dataesr/scientific-paragraphs-categorization huggingface dataset:
Consequently, the oldest valid U-Th date is 127. 2 ± 1. 3 ka from a sample located 7 mm above the base of the stalagmite, and extrapolation of the age model places the beginning of growth at ∼ 128. 8 ka. However, as no other coeval stalag-mites from Kanaan Cave have yet been dated, it is unclear whether the base of our record corresponds to the onset of the LIG optimum. In general, U-Th dates from the upper segment of the stalagmite were more consistent with only one out of seven dates (80. 6 ± 0. 5 ka) clearly out of sequence. Following the exclusion of obvious outliers, the remaining U-Th data set showed a number of age reversals. For the purposes of age-depth modelling, where age reversals were resolvable at the 3σ level, the younger date was assumed to represent the correct age progression and the older dates were excluded such as the basal age in the previous age model  (Verheyden et al.
Conclusions A dated MIS 5 stalagmite record (129-84 ka) from Kanaan Cave, Lebanon, demonstrates the potential of stalagmite records for palaeoclimate reconstruction in the northern Levant. The K1-2010 model age coupled with growth rates and isotopic data provide a more precise record of the climatic changes that occurred during the last interglacial and on into the glacial inception period. The K1-2010 speleothem record indicate a very wet early LIG and during the LIG optimum at the global scale, and is in agreement with warm and humid conditions demonstrated in other speleothem and lacustrine records from the Mediterranean. The K1-2010 isotope record and growth rate curves clearly demonstrate an important change from 126. 3 ± 0. 9 to 120. 3 ka (interpolated). The change seems to be driven mainly by a "source" effect, reflecting the δ 18 O Mediterranean Sea surface water composition and the EM isotopic increase at ∼ 126 ka during the S5 event.
Figure1. The eastern Mediterranean showing the location of palaeoclimate records including this study and the major wind trajectories (Saaroni et al. , 1998) , including the mid-latitude westerlies, and occasional incursions from the Sharav cyclone and the Sharqiya. The northsouth and east-west precipitation gradients are indicated by dashed dark lines (isohyets in mm). CL: Cyprus Low. The locations of Kanaan Cave and other Levantine palaeoclimatic records spanning the MIS 5 period cited in the text are numbered 1 to 14. Records derived from marine studies are indicated by points, lake level reconstructions and pollen data by rectangles, and speleothems by stars. (1) Core MD 70-41 (Emeis et al. , 2003) ; (2) core LC21 (Grant et al. , 2012) ; (3) core ODP Site 967 (Rohling et al. , 2002 (Rohling et al.
Acknowledgements. This study was funded by the  mobility fellowship programme of the Belgian Federal Scientific Policy (BELSPO) , co-funded by the  Marie Curie Actions of the European Commission. We acknowledge  EDYTEM Laboratory  ( UMR-5204 CNRS ) and  Saint-Joseph University  for making stalagmites of Kanaan Cave available for analyses during this study. We would like to thank the support of  ALES  ( Association Libanaise d'Etudes Spéléologiques ) members who accompanied us during field trips. Farrant and Noble publish with the approval of the Executive Director,  British Geological Survey. We thank  Thomas Goovaerts  for sampling preparation,  Daniel Condon  for assisting with the uranium series analyses and  Kevin De Bont  for assisting in water analyses. We also thank the anonymous reviewers for their constructive comments and reviews.
First results and preliminary discussion Both cores are composed of grey lacustrine calcareous marl, with a variable proportion of clay. A preliminary 82 m long composite succession is based on visual correlation after core opening and using the line scan images from the Itrax core scanner (Cox Analytical, Sweden) and the Corewall software package (Corelyzer 2. 0. 1). Core sections that looked disturbed from the coring process were not included in the core composite. Although, theoretically, the drilling depths should have avoided gaps between the individual runs in each hole, overlapping sequences in the lower part of the F3 hole indicated gaps of around 20 cm between the individual runs of the F1 hole due to core expansion after recovery. We therefore assumed 20 cm gaps between the F1 sequences, where there were no overlapping F3 sequences; i.
Acknowledgements. We thank  Michele and Valentino Pietrantoni  for having kindly hosted us and for having permitted to perform the drilling in their farm. Alison Pereira  is thanked for field help during the drilling. Alex Francke  and  Niklas Leicher  provided valuable help during core management at the  University of Koln. Edited by:  U. Harms Reviewed  by:  J. Brigham-Grette  and one anonymous referee.
The classification of archaeological assemblages in the Middle Stone Age of South Africa in terms of diversity and temporal continuity has significant implications with respect to recent cultural evolutionary models which propose either gradual accumulation or discontinuous, episodic processes for the emergence and diffusion of cultural traits. We present the results of a systematic technological and typological analysis of the Still Bay assemblages from Sibudu and Blombos. A similar approach is used in the analysis of the Howiesons Poort (HP) assemblages from Sibudu seen in comparison with broadly contemporaneous assemblages from Rose Cottage and Klasies River Cave 1A. Using our own and published data from other sites we report on the diversity between stone artifact assemblages and discuss to what extent they can be grouped into homogeneous lithic sets.
Methods The study of lithic assemblages is based on quantitative and qualitative analyses including metrical, technological and typological attributes combined with a particular attention to the sequence of manufacture and reworking of Still Bay points at Sibudu, in comparison with the Still Bay points of Blombos. Details of the site context and the stratigraphic sequences of Sibudu and Blombos are discussed by section; detailed analytical protocols are provided in the Supporting Information. Still Bay and Howiesons Poort primary data used here are available both in the body of the paper and in Supporting Information. Supporting Information for the Still Bay at Sibudu and Blombos is provided in S1 File, S2 File and S3 File. Supporting Information for the Howiesons Poort is provided in S4 File, S5 File and S6 File.
The number of shaping flakes raises the question of the in situ production of bifacial pieces. Is the number of flakes in agreement with the number of bifacial pieces? The ratio of shaping flakes to points is 40, that is 2169 flakes >10 mm to 54 bifacial pieces. Too few bifacial pieces are complete to allow counting of the number of flake scars >10mm on their faces. Recent experiments on manufacture of Solutrean bifacial points do not provide data to answer this question  [50] , but Newcomer  [55]  obtained almost 50 shaping flakes per handaxe, including flakes from the initial shaping. So the ratio of shaping flakes to biface we observe at Sibudu seems coherent. Nevertheless, most of the Sibudu bifaces are small fragments, mainly tips. This suggests that the complementary larger fragments have been moved to an unexcavated part of the site or away from the site, before or after retooling.
Funding: Research of PV, SS, and AD at Sibudu was funded by a grant from the  National Science Foundation  ( BCS 0613319  to PV). PV's research was also funded by the  Wenner-Gren Foundation , the  Paleontological Scientific Trust , and the  Frison Institute. The  Paleontological Scientific Trust with its Scatterlings of Africa  programme supported LP with postdoctoral funding in 2008. PV and LP acknowledge support from the  School of Geography, Archaeology and Environmental Sciences, University of the Witwatersrand. Financial support for CSH was provided by a  European Research Council  Advanced Grant, TRACSYMBOLS No. 249587 , awarded under the  FP7 programme  at the  University of Bergen, Norway , and by a  National Research Foundation/ Department of Science and Technology  funded Chair at the  University of the Witwatersrand, South Africa.
Acknowledgements. This special issue has been supported by the  MIAVITA  project, financed by the  European Commission  under the 7th  Framework Programme for Research and Technological Development, Area "Environment", Activity 6. 1  " Climate Change, Pollution and Risks ". We thank the NHESS editors and supporting staff for hosting this special issue. We also thank all participants to the  MIAVITA  project for their contributions, and  Chris Newhall  for very useful advice. We are grateful to the referees for reviewing the papers of this special issue, whose time and work contributed considerably to the quality of the paper. Finally, we thank  Stephen Conway  for proofreading this preface.
Statistical Analysis Chi-squared test, Kruskal-Wallis rank sum test and Fisher's exact test were conducted to identify any significant change over the study period. To assess characteristics of patients with and without FC, non-parametric tests were used for univariate analyses (Wilcoxon and Fisher's exact tests). Variables achieving a p value <0. 17 were entered into a multivariate logistic regression model with a backward stepwise method based on the likelihood ratio test. Regarding repeated exposures during the study period, mixed effects logistic regression to model binary outcome variables, in which the log odds of the outcomes are modeled as a linear combination of the predictor variables, were used to fit a generalized linear mixed model. Seroconversions data were assessed by patient with Fisher's exact test.
These techniques are now increasingly used for data visualisation, clustering and classification of large datasets  (Yin, 2008). They allow representation of a multidimensional dataset by nonlinear projection of artefacts in a lower dimension space, usually represented by discrete locations in a regular 2D lattice. Despite the loss of linearity in the output space, the topological relationships between objects (the order of the distances) are preserved  (Liu and Weisberg, 2005). In our case, the use of SOMs can provide a first, exploratory step for further clustering of large datasets  (Yin, 2008). Artefacts are assigned to the most similar prototype (also called codebook) vectors, which represent a set of locations summarizing the original data. Interestingly, the density of codebook vectors increases with the density of artefacts, making possible (unlike PCA) a high level of detail in the structure of the data where the artefacts are the most numerous.
, 2012) , momocs  (Bonhomme et al. , 2014) , kohonen  (Wehrens and Buydens, 2007) , and ks  (Duong, 2007)  packages, all written for the free R software (R Core  Team, 2014). Multinomial scan statistics and geographical mapping were performed using SaTScan v9. 3. 1 (http://www. satscan. org/; Kulldorff et al. , 1998)  and Quantum GIS 2. 6. 1. (http://www. qgis. org; QGIS Development  Team, 2015) , both of which are freely available software programs.
Note that if the n was lower, say 4 or 3, the observed spatial patterns might have not been identified. In general, increasing n above the optimal value is therefore not as unfavourable as underestimating n. In any case, the determination of meaningful groups in the dataset is a problem shared by all classification techniques (e. g. ,  Orton, 1980; Legendre and Legendre, 1998). The methodology proposed here is unequivocally statistical: the more data available, the better the output, meaning that this method is not suitable for too small a corpus. Artefact assignment is probabilistic, so that it does not formally answer the question of strict attribution to a single type, except when the highest probability is considered. When the classification is based on this highest probability, subsequent analyses take for granted this classification and therefore no longer consider the level of uncertainty associated with it.
It should be mentioned that the archaeologist experimenting typological attributions based on visual observation alone, faces the same restrictions, but unfortunately without any probabilistic control. The groups defined by this approach cannot be thought of as definitive. Any new artefact added to the dataset will marginally modify the output. By its very nature, the proposed model for creating and validating the classification is one stage in a never-ending process. Any new artefact specimen or other observed descriptor for the artefact (e. g. decoration, chemical composition, or fabrication process) can test the stability of the proposed model and may thus contribute to its improvement. It is noteworthy that the morphometric approach proposed here takes into account only the shape of the external outline of the axe.
Socioeconomic variables entailed whether the patient had some professional activity and whether the patient received outpatient rehabilitation and formal care. Mean daily formal care time (FCT) was calculated from the number of publicly funded formal care hours per typical care day and the number of days per typical care week. The Multiple Sclerosis Severity Score (MSSS)  [9, 13]  was used to sort patients by MS progression course over time (scatter plot figures). The MSSS was developed using large databases ( = 9892 patients with a MS duration ranging from 1 to 30 years) and uses a probabilistic method to assign a disease severity score adjusted for disease duration (relating a patient's EDSS to the distribution of disability in patients with the same MS duration). The MSSS table was used in the perspective of the disability severity course, whatever the history of relapsing-remitting or progressive form is; throughout the paper, the terms "fast progressors" and "slow progressors"  [16]  were used to denote patients with a respective rapid and slow disability course at the time point of the study (identified in the matrix of distribution of the MSSS deciles as patients over the 8th decile and under the 2nd decile, resp.
Patients and Primary Informal Caregivers. A total of 216 patients were included, of which 99 primary IC returned the self-administered postal questionnaire. There were no significant differences between the inclusion data (age, gender, and EDSS; all  > 0. 05) of the patients whose IC returned the questionnaire and those whose IC did not return the questionnaire. The patients included were mostly women (63%) with a mean age of 46 years and a wide range of EDSS scores (from 1. 0 to 9. 5) (Table  1 ). A significant level of cognitive impairment was found with the most frequent dysexecutive disorders (DEX) reported in order as follows: apathy and lack of drive, aggression, planning problems, distractibility, poor decision-making ability, euphoria, shallowing of affective responses, and inability to inhibit responses.
Acknowledgments The authors thank  Rana Assouad ,  Marina Bardon ,  Elisabeth Maillart , and  Fanny Pineau  for their help. E. Bayen  received a PhD grant support from the  French Society of Physical Medicine and Rehabilitation  (SOFMER, 2012), from the  Caisse Nationale de Solidarité pour l' Autonomie  (CNSA, 2013), and from the  Fondation de France  (2014). These funding sources were not involved in the study design, in the collection, in the analysis, in the interpretation of data, or in the writing of the paper.
In accordance with French legislation, patients and informal caregivers were informed about the purpose of this study upon inclusion in the database and consent was obtained after oral and written information. Mean (SD; range [min-max]) Count (%) [4,8,21,22]published studies of burden in MS[4,8,21,22], on univariate analyses, and on screening for multicolinearity as follows: (1) patient's variables (EDSS, DEX, urinary intermittent catheterisation, and MS duration); (2) caregiver's variables (gender, coresidency, conflicting role, SF12-MCS, and SF12-PCS); (3) socioeconomic variable (formal care support at home). For each multivariate model an ascending strategy was computed with a progressive addition of each block of variables. The Akaike Information Criterion (AIC) which enables to select the most parsimonious model when dealing with empirical.
3. 3. 1, part of the DENDROPY v. 3. 12. 0 package  [53]. In MRBAYES, we did four independent Markov chain Monte Carlo (MCMC) runs of 25 million generations using four chains per run (nchains = 4) with incremental heating of chains (temp = 0. 1) sampled every 2500 generations. We changed the default branch length prior to unconstrained with an exponential distribution for all partitioned analyses to avoid artificially long branches (prset applyto = (all) brlenspr = unconstrained:exponential(100);  [54] ). We assessed convergence of parameter estimates and tree splits in TRACER v. 1. 5  [55]  and ARE WE THERE YET? (AWTY?;  [56, 57] ), respectively. We assessed topology convergence between runs by the average standard deviation of split frequencies (ASDSF) and  Coloured distributions correspond to the 11 major phylogenetic lineages of the T.
Molecular dating and species delimitation We estimated divergence time in BEAST v. 1. 7. 5  [58, 59]  implemented with BEAGLE  [51]. We included two individuals per nominal subspecies for all Todiramphus taxa, except T. sanctus, for which we included only known breeding populations (e. g. Australia, New Zealand, New Caledonia, Solomon Islands, Vanuatu and the Santa Cruz group; electronic supplementary material, table  S1 ). We linked clock and tree models, but nucleotide substitution models were unlinked. We used MRMODELTEST to partition the data in the same way we did our MRBAYES analyses (electronic supplementary material, table  S3 ). We used a birth-death speciation process for the tree prior. To test for clock-like evolution, we compared likelihoods of runs with a strict clock to those with a relaxed lognormal clock (UCLD).
We failed to reject a strict molecular clock using a likelihood ratio test (p = 1. 0). Additionally, the coefficient of variation frequency histogram of the ucld. std parameter abutted against zero when viewed in TRACER, which is a symptom that the data cannot reject a strict molecular clock  [60]. We ran 10 independent MCMC chains for 100 million generations and sampled every 20 000th generation. We examined burn-in and convergence diagnostics in TRACER; burn-in values were specific to each run with at least 25% of samples discarded, with some runs requiring up to 40% burn-in. Lacking fossil calibration data for this group, we relied on published rates of mtDNA sequence evolution to calibrate our divergence dating analyses. Substitution rate priors derived from ND2 substitution rates for Hawaiian honeycreepers were used (0.
ND3 was omitted from BEAST analyses to simplify mitochondrial rate calibrations. We examined species delimitation and diversification rates to objectively compare patterns of diversity in T. chloris to other published phylogenies of rapid geographical radiations (e. g. Zosterops and Erythropitta). We delimited species with a Bayesian implementation of the general mixed Yule-coalescent model implemented in the R package, bGMYC  [63]. We used the ND2 data and followed the authors' parameter recommendations [mcmc = 50 000; burn-in = 40 000; thinning = 100]. The GMYC model  [64]  is advantageous for single-locus datasets such as those generated by DNA barcodes or when the majority of phylogenetic signal occurs in the mtDNA, including rapid radiations like Todiramphus. We calculated diversification rates assuming a Yule process from the following formula: [ln(N)-ln(N o )]/T, which uses initial diversity (N o = 2), extant diversity (N) and time (T) since origin of the crown clade  [65].
Evaluating the role of children in individual's decisions regarding migration duration can thus help to further understand the multidimensional impact of the hukou system on migration. The overall goal of this paper is to explore the role of children as a motive for return migration in China. We first present a simple illustrative model of migration duration based on  Dustmann (2003)  that accounts for left-behind children through parents' altruistic behaviour. The discussion also points to the potential differentiated impact of children on return decisions depending on their age and gender. Then, using a unique dataset collected in 2008, we provide an empirical test based on a duration model. We estimate the determinants of the length of migration for both on-going migrants with incomplete migration spells and return migrants with complete migration spells.
In particular, we use a recent database that covers a period during which both rural-urban migration and a counter-flow of population back to the countryside increased dramatically, which may enable us to better capture recent changes associated with return migration in China. Based on these data, our analysis also adds to the work of  Connelly et al. (2012)  by examining the whole migrant population (rather than only women at the age of childbearing) and by analysing both complete and incomplete episodes of migration for the entire individual migration history. Second, by examining the determinants of the length of migration, this paper also contributes to filling the lack of research on migration duration in China. Although the length of migration is an important indicator of the flow and scale of migration as well as the economic effects on both receiving and sending regions, it has received rather limited attention in the migration literature, including for international migration 5.
Study area and data description The data used in this paper come from a series of rural household interviews conducted in Wuwei county, Anhui province, from September to November 2008. Wuwei county was selected because of its relatively long labour force export history, the county being famous for sending out female domestic service workers since the beginning of the 1980s. According to local official statistics, individuals working outside the county accounted for 43 per cent of the entire county's rural labour force at the end of 2006  (Wuwei County Government, 2007). Together with the large-scale migration, the county is also characterised by a sizable number of left-behind children 6. Recent data show that in the district of Wuhu, to which Wuwei county belongs, 83,400 of the total 332,000 compulsory education students are left-behind children, accounting for 25% of the entire student population 7.
The sample used in this paper is composed of 284 individuals having a migration and/or return migration history, with 125 return migrants and 159 on-going migrants. Return migrants are individuals who are currently residing and working in the county with at least 6 months migration work experience outside the county. These individuals declared having no plan to migrate again. On-going migrants are individuals who are either in cities at the time of the survey 8 or temporarily returned with a plan to migrate again soon (they are circular migrants). This dataset is unique in that it contains detailed information both on the complete length of migration for each return migrant and the incomplete length of migration for on-going migrants. For return migrants, the length of migration duration is defined as the total cumulated number of years of migration from the year of an individual's first migration up to the year of her last return.