dataesr/scientific-paragraphs-categorization huggingface dataset:
The total ion dose was 4. 28 x 10  16  ions/cm 2. The total ion image acquisition time was approximately 9 hrs. The single-cell drug accumulation and lipid upregulation (Fig. 6d ) correlation calculations were performed in OriginPro software (2015; OriginLab, MA, USA). The correlation test was performed using a Pearson's (2-tailed) correlation test at the 95% confidence level.
ACKNOWLEDGEMENTS: The authors thank  N. Harrison ,  R. Reid  and  A. Harling  and  M. Skingle  for their support during the project and  T. Heller ,  M. Krehl ,  A. Dütting ,  P. Hörster ,  K. Strupat ,  S. Möhring ,  F. Czemper ,  A. Venckus ,.
Data availability The data that support the findings of this study are available in figshare with the identifier 10. 6084/m9. figshare. 5459680. 45  We provide the Orbitrap data in the mzML format for spectra and the centroided ImzML format for images. The ToF image in Figure  2  is in the GRD data (TOF-SIMS generic raw data) format. Data may be opened using Spectral Analysis software 46 available from Github: https://github. com/AlanRace/SpectralAnalysis.
Patients and data collection We evaluated adverse events in 1,241 video-EEG sessions of 998 patients in two groups: group 1 includes 507 consecutive patients with 596 video-EEG sessions. This group was examined retrospectively from the EMU of the Department of Neurology, Medical University, Innsbruck, Austria, between January 1999 and December 2005. The video-EEG investigation was performed without a specific safety protocol ("preimplementation"). The detailed procedures and results have been reported previously. 8 Group 2 includes 491 consecutive patients with 645 video-EEG sessions. These patients were evaluated retrospectively at the Department of Neurology, Paracelsus Medical University, Salzburg, Austria, between January 2013 and September 2015 ("postimplementation"). Based on the results of our previous work, we implemented specific safety measures in group 2 (described in detail below).
Experimental protocol Sixteen to 18 h after completion of the nerve electrode implantation, the arterial catheter and the nerve electrode were connected to the recording system  (Bertram et al. 2005). The renal sympathetic nerve activity (RSNA) signal was amplified (950,000) and bandpass filtered at 300-3,000 Hz. This raw signal was directly sampled at 10,000 Hz. It was also full-wave rectified and low-pass filtered at 150 Hz by an analog custom-made rectifier before being sampled at 5000 Hz  (Chapuis et al. 2010 ). The AP signal was sampled at 500 Hz. Data acquisition was performed using LabVIEW 5. 1 software (National Instruments, Austin, TX). Throughout the experiment, cardiac autonomic blockade was achieved with the combined administration of methylatropine and atenolol (2 mg/kg each followed by 2 mg/kg/h iv).
Data analysis AP and rectified RSNA time series were resampled at 1 Hz by averaging data over consecutive 1-s periods. One-s average values of AP are referred to as MAP values in the following of the text. Within each 1-s segment, heart rate (HR) was calculated from the mean interbeat interval. The background noise level of RSNA was taken as the mean RSNA value calculated over a 2-min period following chlorisondamine. This value was subtracted from all RSNA data for subsequent analyses. Under each condition (Fig. 1 ), i. e. , baseline (before LPS administration), LPS (1 h after LPS administration), LPS + SAL (after LPS administration plus 1 h after starting saline infusion), LPS + DEX (after LPS administration plus 1 h after starting dexmedetomidine infusion), representative MAP, HR and RSNA values were estimated as the 10-min average values preceding a dose-response curve to phenylephrine.
Peak increases in MAP induced by phenylephrine were calculated as the maximum changes of MAP from its value recorded immediately before injection. These pressor responses were plotted against the log 10 of the corresponding phenylephrine dose. Because the MAP phenylephrine dose-response relationship frequently showed asymmetry of curvature, we systematically fitted a five-parameter logistic equation to the data (SigmaPlot 2000; Systat Software, San Jose, CA). The equation actually reduces to a four-parameter equation because it is forced to zero (i. e. , there is no MAP change when the phenylephrine dose is null). The equation is therefore y ¼ a ð1 þ e À xÀx 0 b Þ c where y is the MAP change, x is the log 10 of the phenylephrine dose, a is the plateau of the MAP response (maximum effect), b and c are curvature coefficients, and x 0 ¼ ED 50 þ bLnð2 1 c À 1Þ where ED 50 is the log 10 of the half-maximal effective dose and provides the sensitivity of the response after inverse log transformation (it is then expressed in lg/kg).
Meteorological data We used temperature and precipitation variables to estimate meteorological conditions during migration stopover sites, as well as at breeding and wintering grounds. Furthermore, as the winter North Atlantic Oscillation (NAO) is strongly correlated with spring conditions in Europe, we used this index (following, for example,  Gordo and Sanz 2008; Salewski et al. 2013)  as a variable in one of our models to analyse the effect of weather conditions on survival along the migratory route. Based on recoveries and published data available for the species we considered August-October to be the time of autumn migrations; October-March as the period spent at wintering grounds; and March-May as the time of spring migrations  (Cramp 1992; Korner-Nievergelt et al. 2002; Balanca and Schaub 2005; Prochazka et al.
2008; Zwarts et al. 2014 ). The periods overlap as breeding, arrival and departure of individuals of the species are asynchronous  (Halupka et al. 2014b). For the analyses, we used mean temperatures and total precipitation during the given seasons. Meteorological data for the breeding season were collected at the study site, approximately 800 m from the study plot (Davis Vantage PRO 2 weather station; Davis Instruments, Hayward, CA). Meteorological data for the wintering grounds were obtained from the National Oceanic and Atmospheric Administration's (NOAA) National Centers for Environmental Information (data available online: http://www. ncdc. noaa. gov/). NAO values were obtained from NOAA's National Weather Service, Climate Prediction Center (data available online: http://www. cpc. ncep.
949, df = 26, p = 0. 002; c-hat = 1. 805) and was a result of issues with overdispersion revealed by test 3. SR (p \ 0. 001). These problems were accounted for by implementing a time- since-marking (TSM) model / 2M-t/t p g9t , where two marking classes are present, with the first (M1) class representing survival of individuals in the first year after marking, and the second class (M2?) representing survival in subsequent years. C-hat values were estimated to evaluate whether problems with overdispersion were still present following the structural adjustment of the model. This was measured by dividing the GOF V 2 value of the relevant tests from U-CARE by the degrees of freedom  (Cooch and White 2012). The adjusted (TSM) model accounted for issues in the dataset, therefore meeting assumption 2 of the CJS model.
Variable sensitivity of different habitats to rainfall levels may explain the inconsistent relationships between survival and rainfall in Africa that were found for different bird species  (Boano et al. 2004; Salewski et al. 2013; Ockendon et al. 2014; Johnston et al. 2016). Our results are also in line with the earlier findings of  Thaxter et al. (2006)  and  Zwarts et al. (2009) , who did not find a significant effect of African droughts on the size of Reed Warbler populations. However, our estimates are based on a 9-year dataset. Obtaining a larger dataset would allow for more extensive analyses examining the relationship between survival and meteorological conditions. We did not find a significant effect of weather variables during the breeding season on the survival of adult Reed Warblers.
Acknowledgements We are grateful to  Giulia Cassasole ,  Justyna Chachulska ,  Monika Czuchra ,  Beata Czy_  z,  Alicja Dziachan ,  Stanisław Rusiecki ,  Magdalena Sobon ´ ,  Łukasz Tomasik ,  Paulina Turowicz  and  Katarzyna Turzan  ´ska for their help in the field. We would like to thank to  Tomasz Mokwa  from the  Polish Ringing Centre  for the access to the data associated with migratory routes and wintering sites of Reed Warblers. We are indebted to  Richard Broughton  for constructive comments on the manuscript and for improving the English, and to  Grzegorz Neubauer  for statistical advice. We are also grateful to  Michael Schaub  and two anonymous reviewers whose comments considerably improved the manuscript. The research was financially supported by the  Polish State Committee for Scientific Research  (Grant  2P04F05330 ) and the  University of Wrocław  ( Projects 1018/S/IZ/2010-11  and  1073/S/SORM/2012-13 ).
Study design and population We conducted a prospective, cross-sectional study between  April 1, 2008 and December 31, 2009  in five high-volume referral centers in France  [7]. All 28 endocrine surgeons who surgeons were not informed and did not consent about the eventuality of sharing their personal data outside the study perimeter at a global level. Finally, the dataset analyzed for the purpose of this manuscript contains sensitive information on both patients, surgeons and centers, which could be easily unanonymized. Because the Lyon university hospital was the only structure allowed to store the data in accordance with French regulation, data requests should be sent to Prof. Jean-Christophe Lifante (jean-christophe. lifante@chu-lyon. fr) or Dr Antoine Duclos (antoine. duclos@chu-lyon. fr).
Competing interests: The authors have declared that no competing interests exist. performed thyroid surgery at these centers participated in the study. All patients who underwent a thyroid procedure were eligible for inclusion. The extent of inclusion was measured in relation to the number of eligible thyroidectomies recorded in the hospital administrative databases. This study was approved by the Research Committee for the Protection of Persons (CPP) and the National Advisory Committee on Information Processing in Material Research in the Field of Health (CCTIRS) France, in accordance with French ethical directives. Informed consent was obtained from participating surgeons. The ethics committee waived the requirement for patient consent. Before surgery, patients received written information about personal data use and gave verbal consent for sharing their data.
Outcome measures and data collection The primary outcome was operative time, measured in minutes (min) and defined as the total duration from skin incision to closure of the wound. After each thyroidectomy, a patient report form was completed by the attending surgeon, including items about surgical indication and procedure, as well as the surgeon's identity, the presence of a more experienced supervisory surgeon during the intervention, and the number of surgical procedures performed by that surgeon the same day. Research assistants completed data collection using medical records. These data included patient demographics, information on previous thyroid surgeries, and weight of the thyroid specimens. Variables were divided into two groups: preoperative and intraoperative. Preoperative variables corresponded to information available before the operation.
Statistical analysis Data analyses were performed using SAS and R softwares. Operative time was described using means and standard deviation and compared using the Mann-Whitney test or Kruskal-Wallis test, according to the number of groups. All tests were two tailed, and P < 0. 05 was considered statistically significant. To explore the determinants of operative time, we used a multilevel linear regression model in which patients were nested in surgeons, who in turn were nested in hospitals  [8]. In this model, we classified center variables as a random effect, surgeon variables as both random and fixed effects, and all patient variables as fixed effects. A log transformation on operative time was performed to meet normality, which was successfully checked by using residuals and quantile-quantile plot.
Results were presented in the original scale, as differences compared to the intercept with corresponding 95% confidence intervals (CI) calculated with the function deltamethod of the msm R-package  [9]. To estimate the relative importance of variables to operative time, a multiple linear model was then performed including patients, surgeons, and center variables. The LMG metric via the R-package RELAIMPO was used to assess the relative importance of each variable and of groups of variables by decomposing the overall R 2 into nonnegative R 2 values for each variable and for each group of variables  [10]. Relative importance estimates were then adjusted to sum to 100% for easier interpretation. Confidence intervals were computed via the same R-package with the percentile interval method of bootstrapping based on 1000 replicates, at a 99% confidence level to adjust for multiple comparisons and because these bootstrap CI can be somewhat liberal.
On the contrary, our study focused on a narrower population and recognizes the surgeon's individual experience as a key driver of performance focused on operative time from skin incision to closure. According to the "practice makes perfect" dogma, there is strong evidence that increasing surgical case volume and years of practice are associated with improved outcomes in a procedure-specific manner  [12, 13]. Following an initial learning curve, several studies noted a plateau phase, where increases in case volume or years of practice were no longer associated with improvements in surgical outcomes  [14]. Based on the same dataset used in the present study, we have shown that the plateau for thyroidectomy spanned from 5 to 20 years of a surgeon's practice with regard to both postoperative recurrent laryngeal nerve palsy and hypoparathyroidism  [7].
The strengths of our study include it's a priori design to model the nature of the association between surgeons' experience and performance in thyroid surgery. Because patient recruitment and data were recorded prospectively with great care, we avoided coding bias resulting from secondary utilization of administrative databases  [18]. Based on the accuracy of the data, the patient features and other surgeon factors were partly controlled, and we considered the clustering of patients within surgeon and hospital center levels. However, this work also has several limitations. First, the applicability of our results to other surgical fields is questionable, in view of the limited sample size of endocrine surgeons in academic referral centers. Thyroidectomy is considered a short procedure with quick access to the thyroid site and limited uncertainty in its duration.
Acknowledgments Members of CATHY Study Group:  Laurent Arnalsteen ,  Robert Caizzo ,  Bruno Carnaille ,  Guelareh Dezfoulian ,  Carole Eberle ,  Ziad El Khatib ,  Emmanuel Fernandez ,  Antoine Lamblin ,  Francois Pattou ,  Marie-France Six  (Lille);  Stephanie Bourdy ,  Laetitia Bouveret ,  Francois Chollet ,  Cyrille Colin ,  Ronald Daher ,  Antoine Duclos ,  Benoı ˆt Guibert, Nathalie Laplace, Marie-Annick Le Pogam, Jean-Christophe Lifante, Arnaud Patoir, Cecile Payet, Jean-Louis Peix, Stephanie Polazzi, Gaetan Singier ,  Pietro Soardo ,  Sandrine Touzet ,  Nicolas Voirin  (Lyon);  Pascal Auquier ,  Jean-Francois Henry, Claire Morando, Frederic Sebag, Sam Van Slycke (Marseille) ;  Inès Akrout ,  Fares Benmiloud ,  Jean-Paul Chigot ,  Isabelle Colombet , Gae ¨lle  Godiris-Petit ,  Pierre Leyre ,  Fabrice Menegaux, Severine Noullet, Benoı ˆt Royer, Christophe Tresallet (Paris );  Thibault Desurmont ,  Claudia Dominguez ,  Jean-Louis Kraimps ,  Chiara Odasso ,  Laetitia Rouleau  (Poitiers);  Yves-Louis Chapuis ,  Pierre Durieux ,  Alain Lepape ,  Frederic Triponez (Scientific Committee ).
Numerical modeling For the various PDFs, we neglect any QCD evolution (in practice, we take a fixed factorization scale µ 2 F = 10 GeV 2 ) and we use the following models: PSfrag replacements x H u(-) (x, ξ) H d(-) (x, ξ) Figure  2 : Models for the GPDs H u(-) and H d(-) for ξ =. 1, a value corresponding to S γN = 20 GeV 2 and M 2 γρ = 3. 5 GeV 2. The various curves differ with respect to the ansätze for the PDFs q: GRV-98 (solid black), MSTW2008lo (long-dashed blue), MSTW2008nnlo (short-dashed red), ABM11nnlo (dotted-dashed green), CT10nnlo (dotted brown). Note that the two GPDs H d(-) based on these two last ansätze are hardly distinguishable. • For xq(x), we rely on the GRV-98 parameterization  [30] , as made available from the Durham database. To evaluate the uncertainty of this parameterization, we also consider a few other sets of PDFs, namely MSTW2008lo and MSTW2008nnlo  [31] , ABM11nnlo  [32] , CT10nnlo  [33].
In Fig. 2 , we show the resulting GPDs H u(-) and H d(-) for ξ =. 1 corresponding in our process to the typical value S γN = 20 GeV 2 and M 2 γρ = 3. 5 GeV 2. • For x∆q(x) , we rely on the GRSV-2000 parameterization  [34] , as made available from the Durham database. Two scenarios are proposed in this parameterization: the "standard", i. e. with flavor-symmetric light sea quark and antiquark distributions, and the "valence" scenario with a completely flavor-asymmetric light sea densities. We use both of them in order to evaluate the order of magnitude of the theoretical uncertainty. In Fig. 3 , we show the resulting GPDs Hu(-) and Hu(-) for ξ =. 1 corresponding in our process to the typical value S γN = 20 GeV 2 and M 2 γρ = 3. 5 GeV 2. • For xδq(x) we rely on a parameterization performed for TMDs (based on a fit of azimuthal asymmetries in semi-inclusive deep inelastic scattering), from which the transversity PDFs xδq(x) are obtained as a limiting case  [35].
Acknowledgements We acknowledge several useful discussions with  Mauro Anselmino ,  Marie Boër ,  Goran Duplančić ,  Michel Guidal ,  Stefano Melis ,  Hervé Moutarde ,  Stéphane Munier ,  Kornelija Passek-Kumerički ,  Franck Sabatié  and  Daniel Tapia-Takaki. This work is partly supported by grant  No 2015/17/B/ST2/01838  by the National Science Center in Poland, by the  French grant ANR PARTONS  (Grant No. ANR-12-MONU-0008-01 ), by the  COPIN-IN2P3  agreement, by the  Labex   P2IO  and by the  Polish-French collaboration agreement Polonium.
Although this can be achieved using cpDNA sequences of target and/or related species deposited in public databases, it may be necessary to perform de novo sequencing of chloroplast genomes in species for which no genomic information is available  (Ebert and Peakall, 2009). High-throughput sequencing of total genomic DNA allows for the recovery and assembly of complete chloroplast genomes  (Daniell et al. , 2016) , although most reads in these situations are derived from the nuclear or mitochondrial genomes and contaminating DNA. For example, cpDNA reads represented only 1. 2% of all reads for the conifer Abies nephrolepis (Trautv. ex Maxim. ) Maxim. (.
Characterization of the chloroplast genome sequence of C. sulcata and SSR marker development-To verify the quality of cpDNA assembly by MITObim, gene annotation and synteny analysis were performed for C. sulcata as a case analysis. Gene annotation was performed with the aid of the CPGAVAS AnnoGenome module  (Liu et al. , 2012)  with a cutoff BLASTN E-value of 1e-10. The reference species for annotation was Cryptomeria japonica (GenBank accession no. AP009377). The annotated chloroplast genome was visualized as a circular genome using GenomeVx  (Conant and Wolfe, 2008). PipMaker  (Schwartz et al. , 2000)  was used to create dot plots and gene identity plots between C. sulcata and another Cupressaceae species, Calocedrus formosana (Florin) Florin (GenBank accession no. AB831010). We used MSATCOMMANDER  (Faircloth, 2008)  to screen chloroplast regions, including microsatellites, for ≥10 mononucleotide repeats, ≥6 dinucleotide repeats, and ≥4 tri-to hexanucleotide repeats.
We designed PCR primer pairs for these regions using Primer3  (Rozen and Skaletsky, 1999) ; we chose an optimal annealing temperature of 60(±1)°C, a GC content of 30-70%, and product size ranges of 100-500 bp. Nineteen primer pairs, including seven amplifying mononucleotide, six dinucleotide, four trinucleotide, and one tetranucleotide repeats, were employed to explore PCR amplification and polymorphism in two C. sulcata populations from the northern catchments of the Dumbéa River (n = 21; 22°06′S, 166°30′E) and the Koéalagoguamba River (n = 20; 22°01′S, 166°23′E). Voucher specimen accession numbers were as follows: KYO 00019999 for the Dumbéa River populations and MO 727065 for the Koéalagoguamba River population. For all loci, the forward primer contained one of three different M13 sequences (5′-CACGACGTTGTAAAACGAC-3′, 5′-TGTGGAATTGTGAGCGG-3′, or 5′-CTATAGGGCACGCGTGGT-3′) and the reverse primer was tagged with a PIG-tail sequence (5′-GTTTCTT-3′)  (Brownstein et al.
Chloroplast genome sequencing of eight plant species. The genome sizes are shown as the mean (with minimum and maximum) for each genus. Genome size data were taken from the RBG Kew Angiosperm DNA C-values database, version 8. 0(Bennett and Leitch, 2012). Sequencing of the fragment DNA libraries yielded 4,981,436 cleaned reads in total. The number of reads varied from 81,578 for A. sakawanum Makino var. stellatum (F. Maek. exAkasawa) T. Sugaw. to 1,939,765 for C. sulcata (Table1; DNA Data Bank of Japan [DDBJ] Sequence Read Archive accession numbers: DRA005207-DRA005214). The baiting and iterative mapping approach implemented in MITObim reconstructed the chloroplast genome sequences of the eight species (accession numbers: AP017904-AP017911). The lengths ranged from 131,609 bp in C. sulcata to 161,922 bp in A.
Atherton, r. A. , B. j. mccomiSh, l. D. ShepherD, l. A. Berry, n. W. AlBert, AnD p. j. lockhArt. 2010. Whole genome sequencing of enriched chloroplast DNA using the Illumina GAII platform. Plant Methods 6: 22. Bennett, m. D. , AnD i. j. leitch. 2012. Angiosperm DNA C-values database (release 8. 0, Dec. 2012). Website http://data. kew. org/cvalues/ CvalServlet?querytype=2 [accessed 21 April 2017]. BookjAnS, g. , B. m. StummAnn, AnD k. W. henningSen. 1984. Preparation of chloroplast DNA from pea plastids isolated in a medium of high ionic strength. Analytical Biochemistry 141: 244-247. BroWnStein, m. j. , j. D. cArpten, AnD j. r. Smith. 1996. Modulation of non-templated nucleotide addition by Taq DNA polymerase: Primer modifications that facilitate genotyping. BioTechniques 20: 1004. cAto, S.
Funding was provided by a  Japan Society for the Promotion of Science   Grant-in-Aid for Scientific Research  ( KAKENHI 26850098 ) and the  Environmental Research and Technology Development Fund of the Ministry of the Environment SICORP Program of the Japan Science and Technology Agency  ( 4-1403 ). The southern province of New Caledonia provided financial support and authorized (by exception) the collection of C. sulcata material. The authors thank the traditional owners of the sampling areas for supporting the research program.
sulcata to 161,922 bp in A. sakawanum var. stellatum. GenBank Voucher accession no. accession no. b AP017910 KYO 00019985 AP017907 KYO 00019984 AP017909 KYO 00019986 AP017908 KYO 00019902 AP017911 KYO 00019989 AP017905 KYO 00019987 AP017906 KYO 00019988 AP017904 KYO 00019999 Average read depth 237 52 74 20 64 93 222 220 Percentage of mapped reads 59. 54 24. 6 22. 64 20. 61 17. 76 16. 18 14. 56 10. 13 No. of mapped reads 246,049 48,516 63,443 16,816 53,858 83,609 181,915 196,403 Recovered chloroplast genome length (bp) 158,524 153,981 158,781 161,922 154,809 157,692 156,963 131,609 No. of cleaned reads 413,236 197,233 280,233 81,578 303,264 516,820 1,249,307 1,939,765 Input tissue weight (g) 2. 00 0. 95 1. 40 0. 84 2. 10 2. 00 2. 00 0. 40 Genome size Reference sequence (1C in pg) a (GenBank no.