dataesr/scientific-paragraphs-categorization huggingface dataset:
Re-hospitalisations and mortality during follow-up Information on re-hospitalisations through December 31, 2007 (causes and dates) was obtained for all patients from the Minimum Basic Dataset (CMBD), a national administrative database. According to the 9th revision of the International Classification of Diseases, an admission for COPD exacerbation was defined as any admission with codes 466, 480-486, 490-496, or 518. 81 as the main diagnosis. Survival status until December 31, 2008 was obtained from direct interviews with all patients or their relatives. In cases of death, both hospital and primary care registries were checked to verify the exact date.
Statistical analysis The sample size was fixed by the primary scientific objectives of the PAC-COPD Study  [16]. Before any analysis, we calculated whether the available number of patients (225 patients in the diagnosed group and 117 in the undiagnosed group) would allow for identification of clinically significant differences in outcome between groups (diagnosed vs. undiagnosed). Calculations using the GRANMO 5. 2 software  [24]  showed that, accepting an alpha risk of 0. 05 in a two-sided test, the statistical power was 84 to recognize as statistically significant the difference in proportion admitted (44% vs. 28%, respectively). Descriptive data are presented as the number and percentage, the mean and standard deviation (SD), or the median and 25th or 75th percentiles, as appropriate. We compared the sociodemographic and clinical variables and use of healthcare resources prior to first hospitalisation according to previous COPD diagnosis status, using Student's t-test or Mann-Whitney U test for quantitative variables and a Chi squared or Fisher exact test for qualitative variables.
Multivariate models included as covariates all potential confounders that were related to both the exposure and the outcome, or modified the estimates (>10% change in Hazard Ratio) for the remaining variables. Potential covariates included gender, age, marital status, smoking status, quality of life, degree of dyspnoea, BMI, FFMI, the Charlson index of comorbidity, FEV 1 , DLco, Residual Volume/Total Lung Capacity (RV/TLC), PaO 2 , arterial carbon dioxide tension (PaCO 2 ), 6MWD, and anxiety and depression. The same approach was to be used to assess the effect of undiagnosis on mortality; however, there were very few deaths during follow-up and this multivariate analysis was not completed. Data analyses were conducted using Stata 10. 1 (StataCorp, College Station, TX, USA).
A deltoid muscle biopsy was performed, and histopathologic findings revealed features of myofibrillar myopathy with desmin-positive protein aggregates (Figure  3 ). Genetic molecular analysis revealed a heterozygous missense mutation (c. 38C4T; p. Ser13Phe) in the desmin gene (GenBank accession number NM001927. 3) (Supplementary Figure, available online). The variant fulfilled all the criteria for a diseasecausing mutation. Additionally, this mutation was already published as pathogenic  [1] [2] [3]  and is located in the head domain of the protein. No mutation was identified in the lamin A/C gene. Following these results, the pacemaker was upgraded to an implantable cardioverter-defibrillator (ICD); genetic screening of relatives is ongoing. After 24 months of follow-up, no recurrence of chest pain occurred and echocardiograms remain normal, without evidence for structural cardiomyopathy on echocardiography.
Methods Analyses were performed using data from the CHIMES study, an international, randomized, placebo-controlled, double-blind trial that compared MLC601 with placebo in patients with ischemic stroke of intermediate severity in the preceding 72 hours (clinicaltrials. gov 7] [18]  Subjects were allocated to either MLC601 or placebo for 3 months as add-on to standard stroke care (ie, antiplatelet therapy, control of vascular risk factors, appropriate rehabilitation) and followed for 3 months. The primary outcome measure used in this study was the modified Rankin scale (mRS) score at 3 months. Of 1099 subjects in the CHIMES study, 1006 with complete baseline data and an mRS score at month 3 were included in this post hoc analysis. Logistic regression analyses were performed to identify predictors of mRS score greater than 1 and to assess the association between number of predictors and mRS.
In this analysis, also called sliding dichotomy, subjects are grouped into a number of bands according to baseline prognosis, wherein each band is analyzed according to a customized predefined ''good'' and ''bad'' outcome on a scale. 6 This was performed in the International Surgical Trial in Intracerebral Hemorrhage studies but failed to show statistical significance. 25, 26 here are some limitations in this study. The prognostic variables we identified will need to be externally validated in another data set, although each was already often shown to be important predictive variables in many other studies. 19 The analyses were post hoc, and the trial was not originally planned for such prognosisbased analysis. However, the data used were collected before unblinding. In summary, prognostic factors for functional outcomes are age, sex, stroke severity, and stroke onset to treatment delay.
Acknowledgments: CHIMES Study Investigators-Philippines:  Jose C. Navarro ,  Herminigildo H. Gan ,  Annabelle Lao ,  Alejandro Baroque II ,  Johnny Lokin ,  John Harold B. Hiyadan ,  Ma. Socorro Sarfati ,  Randolph John Fangonillo ,  Neil Ambasing ,  Carlos Chua ,  Ma. Cristina San Jose ,  Joel Advincula ,  Eli John Berame , and  Maria Teresa Canete ; Singapore:  Narayanaswamy Venketasubramanian ,  Sherry H. Y. Young ,  Marlie Jane Mamauag ,  San San Tay ,  Shrikant Pande ,  Umapathi Thirugnanam ,  Rajinder Singh ,  Hui Meng Chang ,  Deidre Anne De Silva ,  Bernard P. L. Chan ,  Vijay Sharma , and  Teoh Hock Luen ;  Thailand: NiphonPoungvarin ,  Sombat Muengtaweepongsa ,  Somchai Towanabut ,  Nijasri Suwanwela ,  Songkram Chotickanuchit ,  Siwaporn Chankrachang , and  Samart Nitinun ;  Sri Lanka: H.
D. Pair k T cuts and binning Similar to previous analyses  [33] [34] [35] 40]   T + m 2 π ) in each bin. These bins yield mean k T values similar to those in the data from previous analyses allowing direct comparison of certain quantities to previously observed trends. In earlier azimuthal HBT studies by CERES  [41]  and STAR  [40]  the analysis was performed in similar, narrow k T bins. For an azimuthally differential HBT analysis the statistics are spread across at least four additional azimuthal bins. At the lowest energies this did not allow for sufficient statistics. For instance, the 7. 7 GeV dataset has both the fewest number of events and the lowest multiplicity per event in each centrality bin. Reliable results could not be obtained from data split into both multiple k T and multiple bins relative to the reaction plane.
VI. RESULTS The azimuthally integrated HBT results are discussed first and compared to historical data from earlier experiments and recent results from ALICE. Later, the azimuthally differential analysis is presented for a wide range of beam energies. The azimuthally differential analysis is also performed in three rapidity bins allowing extraction of the excitation function for the R 2 ol parameter and direct comparison of the freeze-out eccentricity in the same forward rapidity window as an earlier measurement by the CERES collaboration. Finally, the excitation function for the freeze-out eccentricity is discussed along with its implications for the relevant underlying physics as outlined in Sec. II.
When comparing different datasets from previous analyses  [25, 28, 36] , there is an uncertainty on the centrality caused by the different techniques that were used to compute the average charged track multiplicity dN ch /dη. In this analysis, the standard STAR centrality definition was used at all energies, where dN ch /dη is computed using all events that pass the event selection cuts. However, it should be noted that this is an uncorrected value of dN ch /dη that underestimates the true An estimate of the volume of the homogeneity regions, V = (2π) 3/2 R 2 side R long , can be computed using the data in Fig. 6. These values are plotted in Fig. 9  as a function of √ s NN. The STAR results are all for 0-5% central collisions with k T ≈ 0. 22 GeV/c. Since the values are computed using the data in Fig.
Acknowledgments Laboratoire Aimé Cotton  is a member of  Institut Francilien de Recherche sur les Atomes Froids (IFRAF)  and the  LABEX PALM initiative. The work presented here has received funding from the  European Research Council  (grant agreement n. 277762 COLDNANO). We are grateful to  P Yzombard ,  M Doser ,  S Gerber ,  A Fioretti ,  C Gabbanini ,  D DeMille  and  C Koch  for helpful discussions.
The interval used to determine the VLD signal means just before the ELM moves back with dt. FIG. 6. Polar plots of histograms of VLD2 (left panels) and VLD3 (right panels) temporal analytic phase difference D/ just before an ELM in the flat-top of JET plasma 83771. The same data and analysis are used as in Figure  5  except that now Hilbert transform time window (b) is used to determine the VLD temporal analytic phases and it stops at t ELM2 , the time of the second ELM. Phase differences can then only be determined for times dt > dt E ¼ 1ms before t ELM2. From bottom to top dt ¼ ½1, 2, 5, 10ms. differences are aligned, and the statistical significance of any such alignment. Using the procedure described above, we determine the temporal analytic phase differences D/ k for the k ¼ 1; :::; N ELM pairs in given plasma.
We have found that for dt > 5ms, the details of where shortlived fluctuations in R, r / , and p-value occur are not robust, they vary with the dataset and with the detailed parameters of how the Hilbert transform is computed. However, the overall trends are robust, in particular, alignment of the temporal analytic phases around a single value for dt < 5ms, that is, large R and small r / and p-value. In Figure  6 , we only used signals up to, and not beyond, the time of the ELM in order to test for the ELM build-up signature in the full flux loop temporal analytic phases. The corresponding circular statistics are plotted in Figure  9  where the temporal analytic phase differences are obtained for t 0 ¼ t VLDmin. The blue line in Figure  9  replots that in Figure  8 , it is calculated using Hilbert transform time window (a) which extends to times beyond the ELM occurrence time.
We aim to test that the above results are significant compared to a random process. We also aim to quantify trivial correlation, that is, coincidences between ELM arrival time and full flux loop temporal analytic phase. Coincidences could arise in a finite dataset where both the sequence of ELM arrival times, and the full flux loop signals contain time structure that includes periodicity. Here, the ELM waiting times have a mean period and a "comb like" multi-periodic structure, and the full flux loop signals exhibit intervals of oscillatory behaviour. We will calculate the same circular statistics in exactly the same manner as above for the following surrogate datasets, the results are shown in the three panels of Figure  11. (C) Full flux loops are single constant frequency sinusoids (Figure  11 , bottom panel): We replace the full flux loop signal with a sinusoid with period T ¼ 10ms, approximately the characteristic period of the oscillatory response seen following an ELM.
ACKNOWLEDGMENTS We acknowledge the  EPSRC  for financial support. This work has been carried out within the framework of the  EUROfusion Consortium  and has received funding from the  European Unions   Horizon 2020 research and innovation programme  under Grant Agreement No. 633053 , and from the  RCUK Energy Programme  [Grant No. EP/I501045 ]. The views and opinions expressed herein do not necessarily reflect those of the  European Commission. We acknowledge  J. Davidsen  and participants in the 2014 MPIPKS workshop on Causality,  Information Transfer  and  Dynamical Networks  for discussions, and  A. J. Webster  for provision of data.
acknowledgmentsThe authors thank  E. Silva ,  V. V. Ryazanov  and  A. S. Mel 'nikov for stimulating discussions. This work was supported, in part, by the  NanoSC-COST (Belgium) ,  Action   MP1201  , by  ANR (France) , Grant MASH and by the  Russian Foundation for Basic Research. One of the authors (A. V. S. ) is supported by the  Russian Scientific Foundation  Grant No  15-12-10020.
Material and methods The control group (C, n=12) was fed a diet targeting high production (35kg/day Energy-Corrected Milk, ECM). The lower feeding intensity (L, n=12) was achieved by giving -50% concentrate to target 25kg/day ECM. Blood was sampled every 2 weeks until 2 months postpartum and then once a month until Day 120. Plasma was kept frozen until analysis for glucose, insulin and non-esterified fatty acids (NEFA) concentrations. Data were analyzed with Mixed linear models.
Conversely, the impact of pEP1 on the production of autoantibody was less pronounced. Indeed, injection of 10 μg of pEP1 reduces the production of anti-CII antibodies but not significantly (p = 0. 1) (Fig  3D). IgG1/IgG2a ratio of anti-CII antibodies is unchanged in ENO1 and pEP1 treated mice In order to investigate whether ENO1 could favor a Th2 or Th1 cellular response, we measured the levels of anti-CII IgG1 and IgG2a antibodies. Although the production of anti-CII antibodies was reduced in mice treated with ENO1 or pEP1, the IgG1/IgG2a ratio remained unmodified in both experiments (Fig 4A  and 4B ). Levels of anti-CCP antibodies were also measured but no production was observed in our different experiment settings (data not shown).
Comparisons with the operational CDRs We applied the size distribution retrievals to the improved method using the global POLDER L1B data from June 2008. The retrievals were performed with scattering angles of 137-165 • , polarized reflectances at the 865 nm wavelength, and a resolution of 150 km × 150 km. To make the comparison more convenient and reliable, an inversion was performed according to the geolocations of operational size distribution products. We adopted the inversion quality index (Qual) to evaluate our retrievals because the Qual is used in the operational procedure to include the bias generated from the variability of the signal. The Qual is the ratio between the variability in the signal and the RMSE of the fitting. A larger Qual value means a smaller fitting bias. Detailed information on Qual can be found in  Bréon and Doutriaux-Boucher (2005).
Acknowledgements. This work was supported by the  CAS/SAFEA International Partnership Program for Creative Research Teams  (grant no. KZZD-EW-TZ-09 ), the  National Science and Technology Ministry  (grant no. 2014BAC21B03 ) and the  National Natural Science Foundation of China  (grant no. 41471367 ). The data used in this paper were derived from the  CNES/POLDER  instrument onboard NASDA/ADEOS. The authors thank the  Interactions Clouds Aerosols Radiations Etc (ICARE)  thematic center for processing the POLDER data. We are grateful to  K. F. Evans  at the  University of Colorado  for the radiative transfer 3 and  Lorenz-Mie  codes. Edited by:  A. Kokhanovsky.
Objective: There is little evidence to guide treatment strategies for intracerebral hemorrhage on vitamin K antagonists (VKA-ICH). Treatments utilized in clinical practice include fresh frozen plasma (FFP) and prothrombin complex concentrate (PCC). Our aim was to compare case fatality with different reversal strategies. Methods: We pooled individual ICH patient data from 16 stroke registries in 9 countries (n 5 10 282), of whom 1,797 (17%) were on VKA. After excluding 250 patients with international normalized ratio < 1. 3 and/or missing data required for analysis, we compared all-cause 30-day case fatality using Cox regression. Results: We included 1,547 patients treated with FFP (n 5 377, 24%), PCC (n 5 585, 38%), both (n 5 131, 9%), or neither (n 5 454, 29%). The crude case fatality and adjusted hazard ratio (HR) were highest with no reversal (61.
4] [15] [16] [17] [18] [19] [20]  According to a prespecified protocol, we included patients aged 18 years taking any VKA at the time of their ICH. We excluded patients with ICH secondary to trauma or tumor, primary subarachnoid hemorrhage, or hemorrhagic transformation of ischemic stroke, and those with baseline INR < 1. 3. We also excluded those with missing data on reversal therapy received or variables used for adjustment: age, gender, INR, Glasgow Coma Scale (GCS), and imaging parameters (infratentorial location, intraventricular extension, baseline ICH volume). We prespecified exclusion of patients treated >24 hours from ICH onset, but later added these cases following a request from a manuscript reviewer.
No patients died before planned reversal therapy was given. All-cause case fatality by 30 days was 45. 7% (95% CI 5 43. 2-48. 2%) overall and varied markedly by treatment arm, being 61. 7% (95% CI 5 57. 2-66. 2%) with no reversal, 45. 6% (95% CI 5 40. 5-50. 7%) with FFP, 37. 3% (95% CI 5 33. 3-41. 2%) with PCC, and 27. 8% (95% CI 5 20. 1-35. 5%) with the combination therapy (Fig  2 ). Information on immediate palliation was missing in a large proportion of the patients without reversal, but in the 277 with these data recorded, 142 (51%) were immediately palliated. After adjusting for baseline imbalances using Cox regression, the HR for death within 30 days of ICH was the highest with no reversal (HR 5 2. 540, 95% CI 5 1. 784-3. 616, p < 0. 001), followed by PCC alone (HR 5 1. 445, 95% CI 5 1.
Discussion Our international, multicenter, observational study of 1,547 VKA-ICH patients demonstrates considerable variation in the practice of anticoagulation reversal after acute ICH. Most centers used PCC alone and some centers used solely FFP, whereas Australian centers systematically combined both. Treatment with any clotting factor product was associated with less than half the adjusted risk of death by 30 days. When comparing different clotting factor products, we found no significant difference between PCC and FFP. Case fatality was lower in patients treated with a combination of PCC and FFP, when compared to either alone. For those receiving PCC alone, we found that administration of 3-factor PCC was associated with improved survival compared to treatment with 4-factor PCC. An ongoing phase 2 trial (INCH; NCT00928915) is testing FFP versus PCC in 74 patients with correction of INR as the primary outcome,  23  but has not yet reported.
A recent German observational study on VKA-OAC (n 5 1,176), published after the submission of our study, suggested that rapid reversal of INR is associated with less ICH expansion and lower in-hospital mortality, but was not powered to compare reversal therapies due to a uniform national practice of using PCC. 24 Previous observational studies comparing treatments have been small (17-181 patients), with variable results with regard to case fatality,  14, 25, 26  functional outcome,  27, 28  and hematoma expansion. 29 These studies also combined patients receiving FFP with either PCC  26, 29  or no reversal,  14, 23, 27  precluding direct comparison of specific reversal strategies. Our data show that receiving no clotting factor replacement is associated with more than twice the risk of death by 30 days, when compared to all 3 treated groups.
32 Thus, administration of 4-factor PCC with significant FVII content may not be essential to reducing further bleeding and could increase thrombotic complications. The compositions of the multiple PCC products on the market also vary by heparin, antithrombin, and protein C and S concentrations, which all may be relevant. We did not have sufficient numbers to compare individual products, or collect data on thrombotic complications or causes of death, so further studies are required to test this hypothesis. Our study benefits from a large sample size of 1,547 patients. The patients included in the study are representative of routine clinical care in varied health care systems across a range of countries. The majority of patients presented between 2004 and 2013 and thus are more closely representative of current practice than cohorts described in previous studies.
21 Estimates of IVH volume varied and were missing in many patients, as were data on baseline blood glucose. Also, we did not adjust for other factors that have been variably associated with outcome in ICH, such as blood pressure or treatment delays, as they were not in our statistical analysis plan and not available. Finally, we do not have data describing functional outcomes, and thus do not know whether the association of improved survival with treatment is at the expense of a greater proportion of patients living with severe disability. Our findings support the hypothesis that the use of clotting factor replacement therapy in VKA-ICH is associated with improved survival. Although some guidelines recommend PCC over FFP based on limited evidence and expert opinion,  11  we found no evidence to support this policy.
Acknowledgment This study was supported by the  University of Manchester  (A. R. P. -J. ), the  Salford Royal NHS Foundation Trust  (A. R. P. -J. ), the  Academy of Medical Sciences  (A. R. P. -J. ), the  Dowager Countess Eleanor Peel Trust  (A. R. P. -J. ), the  Natalie Kate Moss Trust  (A. R. P. -J. ), the  Italian National Health Service  (M. D. N. ), a grant from the  Helsinki University Central Hospital  for research on intracerebral hemorrhage (T. T. ), a clinical established investigator grant from the  Dutch Heart Foundation  ( 2012T077 , C. J. M. K. ), an  ASPASIA grant  from the  Netherlands Organization for Health Research and Development  ( 015008048 , C. J. M. K. ), the  Finnish Brain Foundation  (J. H. ), the  Finnish Medical Foundation  (J. H. , A. M. ), the  NIH National Institute of Neurological Disorders and Stroke  ( R01 NS059727 , J.
R. ), the  Academy of Finland  (A. M. ), and the  National Health and Medical Research Council  (Australia; A. M. ). CROMIS-2 was funded by a grant from the  British Heart Foundation  and the  Stroke Association (UK ; chief investigator: D. We. ). LATCH was funded by an  MRC  senior clinical fellowship (R. A. -S. S. ) and an  MRC/Stroke Association  clinical research training fellowship.
7a  as a function of Dt, see Methods for further details). and refine the use of animals for scientific purposes, we used the minimal number of monkeys necessary to ensure replicability of the data: N ¼ 2. Following standard practice in awake monkey research, we validated the appropriateness of the chosen N by showing that statistical tests of significance were positive (Pr0. 01). Eye movements were sampled at 1 kHz with a Riverbend system, and single units were recorded extracellularly with lacquer-coated electropolished tungsten electrodes (FHC Inc. ). A small portion of the dura mater was removed to facilitate the penetrations. All other details are as in ref. 46. We recorded from199 single neurons but discarded 44 of them before analysing the data because of technical problems such as noise in the eye coil signal or instability of the recording throughout the session (39 neurons) or poor fixation performance (five neurons).
We present data from 155 neurons, 80 from monkey H (57 foveal/parafoveal neurons and 23 peripheral neurons) and 108 from monkey Y (19 foveal/parafoveal neurons and 56 peripheral neurons), with RF eccentricities ranging from 0. 2 to 35 degrees of visual angle. Experimental design. Monkeys were contained in a dark box during the experiments. They fixated a small red cross (0. 5-deg width) on a cathode ray tube (CRT) video monitor (BarcoReference Calibrator V, 120 Hz refresh rate) placed 57 cm away from their eyes and received fruit juice rewards approximately every 1. 5-2 s when properly fixated. Excursions of gaze outside of an invisible 2 Â 2-deg fixation window were recorded but not rewarded. Monkeys kept their gaze within the fixation window for 94% of the experimental time. We note that we used a relatively large fixation window so as to avoid fixation overtraining, and thus allow the monkeys to naturally produce sufficient numbers of microsaccades of varied sizes.