx1 +/+ and Tbx1 neo2/-on the other side. For HR measurements, data are reported as individual values, means ± SEM, or as box and whiskers plots. Statistical significance of repeated measures at different time intervals was determined using mixed level modelling and statistical significance between 2 groups was determined by Mann-Whitney test. Differences between the groups with p < 0. 05 were considered to be significant. Statistical analysis was performed using prism 5, GraphPad (San Diego, CA, USA) and IBM SPSS Statistics (New York, NY, USA) Subscription.
Funding: This research was funded by a  British Heart Foundation  grant number  RG/15/13/28570  to  P. J. Scambler  and  A. Calmont  and number  RG/15/15/31742  to  A Tinker  and was facilitated by the  NIHR Biomedical Research Centre at Barts. R. A. was supported by a  Medical Research Council   Clinical Research Training Fellowship. This research was supported by the  NIHR Great Ormond Street Hospital Biomedical Research Centre. The views expressed are those of the author(s) and not necessarily those of the  NHS , the  NIHR  or the Department of Health.
Participants whose blood was collected before the dengue outbreak were invited to provide another blood sample after the dengue outbreak. This allowed  Kucharski et al. to identify individuals who had already developed immunity to dengue before the outbreak and those who were likely infected during the outbreak. Comparing blood samples taken before and after the outbreak revealed that children and teenagers between the ages of 10 and 19 had the greatest risk of infection during the outbreak. No other demographic or environmental factors were strongly linked to the likelihood of infection. Computer models using the data also showed that the clean-up efforts could explain the reduced dengue transmission during the outbreak. These findings suggest that studying immunity against dengue can lead to a better understanding of disease transmission.
, 2011; Reiner et al. , 2014) , which can be difficult to implement in island settings where outbreaks are infrequent and difficult to predict. Immediately before the 2013/14 dengue outbreak in Fiji, a population-representative serological survey had been conducted to study leptospirosis and typhoid  (Lau et al. , 2016). To investigate patterns of dengue infection in 2013/14, we followed up participants from this survey in Central Division, to obtain a set of paired pre-and post-outbreak serological samples (see Materials and methods). We tested the paired samples for anti-DENV IgG antibodies using ELISA and a recombinant antigen-based microsphere immunoassay (MIA), and combined these data with dengue surveillance data to compare possible explanations for the outbreak dynamics. We measured age-specific and spatial patterns of infection and reported disease, and tested whether there were demographic and environmental risk factors associated with infection.
Discussion We analysed surveillance reports and serological survey data to examine the dynamics of a major 2013/14 dengue outbreak in Fiji. Owing to the sporadic and unpredictable nature of dengue outbreaks in the Pacific  (Cao-Lormeau et al. , 2014) , it is rare to have access to paired population-representative sera collected before and after such an epidemic. Comparing surveillance and serological survey data made it possible to investigate the relationship between observed reported cases and the true attack rate and quantify the relative role of climate, herd immunity and control measures in shaping transmission. Analysis of detailed serological data provided insights into age-specific patterns of infection that would not be identified from seropositivity thresholds alone. We estimated the highest infection rate was in the 10-19-year-old age group, whereas proportionally the most reported cases were in the 20-29-year-old group.
Surveillance data In December 2013, the dengue outbreak in Fiji was determined to be due to DENV-3 by RT-PCR performed on serum samples sent to the World Health Organization Collaborating Centre for Arbovirus Reference and Research at the Queensland University of Technology (QUT, Brisbane). Hereafter, samples that were ELISA reactive for NS1 antigen or IgM were presumed to be to DENV-3 infections with a sub-sample of them sent for confirmatory serotyping at QUT, the Institut Louis Malarde(ILM) and the US Centers for Disease Control and Prevention. Of the 10,442 laboratory tested cases that were notified to the Fiji National Centre for Communicable Disease Control between 27th October 2013 and 4th March 2014, 4115 (39. 4%) were reactive for DENV NS1 and/or anti-DENV IgM. After this time period, dengue surveillance was transitioned from laboratory to clinical-based reporting (i.
It was not possible to perform the same analysis using the MIA data because unlike the ELISA and neutralisation assay data, the raw MIA values did not follow a bimodal distribution that indicated likely naive and previously exposed individuals (Figure  6 -figure  supplement 2B ). We used a generalized additive model with binomial link function to examine the relationship between ELISA value in 2013 and rise between 2013 and 2015, with data points weighted by probability that the change in ELISA value was the result of a genuine rise rather than measurement error. Risk factor analysis was performed using a univariable logistic regression model. Both were implemented using the mgcv package in R version 3. 3. 1  (Wood, 2006; R Core Team, 2015).
Model fitting The model was jointly fitted to laboratory-confirmed case data and serological data using Markov chain Monte Carlo (MCMC) via a Metropolis-Hastings algorithm. For the case data, we considered time units of one week. To construct a likelihood for the observed cases, we defined case count for week t as c t ¼ C t À C tÀ1. Because reporting switched from lab tested to DLI during the outbreak, we jointly fitted two sets of time series data. The first dataset was lab tested cases. We defined the first observation as 4th November 2013, the week of the first confirmed case in Central Division, and the last observation as 26th May. The second dataset was DLI cases, which we fitted from 1st February until 26th May. Earlier DLI cases were not included as these were likely to reflect reporting artefacts rather than genuine infections.
We fitted the model to either MIA or ELISA data, to reflect two different assumptions about the relationship between seroprevalence and immunity. The model using MIA data made the assumption that only individuals who were seropositive to DENV-3 were immune to this serotype. As a sensitivity analysis, the model using ELISA data assumed that seropositivity to any DENV serotype indicated immunity to DENV-3. All observations were given equal weight in the model fitting. The joint posterior distribution of the parameter set was obtained from 200,000 MCMC iterations, each with a burn-in period of 20,000 iterations. We used adaptive MCMC to improve efficiency of mixing: we iteratively adjusted the magnitude of the covariance matrix used to resample to obtain a target acceptance rate of 0. 234  (Roberts and Rosenthal, 2009).