dataesr/scientific-paragraphs-categorization huggingface dataset:
Acknowledgement. We thank  Anne Petrenko ,  Louise Rousselet ,  Alain De Verneil ,  Christophe Yohia  and  Christelle Pinazo  for modeling outputs. We also thank  Nicole Garcia  for Chl a measurements. Special thanks go to the  DT-INSU people from la Seyne sur Mer , and in particular to  Malika Oudia  for her help in administration work and  Céline Heyndrickx  and  Frédéric Le Moal  for their technical support. Genavir and in particular  Julien Fenouil  are acknowledged for providing and assisting us with the MVP. MVP and the associated captors were brought by  IFREMER  and LOPB with co-funding by the  Centre Européen de Technologies Sous-Marine  (CETSM -Contrat de Projet Etat Région 2007-2013 en PACA) and the  French ANR   FOCEA  (project  ANR-09-CEXC-006-01  to  M. Zhou  and  F. Carlotti ).
We also thank  J. Thomas Farrar  ( MIT ) for his suggestions on the cruise strategy,  Alain Lefèbvre  ( IFREMER ) for the Pocket FerryBox, and  Frédéric Partensky  for his constructive discussion. The OSCAHR cruise was supported by the  MIO "Axes Transverses" program  ( AT-COUPLAGE ), by  FEDER  fundings (PRECYM flow cytometry platform) and by the following projects: CHROME (PI M. Thyssen, funded by the  Excellence Initiative of Aix-Marseille University  -A*MIDEX, a  French "Investissements d'Avenir" program ),  SeaQUEST (PI O. Ross , funded by the  UE FP7 people) ,  AMICO (PI C. Pinazo , funded by  Copernicus -MEDDE French Ministery MDE) , and  BIOSWOT (PI F. d'Ovidio , funded by  TOSCA/CNES ). We also thank the captain and crew of the Téthys II research vessel. The project leading to this publication received funding from the  European FEDER Fund  under project  1166-39417.
, the most commonly used taxon in Mg/Ca-based palaeotemperature reconstruction. The reliability and accuracy of the new Mg/Ca-temperature calibration is corroborated in the fossil record by a case study in the Timor Sea which demonstrates the presence of southern-sourced waters at intermediate depths for the past 26,000 years. Costate species of Bulimina might thus provide a valuable alternative for BWT reconstruction in mesotrophic to eutrophic settings where many of the commonly used (more oligotrophic) species are rare or absent, and be particularly useful in hypoxic settings such as permanent upwelling zones where costate buliminids often dominate foraminiferal assemblages. The evaluation further reveals a mean positive offset of $0. 2 mmol/mol of the Atlantic data-set over the Indo-Pacific dataset which contributes to the scatter in our calibration.
Mg/Ca trends in the core top data-set Bulimina tests from the core top samples show Mg/Ca ratios between 1. 15 and 2. 57 mmol/mol over the 3-13. 1 °C temperature range (Table  2 ). In all study areas, Mg/Ca ratios decrease with increasing water depth and parallel the decrease of in situ temperatures (Fig. 2 ). The agreement between Mg/Ca ratios and in situ temperatures is not only observed along continuous transects in the Makassar Strait, the Timor Sea and off Cape Blanc, but also between individual sites bathed by different water masses in the NE Atlantic. There, the temperature anomaly created by MOW at 1073 m water depth in the Gulf of Cadiz is well expressed in the Mg/Ca data (Fig. 2 ). Parallel analyses of the different morphotypes at six sites (BJ8-03-81MC, -88MC, -96MC; SO185-18494, -18501, -18502) reveal minimal and statistically insignificant (t < À0.
2 ; Table  2 ). Ontogenetic variability has been considered in previous studies as potential source of bias on Mg incorporation in foraminiferal shells, in particular for non-costate Bulimina species  (Hintz et al. , 2006b; Wit et al. , 2012; Groeneveld and Filipsson, 2013 ). To minimize a potential bias, it has been recommended to use tests of a limited range of Fig. 7. Offset between Atlantic (blue = new data; grey = published data of  Lear et al. , 2002, and Tachikawa and Elderfield, 2004)  and Indo-Pacific (green = new data; black = published data of  Lear et al. , 2002)  data-sets of costate Bulimina species. Samples with particularly low Mg/Ca values in the Indo-Pacific data-set are highlighted in grey (see text for discussion). test-size  (Wit et al. , 2012; Groeneveld and Filipsson, 2013).
Due to the general restriction of test-size to >300 mm in our study we consider the contribution of a potential ontogenetically induced test-size effect to the scatter in our data set as minimal (Table  2 ). Samples composed of smaller tests (150-300 mm) show no systematic deviations from the general trend; however, the small number of samples (n = 3) does not allow for a reliable evaluation of ontogenetic effects. A characteristic feature of our data-set that considerably contributes to the scatter and SEE is the mean positive offset of Mg/Ca ratios between Atlantic and Indo-Pacific sites (D Mg/Ca A-IP ). Separate regression analyses for the two study areas result in Mg=Ca Atlantic ¼ 0:119ðAE0:014ÞT þ 1:059ðAE0:100Þ ðR 2 ¼ 0:87; SEE ¼ 0:15; p < 0:0001; n ¼ 13Þ ð3Þ Mg=Ca IndoÀPacific ¼ 0:103ðAE0:016ÞT þ 0:957ðAE0:101Þ ðR 2 ¼ 0:54; SEE ¼ 0:17; p < 0:0001; n ¼ 40Þ ð4Þ While the slopes of the two regression lines are not significantly different (t = À0.
9; p = 0. 18), they show an offset between 0. 15 and 0. 26 mmol/mol over BWTs of 3-10 °C (temperature range covered by both data-sets; Fig. 7 ). To our knowledge, no such geographic offset has been identified in previous core top studies on calcitic benthic foraminiferal tests. Extended data sets from the Atlantic and Indo-Pacific regions are available only for few benthic foraminiferal taxa. A review of data for the epifaunal Cibicidoides wuellerstorfi shows no systematic offset between the two regions (Fig. S1a ;  Russell et al. , 1994; Rosenthal et al. , 1997; Lear et al. , 2002; Martin et al. , 2002; Elderfield et al. , 2006). However, due to the limitation to BWTs from 1 to 3. 5 °C, Mg/Ca ratios are likely compromised by low levels of carbonate ion saturation (see below;  Elderfield et al.
, 2006). Data over a much broader BWT range are available for the shallow infaunal Uvigerina spp. and Hyalinea balthica. Regression analyses for Uvigerina spp. show no discernable discrepancies between the Atlantic and Pacific data-sets (Fig. S1b ;  Lear et al. , 2002; Tachikawa and Elderfield, 2004; Bryan and Marchitto, 2008; Elderfield et al. , 2010; Tisserand et al. , 2013) ; only if data from the Indian Ocean  (Elderfield et al. , 2006)  are added to the Pacific data-set, Atlantic data show a minimal positive offset of $0. 05 mmol/mol. However, potential biases from different cleaning protocols and species-specific vital effects prevent a more thorough assessment of the Uvigerina data  (Bryan and Marchitto, 2008; Elderfield et al. , 2010). For Hyalinea balthica, regression analyses for the Atlantic (BWT range: 7-13 °C) and Pacific (4-10 °C) data-sets suggest an offset comparable to that of costate buliminids, which has not been addressed in the original publication (Fig.
Based on an extensive data-set from the Atlantic,  Weldeab et al. (2016)   ] pore water at the Atlantic sites close to or slightly above zero whereas Indo-Pacific pore waters would be clearly undersaturated. The fact that almost all our data are well between the discussed thresholds for BWT and D [CO 3 2À ] bottom water should thus not be taken at face value and a carbonate ion effect discarded. Instead, a potential carbonate ion effect from undersaturated D [CO 3 2À ] pore water should be considered, particularly for the Indo-Pacific sites. This effect thus offers a potential explanation for the general offset of Atlantic and Pacific data which would also be in line with the pattern observed for the aragonitic Hoeglundina elegans. A closer inspection of our data-set shows that D Mg/ Ca A-IP is largely driven by strongly depleted Mg/Ca values at BWT >7 °C at Indo-Pacific sites located in the thermocline between $360 and 600 m (Figs.
ACKNOWLEDGEMENTS The authors thank  William A. Berggren  (Rutgers) and  Paola Moffa Sanchez (Cardiff)  for helpful discussion. Ryan Bu  (Rutgers) is thanked for assistance with ICP-MS measurements. The final manuscript has benefitted from the constructive comments of two anonymous reviewers and associate editor  Thomas M. Marchitto. This study was financially supported by a grant of the  Max-Kade-Foundation  to P. G. and contributes to projects FWF  P25831-N29  of the  Austrian Science Fund  (W. E. P. , P. G. ) and  NSF   OCe 1334691  (YR).
, 2017b). Previous studies in this system reported lower reproductive success in intensive farmlands than in non-intensive areas  (Ghilain & Bélisle, 2008; Lessard et al. , 2014). All nest boxes are visited every two days during breeding seasons (early May to mid-July). Breeding females and social males are captured and marked at the nest box during incubation and food provisioning, respectively. Blood samples are collected for subsequent DNA analyses. Data were collected in compliance with the Canadian Council on Animal Care, under the approval of the Université de Sherbrooke Animal Ethics Committee (protocols DG2014-01 and FP2014-01).
Rising wage inequality From 1975 to today, wage inequality has increased dramatically in Germany, with a substantial acceleration around the year 1993. We use administrative worker-level data based on social security records to document these facts. More precisely, we work with a 2% random sample of the universe of employed persons under social security (i. e. , excluding the self-insured). These data are provided by the Institute for Labor Market Research, the official research body of the Federal Employment Agency, and can be accessed at the Institute's premises in Nuremberg. 36   Following common practice, we focus on male full-time workers aged 20 to 60 with work places in Western Germany for the years 1975 to 2007, the last year before the Lehman Brothers collapse triggered a massive crisis of world trade.
16 points from about 0. 27 to 0. 43 (see Figure  2 ). This is about 4/5ths of the rise in raw wage inequality in this period. Again, there is evidence for a break in the trend around the year 1993. Before that point, residual inequality was increasing at an average yearly rate of 0. 0030, and after that point by about 2. 5 times that rate. While less pronounced, other countries exhibit similar trends. 40   As a next step, we wish to highlight the role of firm characteristics in explaining wage inequality. We use matched employer-employee data, the so called LIAB data set. It consists of 36 German social security data covers the whole workforce subject to unemployment insurance. The data is deemed of very high quality. Wage income is censored at the annual Social Security earnings maximum. This affects at most the 14% highest incomes.
We use well-established imputation algorithms to deal with this problem; see  Dustmann et al. (2009)  for a discussion. This paper also shows that the inequality measures obtained from these data are broadly comparable to the ones obtained from survey data. The 2% sample is known as the SIAB data base; a matched employer-employee data set is also available (LIAB data). See Appendix C. 1 for details. 37 Daily wages are computed by averaging yearly wages. There is no information on hours worked so that increased inequality in working hours across workers cannot be captured. The use of log wages ensures that the inequality measure is not sensitive to the choice of units for wages. 38 A similar finding, based on variants of the same data set, is reported by  Dustmann et al. (2009)  and  Card et al.
Notes. Decomposition using log daily wage data for West German, male, full-time workers, aged 20-60, as reported in German social security data (LIAB data). Also see  Baumgarten (2013). Finally, we decompose the log wage variance into a within-plant and a between-plant component. 46 We do so using the LIAB data set, and report the years 1996 and 2007. We find that, in 1996, both components were of similar importance for total wage inequality; in 2007, the relative weight of the within component has decreased to about 44%. Thus, about two thirds of the total increase in inequality is due to the between component. 45 The variance components missing in Table  1  (the variance of the residual, the covariances between plant or person effects with the covariate index) make up less than 8% and do not show any significant trend.
oecd. org/els/social/workincentives. 52 Another important change in labor market institutions in those years is the reduction in collective bargaining coverage which followed German reunification. The German LIAB data show that about 80% of all firms were covered by industry agreements in 1996; this share declined to about 60% in 2007, with most of the decline occurring prior to 2005. We do not study this institutional change here; for an empirical analysis of the links between the fall in collective bargaining and wage inequality in Germany see  Dustmann et al. (2009)  and  Card et al. (2013). 53 We chose σ in line with  Bernard et al. (2007). Kohlbrecher et al. (2013)  present estimates of η based on the SIAB data base (which contains information about unemployment spells of workers) for the period 1993 to 2007.
Controlling for workforce heterogeneity, they cannot reject a constant returns to scale specification of the matching function. are, therefore, naturally compatible with the SIAB and LIAB data bases provided by the same agency and used for other moments in our quantitative exercise. The firm and job destruction rates are taken from  Fuchs and Weyh (2010). These authors use the Establishment History Panel of the IAB for the period 2000 to 2006. The data base includes all plants in Germany with at least one employee subject to social security. We use their empirical estimates to set the yearly plant exit rate δ = 5% per year, and the yearly job destruction rate due to match dissolutions χ = 7%. For the parameter b, we use the replacement rate 54 of 0. 35 reported in  Kohlbrecher et al. (2013).
They rated on 12 self-presentation traits how they wanted to appear during the interview to make the employer select them. The traits, derived from  Kervyn, Yzerbyt, Judd, and Nunes (2009) , measured the extent to which participants wanted to appear competent (positive valence: skilled, determined, competent; negative valence: disorganized, negligent, messy) and warm (positive valence: caring, tolerant, warm; negative valence: cold, selfish, insensitive). Scales ranged from 1 (not at all) to 7 (very). All data from the three experiments can be retrieved online on the Open Science Framework (OSF) repository at the following link: https://osf. io/pbfh4.
Participants were then asked to imagine they had been called to a police interrogation regarding the crime. Instructions and ratings of preferred appearance followed those of Experiment 2. In addition, ostensibly to help the police gauge their credibility, participants were asked to write a few sentences describing themselves. We used the Linguistic Inquiry and Word Count  (Pennebaker, Booth, Boyd, & Francis, 2015)  text analysis software to determine the semantic content related to competence and warmth in participants' self-descriptions. We used default categories to estimate competence (words longer than six letters, power, and work) and warmth (positive and negative emotion, social processes, family, and friend).
We also explicitly exclude definitions  (Hutton 2004; Pratt 2009; Foord 2013 ) that overlap with 'creative' industries (i. e. , design and architecture), which are not necessarily 'new' and which are historically associated with central neighbourhoods. Rather, 'creative' industries act as independent variables in the econometric analysis. Compared to previous studies, we thus focus on a part of the new economy. All data are based on Statistics Canada micro-data drawn from the 1996, 2001 and 2006 censuses, and the National Household Survey (NHS) for 2011. Unlike the census (20% representative sample), the NHS is based on a voluntary questionnaire (sent to 33% of households) and must thus be used with some caution. Consequently, parallel analyses (with and without NHS data) are provided as a robustness check of results.
Micro-data for employed persons were aggregated at the census tract level by place of work. The census tract level 5  is the most appropriate scale as it allows us to accurately identify intrametropolitan NE employment dynamics and distinguish between central and suburban poles. Most previous studies on employment poles have used census tract-level data when available (see for example, for Canadian CMAs,  Coffey and Shearmur 2002, Shearmur and Coffey 2002). The micro data contain information on individual employment by industry and occupational class (at the 4-digit level). Most studies use industry classes; that is, the industry of the establishment in which the individual works. However, Statistics Canada changed classification schemes in 1997: industries prior to 1997 are based on SIC codes (Standard Industrial Classification) and on NAICS codes (North American Industry Classification System) thereafter, periodically updated.
Bars and Restaurants measures the number of jobs in bars and restaurants per capita, an indicator of urban amenities. Finally, for the CBD variable geo-coded street data was again used to compute the shortest road network distance between each CT and the CBD. 22  The coefficients estimated with the baseline OLS model (Equation  4 ) may suffer from endogeneity problems because of possible reverse causality and/or omitted variables bias. To reduce possible reverse causality problems, in a second step, we estimate a model using lagged explanatory variables: 23 Y it ¼ α þ ∑ 5 k¼1 β k Á Ind k;itÀ1 þ ∑ 5 l¼1 γ l Á Env l;itÀ1 þ δ Á CBD i þ γ t þ ε it : (5) Given that NE employment accounted for less than 3 per cent of total employment in 2006 in all three metro areas (less than 5. 5% using NOC), reverse causality should not be 18 The variable was mapped for the three metro areas; visual inspection confirmed that the principal concentrations were suburban, often around well-known office parks.
67) Computer systems analysts Computer programmers 65,035 Computer and information systems managers (3. 56) Computer engineers (except software engineers) Information systems analysts and consultants Database analysts and data administrators Software engineers Computer programmers and interactive media developers Web designers and developers Computer and network operators and web technicians User support technicians Systems testing technicians Codes M772 M772 5112 5142 5415 5112 5182 5415 5112 5182 5415 Codes 0213 2162 2163 0213 2147 2171 2172 2173 2174 2175 2281 2282 2283 Classification SIC 1980 SIC 1980 NAICS 1997 NAICS 2002 NAICS 2007 Classification NOC 1990 NOC 2001 Year 1996 2001 2006 2011 Year 1996 2001 Notes:Share of new economy jobs in total metropolitan employment in brackets.
High-throughput sequencing HTS was performed on a GAIIx instrument (Illumina, Little Chesterford, UK). Adapter and indexing sequences re-quired for Illumina multiplexing sequencing were added to the DNA libraries by PCR. The PCR products were purified on a 3% agarose gel and recovered by passive elution in TE-NaCl buffer (1× Tris-EDTA, 25 mM NaCl). Each eluate was concentrated and precipitated with ethanol before resuspension in distilled water. Samples were then mixed and loaded with 5 to 10% PhiX into a flow-cell and sequenced according to the provider's instructions. HTS data were demultiplexed and recovered in FASTQ format using bcl2fastq Conversion Software v1. 8. 4 (https://support. illumina. com/ downloads/bcl2fastq conversion software 184. html).
Primer trimming and quality filtering of sequences. All FASTQ files were processed using a software suite called PATTERNITY-SEQ, developed from the aptamer platform in MIRCen (http://jacob. cea. fr/drf/ifrancoisjacob/ english/Pages/Departments/MIRCen/Platforms. aspx? Type=Chapitre&numero=6). The different steps of the sequencing analysis are provided in Supplementary Figure  S1. First, adapter and primer sequences were removed from each sequence, keeping only variable regions. Then, sequences that contained at least one base with a quality score (Q) below 30 were removed before being saved in a FASTA format. This quality score can be converted to a probability of error (P) using the formula P = 10 (-Q/10). Thus, the recovered sequences contain bases with a potential probability of error below 0.
Phylogenetic tree of the ACE4 aptamer family All variants of the ACE4 family representing at least 0. 01% of the library in one round were used to build a phylogenetic tree using the software MEGA7  (38). The tree was generated by the Maximum Parsimony method using the Subtree-Pruning-Regrafting (SPR) algorithm  (39). The branch lengths were calculated using the average pathway method and are in the units of the number of changes over the whole sequence.
The ratio of competitive binding compared to ACE4 was calculated for each ACE4 variant by dividing the amount of radioactive ACE4 bound in the presence of wild-type ACE4 by the amount of radioactive ACE4 bound in the presence of a variant. Saturation binding experiments were performed using a protocol extensively described elsewhere  (45). Briefly, cells were incubated with radiolabelled 2 F-Py RNA aptamers at different concentrations, in 200L RPMI 1640, containing 100 g/ml of both tRNA and Poly(I), at 37 • C for 15 min. After washing five times with 500 l RPMI 1640, the amount of radioactive ACE4 aptamer attached to the cells was counted using a MicroBeta TriLux counter. The specific binding of aptamers was measured by subtracting the background values obtained with ACE4scr and apparent K d values were determined by fitting the binding curves with GraphPad Prism 6 (GraphPad Software, La Jolla, USA) using a one site-specific binding model.
The media was then replaced by fresh RPMI 1640 containing 0. 1% sodium azide and the dissociation of aptamers from the cell surface was measured for 60 min. Measurements were performed using the 2 × 3 opposite positions mode and a 4 s integration time per measure. The dissociation rate constant (k off ) was calculated using Trace Drawer software (Ridgeview Instruments AB). The association rate constant (k on ) was calculated by dividing the k off by the apparent K d previously measured during the saturation binding experiments.
Analysis of the mutational landscape from the doped cell-SELEX Galaxy Project was used to randomly select 800 000 sequences from the starting doped library and each round of selection. The positional nucleotide frequency of each library was obtained using BioEdit  (43). For each position p, a normalized enrichment ratio (RN) between rounds 4 and 0 (starting doped library) was calculated for each base b as described by the formula:   (46). They were added, using the software VARNA  (47) , as a color scale to the predicted secondary structure of ACE4 built by MFold  (48). The percentage of each sequence x and their Levenshtein distance to the wildtype ACE4 were obtained by PATTERNITY-SEQ to reconstruct a 2D mutational landscape of the ACE4 aptamer. Only sequences with a frequency >0. 001% in one round were recovered.
Thus, we amplified a chemically synthesized sequence of the ACE4 aptamer for 7 or 24 PCR cycles under the same conditions of our SE-LEX, but using primers that were elongated with adapter sequences for multiplexing HTS. We then analysed two million sequences per PCR products by HTS in triplicate. The number of variants dramatically increased between 7 and 24 PCR cycles and, consequently, the non-mutated ACE4 sequences decreased from 96% to 83%, respectively (data not shown). This suggests that the mutations were mainly due to polymerase errors rather than sequencing errors. This was further confirmed by comparing the mutation frequency per position. Indeed, the primer sequence, which is not enzymatically synthesized during PCR but elongated, was seven times less mutated after 24 PCR cycles than the sequence synthesized by Taq polymerase (Figure  2A ).