dataesr/scientific-paragraphs-categorization huggingface dataset:
Patient population Validated ICD-10-CM codes were used to identify patients in the SNDS database. 20  Adults with an RA diagnosis (ICD-10-CM codes M05 (RA with rheumatoid factor), M06. 0 (RA without rheumatoid factor), M06. 8 (other specified RA) and M06. 9 (RA unspecified)) and ≥2 distinct delivery dates of disease-modifying antirheumatic drugs (DMARDs) were included. DMARDs included were abatacept, adalimumab, anakinra, azathioprine, baricitinib, certolizumab pegol, cyclophosphamide, ciclosporin, etanercept, golimumab, hydroxychloroquine, infliximab, leflunomide, methotrexate (MTX), rituximab, sulfasalazine, tocilizumab and tofacitinib. Onset of RA was defined as the first date of occurrence between RA diagnosis and the first known DMARD reimbursement. For patients with RA-ILD, ILD diagnosis was defined using the ICD-10-CM code J84 (other interstitial pulmonary diseases) and as having ≥1 CT scan during the observation period.
Rheumatoid arthritis Rheumatoid arthritis Rheumatoid arthritis which is similar to what was found in a recent Danish national healthcare database and other American health insurance registry studies. 10 21 22  Variation among previous estimates using administrative claims to identify patients with RA-ILD is by definition limited to the population of patients meeting the criteria for clinically significant ILD, and concomitantly excludes patients with subclinical RA-ILD. Nonetheless, this approach may be relevant to evaluate the disease burden of RA-ILD, as it considers the most significant cases of ILD. In previous decades, progress in RA treatment led to a decrease in mortality due to comorbidities, such as cardiovascular events, in patients with RA. 18  However, in a recent American epidemiological study, while RA-related overall mortality rates decreased from 2005 to 2018, RA-ILD-related mortality rates remained stable over the same period, suggesting that improvement of overall outcomes in RA had limited effect in patients with RA-ILD.
23  The increased mortality rate observed in our study is supported by other healthcare claims database analyses showing significantly increased mortality rates in these patients. 10 21 24 25  However, we did have some novel findings. In the current study, the difference in mortality rate between patients with RA-ILD and those with RA-n-oILD was greatest in patients aged <75 years, in patients who developed ILD prior to RA onset, in male patients and in patients with cardiometabolic risk factors. The excess mortality observed in patients with ILD prior to RA may correspond to a more advanced, and therefore more severe, lung disease. Regarding the finding of a greater difference in mortality rate in patients aged <75 years, the burden of RA-ILD appears to be higher in young patients free from other comorbidities.
These findings align with the perception that some DMARDs, notably MTX, may exert a potentially harmful impact on patients with RA-ILD. 29  Regarding MTX, this perception contradicts recent evidence demonstrating the lack of contribution of MTX to RA-ILD occurrence and progression. [30] [31] [32] [33] [34]  Besides, recent studies have identified an association between high RA disease activity and mortality in patients with RA-ILD, meaning that inappropriate cessation of the recommended first-line DMARD could strongly impact prognosis in such patients. 18 33  Nevertheless, future studies identifying how specific treatments may affect mortality rates among patients with RA-ILD will better inform clinical decision-making for these patients. This current study has limitations. First, a healthcare claims database is not adequate to assess preclinical RA-ILD, leading to a potential misclassification bias.
Second, the absence of information about HRCT pattern data did not allow for detailed analyses identifying potential characteristics specific to a given RA-ILD pattern. Third, the limited period of analysis (January 2013 to December 2018) did not enable comparative analyses over time. Additionally, data not collected by the SNDS database, such as smoking status, limited the ability to examine alternative variables that may impact mortality rates among patients with RA-ILD. Last, the treatment modalities reported herein reflect everyday clinical practice in France and may not be generalisable to other countries. However, the current study contained notable strengths, such as the size of the overall cohort, which comprised nearly all of the French population. Additionally, congruent with previous work, we used validated diagnostic codes for both RA and ILD stratification, reinforcing the reliability of our findings.
Acknowledgements Professional medical writing and editorial assistance was provided by  Claire Line, PhD , and  Ryan Miller, PhD , at  Caudex , a  division of IPG Health Medical Communications , and was funded by  Bristol Myers Squibb. Contributors Study conception and design: P-AJ, GD, JZ, VV-M, BC, R-MF, PD. Acquisition of data: GD, PD. Data analysis: GD, PD. Data interpretation: P-AJ, LW, SO, GD, JZ, VV-M, R-MF, BC, PD. All authors were involved in drafting the article or revising it critically for important intellectual content, and all authors approved the final version to be published. PD is the guarantor for this manuscript. Funding This study was supported by  Bristol Myers Squibb. Competing interests PA-J has received grant/research support from  AstraZeneca ,  Boehringer Ingelheim ,  Bristol Myers Squibb ,  Medac ,  Novartis ,  Roche Chugai  and  Societe Francaise de Rhumatologie ; and consultancy fees from  Bristol Myers Squibb.
Research in context Evidence before this study We searched PubMed for all English language publications from database inception to Dec 31, 2022, using the terms: ("Hepatitis B" OR "HBV") AND ("peripartum treatment" OR "PPT" or "peripartum antiretroviral prophylaxis" OR "peripartum antiviral prophylaxis" OR "PAP" OR "peripartum prophylaxis"). WHO has set a goal to reach less than 0•1% prevalence of chronic hepatitis B virus (HBV) infection in children younger than 5 years by 2030, which will require substantial efforts in reducing HBV mother-to-child transmission-now the major route of ongoing transmission. Building on the longstanding WHO recommendation of universal neonatal and infant vaccination, the first guidelines for the addition of peripartum antiviral prophylaxis (PAP) in pregnancy were released in 2020.
Model In this modelling study, we adapted de Villiers and colleagues'  12  dynamic simulation model of the global HBV epidemic structured by age, sex, and country. 110 countries that collectively represent a large share of the burden of HBV (appendix 1 p 12) were each modelled independently by use of national demographic data on fertility, mortality, and population structure and national intervention coverage levels of infant vaccination (HepB3) and HepB-BD vaccination (appendix 1 p 3). Transmission and natural history progression rates were parameterised following reviews of the literature (appendix 1 pp 2-3). In the model, transmission occurred from mother to child, from child to child, and across the whole population, and the relative strength of each mode of transmission was inferred through a calibration procedure that fitted the model to data on HBsAg prevalence, the proportion of women of childbearing age who were HBsAg-positive, and HBV-related deaths (appendix 1 p 8).
When comparisons were made against a cost-effectiveness threshold, we used the country-specific values for health opportunity costs as calculated by Ochalek and colleagues. 21 Cost-effectiveness thresholds were calculated for 106 (96%) of the 110 countries, as there were insufficient data to calculate thresholds for North Korea, Somalia, Syria, and Yemen (appendix 1 p 12). All newborns are eligible for both HepB-BD and HepB3, irrespective of maternal HBsAg status. For the PAP strategies, PAP is accompanied by antenatal screening and it is assumed in the main analysis that each pregnant woman is tested only once (on their first pregnancy) as part of routine antenatal care. HBV=hepatitis B virus. HepB-BD=hepatitis B birth-dose vaccination. HepB3=hepatitis B infant vaccination. PAP=peripartum antiviral prophylaxis.
Acknowledgments SN, TBH, and MJdV acknowledge support from the  MRC Centre for Global Infectious Disease Analysis , based at  Imperial College London, London, UK ; this award is jointly funded by the  UK MRC  and the  UK Department for International Development (DFID)  under the  MRC-DFID  Concordat agreement and is also part of the  Second European & Developing Countries Clinical Trials Partnership programme  supported by the  EU. SN is also supported by the  UK National Institute for Health and Care Research Imperial Biomedical Research Centre.
Context: Retirement is an opportune time for people to establish new healthy routines. Exercise and nutritional interventions are promising in the prevention and treatment of sarcopenic obesity. Objective: This systematic review aimed to assess the effectiveness of nutritional and exercise interventions for the treatment of sarcopenic obesity in persons of retirement age. Data Sources: PubMed, Embase, CINAHL, and CENTRAL databases were searched in September 2021 for randomized controlled trials; a manual search was also conducted. The search yielded 261 studies, of which 11 were eligible for inclusion. Data Extraction: Studies of community-dwelling individuals with sarcopenic obesity receiving any nutritional or exercise intervention ! 8 weeks with the mean age 6 standard deviation between 50 and 70 years were included.
METHODS To answer the research question, a systematic review of randomized controlled trials (RCTs) was performed, and meta-analyses were conducted where possible. This systematic review was registered in the PROSPERO database (no. CRD42021276461). The review protocol is available at https://www. crd. york. ac. uk/prospero/. No substantial methodological changes were made to the registered protocol. This systematic review was reported according to the 2020 PRISMA (Preferred Reporting Items for Systematic Reviews and Meta-Analyses) guidelines. 25 The PRISMA checklist is available as Table  S1  and the PRISMA 2020 checklist as Table  S2  in the Supporting Information online.
Search strategy A comprehensive literature search of the following databases was conducted to identify RCTs published up to September 16, 2021: PubMed (MEDLINE) from 1946 onward, Embase via OVID from 1974 onward, CINAHL via EBSCOhost, and the Cochrane Central Register of Controlled Trails (CENTRAL) in the Cochrane Library. Two trial registers (clinicaltrials. gov and the WHO International Clinical Trials Registry Platform) were also searched, and the reference lists of eligible articles (ie, snowballing), as well as Google Scholar, were checked to identify additional potentially relevant studies. The literature review was conducted by two authors independently. There were no restrictions on language. The following search terms were used to identify relevant studies: "obesity," "sarcopenia," "sarcopenic obesity," "utr*," "oral nutritional supplements," "diet," "train*," "physical activity," "exercise," and the MeSH terms "diet, food, and nutrition," "nutrition therapy," "diet," and "exercise.
Study selection Study selection was carried out with the Covidence systematic review software (Veritas Health Innovation; Melbourne, Australia). Title and abstract screening as well as full-text screening was conducted by two authors independently (D. E. , M. T, S. B. ). In case of conflicts, a consensus was reached by discussion and, if necessary, a third author was consulted to make a final decision (D. E. ). Articles were included on the basis of predefined PICOS (Participants, Intervention, Control, Outcomes, Study design) criteria (Table  1 ). Conference abstracts were excluded. Primary endpoints were body composition (eg, body fat mass, muscle mass, lean mass). Secondary endpoints were body mass index, muscle strength, and physical function (eg, gait speed). Studies in which any pharmaceutical agent was administered were excluded.
Data extraction and quality assessment For the purpose of data extraction, a standardized data extraction template, which included study characteristics, patient characteristics, and outcomes, was used. Two independent reviewers (D. E. , M. T. ) extracted data from the full-text articles that were included. The methodological quality of the RCTs was assessed by two authors independently, using the Cochrane Risk of Bias Tool. 26 Use of this tool enabled the following potential sources of bias to be evaluated: sequence generation; allocation concealment; blinding of participants, personnel, and outcome assessment; incomplete outcome data; selective outcome reporting; and other sources of bias.
Data synthesis and data analysis The data synthesis and data analysis processes in this study were carried out as recommended in the Cochrane Handbook for Systematic Reviews of Interventions. 26 Comparable studies were grouped on the basis of the type of intervention. Meta-analysis was conducted when at least two comparable studies were identified for the respective intervention and outcome. An intention-to-treat analysis of the primary studies was preferred, if available. If there were considerably different measurement methods for one outcome, subgroup analyses were performed. For interventions for which data could not be pooled, results were summarized narratively. The statistical analysis was conducted with Review Manager software (version 5. 4. 1; Cochrane, Oxford, UK). The primary outcome was the mean difference (MD) in postintervention body composition (fat mass and muscle mass) between the intervention group and the control group.
Continuous outcome data were presented as mean differences with standard deviations (SDs) (ie, intervention group minus control group) and 95% confidence intervals. For trials with more than one control group, the control group without any intervention was included. Random-effects models were used to calculate the overall estimated weighted mean differences. The statistical heterogeneity of the included studies was assessed using the X 2 test (statistical significance was set at P < 0. 05) and I 2 statistics. An I 2 value up to 25% was assumed to indicate low heterogeneity, a value up to 50%, moderate heterogeneity, and a value more than 75%, high heterogeneity among the studies. 27 Reporting bias was evaluated by funnel plots if more than 10 studies were available.
Meta-analysis for the effect of resistance training [32] 36, 38  Frequency of the training was 2 to 3 times a week, and the duration ranged from 20 to 60 minutes per session. The results of the meta-analysis for all outcomes are shown in Figure  3. [29] [30] [31] [32] 36, 38  Percent body fat. Four  dies with a total of 198 participants reported results for the effects of resistance training on percent body fat; therefore, these could be pooled. [29] [30] [31] [32] 36  The meta-analysis results reveal a significant reduction in percent body fat (MD À1. 53%; 95%CI, À2. 91 to À0. 15) through resistance training. The I 2 of the data was 28%, indicating a moderate level of heterogeneity between the studies. Muscle mass. Muscle mass in kilograms was reported in 3 studies with a total of 136 participants, and muscle mass as a percentage of body weight was reported in 2 studies with 88 participants.
For example, they tend to have different body fat content and increased abdominal adiposity compared with Caucasians. 56 The generalizability of the results of the individual studies, therefore, is limited, and any application of the results to other ethnic groups should be done with caution. Second, the authors of the included studies used different definitions for both sarcopenia and obesity, which may have reduced the comparability of results. Third, even though a comprehensive search in several databases and trial registers was conducted, it is possible that some studies may have been overlooked.
Acknowledgment Author contributions. D. E. contributed to project administration, study conception, data extraction, literature screening, data interpretation and writing of the original draft. S. B. contributed to project administration, study conception, data analysis, and writing, critical reviewing, and editing of the original draft. M. T. contributed to the literature search, data extraction, data interpretation, and writing of the original draft. L. R. contributed to critical reviewing and editing of the original draft. J. D. S. , P. M. J. W. , T. V. , Y. B. , and A. J. C-J. contributed to study conception and critical reviewing and editing of the original draft. Funding. The SO-NUTS (preventing obesity, sarcopenia, and Sarcopenic Obesity in retirement through.
A search for a new heavy scalar or pseudo-scalar Higgs boson (H/A) produced in association with a pair of top quarks, with the Higgs boson decaying into a pair of top quarks (H/A → t t) is reported. The search targets a final state with exactly two leptons with same-sign electric charges or at least three leptons. The analysed dataset corresponds to an integrated luminosity of 139 fb -1 of proton-proton collisions collected at a centre-ofmass energy of 13 TeV with the ATLAS detector at the LHC. Two multivariate classifiers are used to separate the signal from the background. No significant excess of events over the Standard Model expectation is observed. The results are interpreted in the context of a type-II two-Higgs-doublet model. The observed (expected) upper limits at 95% confidence level on the t tH/A production cross-section times the branching ratio of H/A → t t range between 14 (10) fb and 6 (5) fb for a heavy Higgs boson with mass between 400 GeV and 1000 GeV, respectively.
This particular signature is experimentally favoured due to the low level of background contamination, with the main contribution originating from the SM production of four top quarks and of t t in association with a W boson (t tW ), Z boson (t tZ), or Higgs boson (t tH). Other significant sources of background are events where one of the leptons has a mis-assigned charge and events which contain a fake/non-prompt lepton. Backgrounds from multiboson, t tW W , single-top-quark or other rare top-quark processes are expected to be minor. The analysed dataset corresponds to an integrated luminosity of 139 fb -1 of proton-proton (pp) collisions collected at a centre-of-mass energy of √ s =13 TeV with the ATLAS detector at the LHC. Previous searches for ttH/A(→ tt) in the SSML channel were performed by the AT-LAS and CMS Collaborations  [29, 30].
JHEP07(2023)203 The results obtained in this search are interpreted in the context of a type-II 2HDM; upper limits are placed on the ttH/A production cross-section times the branching ratio of H/A → t t as well as on tan β as a function of the heavy Higgs boson's mass. Due to a significantly larger dataset and improved analysis techniques, including the use of multivariate classifiers based on boosted decision trees (BDTs) to distinguish the signal from the SM background, the expected sensitivity of the present ttH/A(→ tt) search exceeds that of the previous ATLAS ttH/A(→ tt) search with 36 fb -1  [29]  by about a factor of four. This paper is structured as follows. The ATLAS detector is presented in section 2. The data and samples of simulated events are described in section 3. The object and event selections are detailed in section 4.
Data and samples of simulated events This analysis uses pp collision data collected between 2015 and 2018 by the ATLAS detector at √ s = 13 TeV. The full dataset corresponds to an integrated luminosity of 139 fb -1. Only data recorded during stable beam conditions in which all detector subsystems were operational  [40]  are included. Events were collected using single-lepton or dilepton triggers. The trigger p T threshold depended on the lepton flavour and the data-taking period  [41, 42].
Experimental uncertainties The uncertainty in the integrated luminosity of the combined 2015-2018 dataset is 1. 7%  [101] , obtained using the LUCID-2 detector  [102]  for the primary luminosity measurements. The uncertainties related to the reweighting factors that correct the pile-up profile in simulations to match the one in data are also included. Uncertainties in the modelling of leptons arise from their momentum resolution and scale, as well as the trigger, reconstruction, identification and isolation efficiencies used to correct for the difference between the simulation and data  [41, 42, 85, 86]. Uncertainties in the modelling of jets come from their energy scale and resolution, containing the effect of jet flavour composition, single-particle response, and pile-up  [90]. The uncertainty in the efficiency to pass the JVT requirement for pile-up suppression is also applied  [103].
The test statistic q µ is evaluated with the RooFit package  [109, 110]. A related test statistic is used to determine whether the observed data is compatible with the background-only hypothesis by setting µ = 0 in the profile likelihood ratio and leaving μ unconstrained: q 0 = -2 ln(L(0, θ0 , λ0 )/L(μ, θ, λ)). The p-value representing the level of agreement between the data and the backgroundonly hypothesis is estimated by integrating the distribution of q 0 , based on the asymptotic formulae in ref. [111] , above the q 0 value observed in the data. Upper limits on µ are derived by using q µ in the CL s method  [112, 113]. For a given signal scenario, values of µ yielding CL s < 0. 05, where CL s is computed using the asymptotic approximation  [111] , are excluded at ≥95% confidence level (CL).
In addition, individual groups and members have received support from  BCKDF ,  CANARIE ,  Compute Canada  and  CRC ,  Canada ;  PRIMUS 21/SCI/017  and  UNCE SCI/013 , Czech Republic; COST,  ERC ,  ERDF ,  Horizon 2020  and  Marie Skłodowska-Curie Actions ,  European Union ;  Investissements d'Avenir Labex, Investissements d'Avenir Idex  and  ANR, France ;  DFG and AvH Foundation ,  Germany; Herakleitos ,  Thales  and  Aristeia  programmes co-financed by  EU-ESF  and the  Greek NSRF, Greece; BSF-NSF  and MINERVA, Israel; Norwegian Financial Mechanism 2014-2021, Norway;  NCN  and  NAWA, Poland ;  La Caixa Banking Foundation ,  CERCA Programme Generalitat de Catalunya and PROMETEO and GenT Programmes Generalitat Valenciana ,  JHEP07(2023 )203  Département de Physique Nucléaire et Corpusculaire ,  Université de Genève, Genève; Switzerland (a) Dipartimento di Fisica, Università di Genova, Genova; (b) INFN Sezione di Genova; Italy II.
Acknowledgements The research leading to the results detailed in this manuscript received funding from the  Fondation pour la Recherche Médicale (MJ)  and the  CPER-FEDER program  (MJ),  FRC CONNECT (SL)  and  Région Nouvelle Aquitaine  (SL and MJ,  2019-1R3M08 ). VT is a recipient of a joint PhD fellowship from  Nouvelle-Aquitaine Region/Fondation pour la Recherche Médicale. MT is a recipient of a joint PhD fellowship from  Nouvelle-Aquitaine Region/Inserm. This work has benefited from the facilities and expertise of the  PREBIOS platform (Université de Poitiers ). Authors are grateful for the research staff at the  Eye and Nutrition Research Group  for lipidomic analyses ( UMR1324   INRAE, Dijon, France). Authors are thankful to  Jeffrey Arsham  and  Nathaniel L. Ritz  for careful language editing.
Data availability Raw data for gut microbiota analysis are available on the following site (https:// www. ncbi. nlm. nih. gov/ biopr oject/ PRJNA 854167). Accession SRA# are from SRR 19906240 to SRR19906299. Detailed statistical analysis is included in the supplementary material (Suppl. Tables  S3-S15 ). Other details and raw data of all other experiments can be obtained upon reasonable request to the authors.