dataesr/scientific-paragraphs-categorization huggingface dataset:
RSM can be used to optimize the operating parameters by coupling various studied responses  (Benoist et al. , 1994). α (axial distance) = 4 √ 2 N. In the present case N = 2 and α = 1. 4142. In the present study DIC operating parameters were: saturated steam pressure (P) ranged from 0. 2 MPa to 0. 7 MPa and treatment time (t) ranged from 20 to 120 s. The responses regarding extraction kinetics and extraction yields were analyzed. Statistical treatment of obtained results was executed using the analysis design procedure of Statgraphics Plus software for Windows (version 16. 0, Levallois-Perret, France).
Gas chromatography analysis GC-MS analyses were performed by using an Agilent (Kyoto, Japan) gas chromatography. The instrument was equipped with a BD-EN14103 capillary column 30 m × 320 μm × 0. 25 μm (Agilent). The velocity of the carrier gas (He) was at 33 cm s -1. Injection of 2 μl of the various samples was carried out with a split mode (ratio 1:20) and the injector temperature was set at 250 • C. The oven temperature increased from 50 • C (1 min) to 180 • C at a rate of 20 • C min -1 , increased from 180 • C to 220 • C at a rate of 2 • C min -1 , and then held at 230 • C for 10 min. The mass spectra were recorded at 3 scan/s between 50 to 400 amu. The ionization mode was electron impact (EI) at 70 eV. Identification of common fatty acids was performed using the NIST'98 [US National Institute of Standards and Technology (NIST), Gaithersburg, MD, USA] mass spectral database.
Impact of DIC parameters on oil extraction yields To determine and optimize DIC operating parameters this time vis-à-vis the final extracted lipid yield, experimental results were analyzed using Statgraphics software. The mathematical relationship obtained was a polynomial empirical model representing the quantitative effect of process variables and their interactions on the measured response. The values  of the coefficients of P and t were related to the effect of these variables on the response. Statgraphics software tested the statistical significance of each effect by comparing the mean square against an estimate of the experimental error. In this case, 2 effects have P-values less than 0. 05, indicating that they are significantly different from zero at the 95% confidence level. Pareto chart of standardized effects (Fig.
Additional neuropsychological testing included the Token-test  (De Renzi and Vignolo, 1962) , the consortium to establish a registry for Alzheimer's disease (CERAD)  (Welsh et al. , 1994)  and the Parkinson neuropsychometric dementia assessment (PANDA)  (Kalbe et al. , 2008). No other severe neurological or psychiatric illness was reported. Twenty (10 males) right-handed healthy adults, who matched the patients in age (M = 66. 4, SD = 7. 8) and education (M = 14. 4 years; SD = 3. 0) formed the control group. Healthy controls had no history of neurological or psychiatric disorders, showed no hearing impairments and did not actively practice music. All participants, recruited via the database of the Max Planck Institute for Human Cognitive and Brain Sciences, Leipzig, Germany, gave informed consent and were remunerated for their participation.
PERCEPTUAL TIMING TASKS The first three tasks (duration discrimination, anisochrony detection with tones, and anisochrony detection with music) allow estimating thresholds of duration discrimination of two tones and to detect an interval embedded in an isochronous sequence of tones or in a musical excerpt. Thresholds are estimated using a maximum-likelihood adaptive procedure (MLP)  (Green, 1993)  implemented in the MLP toolbox  (Grassi and Soranzo, 2009)  in MATLAB. Participants performed 3 blocks of 16 trials each. In each trial, the stimulus difference was changed adaptively depending on the participants' response. Thresholds corresponded to the midpoint of the psychometric curve defined as a probability of 63. 1% of correct detection  (Grassi and Soranzo, 2009). Stimuli were delivered via headphones (Sennheiser HD201) at a comfortable sound pressure level.
GAIT ASSESSMENT Gait kinematics in the absence of auditory cues was assessed with a Vicon MX Motion Capture System during the second day of testing. Sixteen passive reflective markers (14 mm) were attached to participants' lower body (four on the hip, three on each leg and foot, respectively) in accordance with the Conventional Gait Model  (Baker, 2006). Participants were asked to walk for 1 min at their spontaneous walking speed. The trial was repeated twice. The performance was recorded using Vicon Nexus Software.
BAASTA Before submitting data to the following analyses, trials were screened for outliers (e. g. , for perceptual tasks, blocks with a higher false alarm rate higher than 30%; for motor tasks, taps with ITI deviating by more than three times the interquartile range from the median). A low number of outliers was found in both patients and controls. Overall in the perceptual tasks, 3. 3% of the trials were rejected for patients and 3. 8% for controls. In the motor tasks, <1% of taps were discarded for both patients and controls. The effect of training on perceptual and motor timing abilities was examined for the tasks of the BAASTA where patients showed impaired performance relative to controls pre-training. The mean results obtained in these tasks are shown in Figure  1  (perceptual tasks), and Figure  2  (motor tasks).
DATA ANALYSIS For each participant, we recorded intensity and pleasantness ratings (0-100) and sniff parameters on 20 occasions per odor (T1 to T20, for anise and chocolate). Sniffs (see Figure  1B ) were preprocessed by removing baseline offsets and aligned in time by setting the point where the sniff entered the inspiratory phase as time zero. Sniff maximum flow rate, duration, and volume (see Figure  1B ) were calculated for the first sniff of every trial, for every participant. Before analyzing how the ratings and sniff parameters changed with repeated exposure, outliers defined as values exceeding three standard deviations from the participant's mean were removed (0. 65% of the trials). Then, analyses of the time-related changes in ratings and sniff parameters were performed (1) at the group level, by comparing time-related changes of "likers" (participants giving the highest pleasantness scores) and "dislikers" (participants giving the lowest pleasantness scores), and (2) at the individual level, by correlating each participant's initial pleasantness rating at T1 with the time-related changes across the trials T1 to T20.
DATA ANALYSIS As the participants explored the environment, the position and orientation of their heads were recorded on average every 60 ms (15 Hz) and subsequently used to calculate their paths taken during the different phases of the experiment (Figure  1 ). From these recordings, we assessed search time, i. e. , the time to reach the hidden target, path length, and boundary crossings for all the trials. In order to study the effect of the tests, the following performance measures were also calculated: 1. Percent quadrant time: Amount of time participants searched in a virtual quadrant (i. e. , 25% of total exploration area). 2. Boundary crossing per quadrant: Number of times the participant crossed the boundary of the exploration area per virtual quadrant. For the tests and probe trial we computed the spatial distribution of the time elapsed in the exploration area, i.
EEG RECORDING AND DATA PREPROCESSING EEG was recorded with 64 electrodes (actiCAP, Brain Prod-uctsGmbH, Germany). The EEG was digitized at 500 Hz. EEG analysis was conducted using EEGLAB  (Delorme and Makeig, 2004)  and custom-built Matlab scripts. The data were resampled offline to a 250 Hz sample rate, low band-pass filtered at 45 Hz to remove line noise. Epochs were generated from -500 to 980 ms relative to stimulus onset, with a 200 ms prestimulus baseline correction. Initial artifact rejection was conducted in a semiautomatic manner (in EEGLAB) by rejecting epochs with activity above 100 µV or below -100 µV, as well as rejecting trials where activity at any time point for any electrode was more than 5 standard deviations from the mean activity for that epoch. Any channels that contributed to the rejection of many epochs were considered for removal and later interpolation.
Frontal channels that showed large amplitude blink activity were also excluded from further analysis during this this first semiautomatic artifact rejection. Ocular artifact correction was conducted in EEGLAB in Matlab using independent component analysis  (Delorme and Makeig, 2004). Following removal of eye blinks and eye movements, noisy channels were replaced by an interpolated weighted average from surrounding electrodes. Data were then re-referenced to the common average. A final round of semiautomatic artifact rejection with a threshold of +/-80 µV was used to remove any remaining artifacts. All ERPs are presented with a low-pass filter of 20 Hz for visual presentation purposes. An ANOVA with the factors congruency (congruent, incongruent) and stimulus contrast (C0, C1) was run on ERPs averaged for each participant.
PSYCHOPHYSICAL RESULTS We analyzed our data dependent on participants' contrast discrimination. Discrimination performance (d ) was lower in the congruent condition (d congruent: 1. 22, SD = 0. 46) than in the incongruent (d incongruent: 1. 42, SD = 0. 45) condition. A one factor repeated measures analysis of variance including the factor of congruency showed this effect of congruency on d to be significant (F (1,14) = 5. 36, p = 0. 03). At the same time, the criterion (c) was not different in the two congruency conditions (1 factor repeated measures ANOVA; F (1,14) = 0. 165, p = 0. 69). A bias free measure of sensitivity (A ) also confirms our finding. Sensitivity appears to be better for incongruent trials (A : 0. 81, SD = 0. 075) than for congruent trials (A : 0. 79, SD = 0. 067) (F (1,14) = 4.
However, note that the psychophysical effect corresponds to an interaction between contrast and congruency: the discrimination between the two contrast levels is more difficult in congruent than in incongruent trials. If sensory attenuation assessed with psychophysical methods were a direct reflection of the ERPs assessed at the same time, we would have expected to see an interaction between these two factors in our ERP data as well. However, this was not the case. Of course, it might be that the ERP data simply lack sufficient statistical power. However, it is also possible that psychophysical and ERP indices of sensory attenuation (at least those assessed in the current experiment) are not in a simple one-to-one relationship. Perception might be dependent not only on early cortical responses, but also on later processing and/or recurrent processing, tweaking the relationship between perceptual measures and observable neurophysiological measures.
The film was then digitized and coded frame by frame by two trained research assistants who were naı ¨ve to the hypotheses under investigation. One assistant coded the videos of all infants, while a second coder scored 50% of the data to verify the reliability of the codes. Intercoder reliability exceeded 0. 90. To be considered in the final analysis, during each trial infants had to look at the stimuli for a minimum of 25% of each trial duration and for a minimum of 5% toward each video of the sideby-side stimuli presentation. In all Experiments, all participants met these criteria for inclusion. We computed four preference scores by dividing the looking time to one face (German talking face or French talking face) by the amount of total looking time (sum of looking times to the German and French talking faces) separately for the baseline and test trials.
These scores were then converted to percentages. For all subsequently performed ANOVAs, these four preference scores were then used as two dependent variables, ''Baseline'' and ''Test'' depending on the auditory-only familiarization (French, German). These variables only included the audio-matching preference scores on either the German or French talking faces in baseline and test trials, respectively. Because preliminary analyses in all experiments did not reveal any significant effects of infants' gender or of the bilingual speakers' identity on infants' looking times, the data for these two factors were collapsed in all analyses.
Method Participants. The sample consisted of a total of 88 monolingual German-learning infants. All infants were full-term with no visual or auditory deficits, as reported by parents. The data from nine additional infants were discarded from the final sample due to equipment failure (n = 2) or due to extreme fussiness (n = 7). There were forty-three participants in the 6-month-old group (mean age = 195. 0 days; SD = 8. 2 days; 20 females), and forty-five participants in the 12-month-old group (mean age = 368. 7 days; SD = 12. 1 days; 28 females). Stimuli. All stimuli were identical to Experiment 1a. Additionally, for audio-visual (i. e. , in-sound) stimuli, the same video clips were presented at conversational sound pressure level (65 dB 65 dB). Apparatus and procedure. After showing an attention getter, we used the intersensory matching procedure.
Acknowledgments The authors are indebted to the parents who donated their time to participate with their infants in the study. We are grateful to  Nicole Deisenroth ,  Janika Epe , and  Franziska Schwarzer-Geraedts  for their assistance with data collection and coding. We would also like to thank  David J. Lewkowicz  and two anonymous reviewers for their constructive comments on an earlier version of this paper.
Studied population and missing data Variables collected at baseline were described with median and range for continuous variables and percentage for qualitative variables, with percentage of missing data. No imputation was performed on missing items. Scores were calculated if at least half of the items were answered according to the recommendations of the EORTC scoring manual  [23]. No imputation was performed on missing scores. MCA and LLRA were both performed on patients with all items of the studied dimension (each dimension of the QLQ-C30 and the QLQ-BR23) filled out at the then-test and the pre-test measurement time points and with a MCID between then-test and pre-test of at least 5 points for the given dimension. This selection was done in order to retain a clinically meaningful difference of the recalibration occurrence.
Changes occurring at the first principal component are considered as major and those occurring at the second principal component as minor. In this way, changes were qualified in the first axis of "major reprioritization" and in the second axis of "secondary reprioritization". The study was limited to the first two principal components, according to the Scree test  [37]. Reconceptualization was reflected by a change in the structure of the graph of correlations between scores and principal components, as well as in the connection or opposition of some scores. Concerning the module BR23, sexual enjoyment and hair loss were excluded from the analysis given the number of missing values. All analyses were performed with R software  [38]  using FactoMineR library for factor analyses  [39]  and eRm library for LLRA  [34, 35, 40].
Insertion Protocol and Insertion Force Measurements. Frictions forces between the array and the cochlea have been recorded with a 6-axis force sensor (ATI Nano 17, calibration type SI-12-0. 12, resolution: 3 mN, Apex, NC). Sensor data have been recorded in real-time via the same analog to digital interface card controlling the actuator input power at (ii) insertion with the Hifocus 1J electrode insertion tool (Figure  3 (b)), a commercially available tool distributed with Hifocus 1J and helix array; it is composed of a handle comprising a flexible shaft connected to a slide that can eject out of an insertion tube the array by pushing its silicon jog. We have been using the metal insertion tube (AB-6135, Advanced bionics, Valencia, CA) in this study. This tool was held manually during insertion; (iii) insertion with an in-house made motorized insertion tool (Figure  3(c) ).
We found that force peaks of the motorized insertion on third position were significantly different compared to measurement of first and second insertion ( = 0. 0362). Thus, third insertion could not be used for analysis and all data collected during the third insertion were discarded in all temporal bones. Consequently, insertions forces data were used for analysis in 10 manual insertions, 8 Hifocus 1J electrode insertion tool insertions, and 8 motorized insertions.
An increase in the time spent by the animals in the light chamber was interpreted as a sign of passive avoidance. Statistical analyses. Animal treatments were randomized within each cage. For the BDNF lentiviral vectors and rBDNF experiments, we performed a two-way ANOVA followed by a Student-Newman-Keuls post hoc test for multiple comparisons using the Sigmastat software. For the CTX-B experiments, we performed three-way ANOVAs, followed by a Student-Newman-Keuls post hoc test for multiple comparisons. ANOVA results are given as the value of the Fisher distribution Fx,y obtained from the data, where x is the degrees of freedom for groups and y is the total degrees of freedom of the distribution. Data were expressed as mean Ϯ SEM, and the levels of significance were set as follows: *p Ͻ 0.
Electrophysiological stimulation protocol The stimulating electrode was placed at the border between the white matter and layer VI to stimulate vertical pathways projecting to cortical layers II/III and IV  (Woodward et al. , 1990; Kenan-Vaknin and Teyler, 1994). Field potentials were recorded at 32°C in the current-clamp mode of a MultiClamp 700B amplifier (Molecular Devices). Signals were filtered at 10 kHz and digitized at 50 kHz with an analog signal converter (Digidata 1322A; Molecular Devices), and analyzed using pClamp 9. 2 software (Molecular Devices). The stimulation electrode was set at 0. 2-1 mA for 200 s, with a stimulation intensity that yielded a half-maximal response. Baseline responses were obtained every 20 s over 10 min. LTP was induced by TBS: three trains of 10 theta bursts (with 200 ms interbursts) separated by 15 s, with 1 theta burst consisting of four pulses (0.
2 ms/pulse) at 100 Hz  (He et al. , 2009). Post-TBS acquisition was performed over a 35 min period with stimulations every 20 s. Recorded data were analyzed using the Clampfit, Igor pro, and Excel software programs. The amplitude of the maximum negative field potential in layer II was used as a measure of the evoked population excitatory current. The last 5 min of the recordings were averaged and compared with the baseline; all recordings with an average greater than twice the noise of the baseline were considered as LTP. Recordings whose baseline varied by Ͼ10% were excluded. Data were then pooled together for each experimental group and expressed as the percentage change from the control baseline Ϯ SEM. Statistical comparisons were performed by applying a Mann-Whitney rank sum test. Differences were considered significant at p Ͻ 0.
To count the number of positive cells, we performed a cell counter analysis using the ImageJ software. The same regions of interest were used for each animal as illustrated in Figure  1D. Cell counting was performed separately for the ipsilateral and contralateral sides of the brain for each animal. Statistical significance was determined by two-way ANOVA followed by Student-Newman-Keuls comparison post hoc test, using the Sigmastat software. The first pa-rameter was the treatment of the animals (saline vs CFA) and the second parameter was the cortex side counted (ipsilateral vs contralateral side). ANOVA results are given as the value of the Fisher distribution Fx,y obtained from the data, where x is the degrees-offreedom for groups and y is the total degrees of freedom of the distribution.
ACKNOWLEDGMENTS This work was supported by grants from  ANR (Agence Nationale pour la Recherche ;  Project Démences et Utilisation d'Outils/Dementia  and  Tool Use ,  N • ANR 2011 MALZ 006 03 ), and was performed within the framework of the  LABEX CORTEX  ( ANR-11-LABX-0042 ) of  Université de Lyon , within the program " Investissements d'Avenir"  ( ANR-11-IDEX-0007 ) operated by the  French National Research Agency (ANR).
All participants were given a semi-structured questionnaire that had been specially designed for the study in order to rate demographic data (gender, age, education and occupation) (cf. Table  1 ). The stimulating quality of the participants' occupations was rated as follows: a value of 1 was attributed for jobless individuals; 2 for manual workers; 3 for office workers, publicsector workers, tradespeople, and storekeepers;4 for the self-employed, executive managers, and students; and 5 for physicians and company directors. The musicians were also asked to evaluate their age at training onset, and its average frequency for the last 6 months. Based on this questionnaire, the groups of participants were matched on gender, age, education, and occupational level. Moreover, middle-aged and older musician groups differed only marginally on age at training onset as well as frequency of practice (cf.
Statistical analyses All the statistical analyses were performed with STATISTICA software  (StatSoft,(2011). STATISTICA (data analyses software), 10 th version www. statsoft. fr). The weakest significance threshold was set at p=. 05,uncorrected for multiple comparisons. Twochi-squaretests of independence were run to attest that the musician and nonmusician groups did not differ significantly in gender frequencies. The first compared all four groups (middle-aged controls, middle-aged musicians, older controls, and older musicians), and the second tested themusician and nonmusician groups, collapsed across age (cf. Table  1 ). Twoanalyses of variance (ANOVAs) were conducted to attest that the four groups did not differ significantly on education and occupation level (cf. Table  1 ). For these variables, we also ran two ttests, where themusician and nonmusician groups were collapsed across age.
Acknowledgements This research was supported by a  French Ministry of Research  grant. We would like to thank  Alice RODAIX ,  Isabelle RAMEAU ,  Elhia DEMBA ,  Manuella BOISSET , and  Camille HOU  for their help in collecting data and their investment in this project. Many thanks, too,to  Mona LEBLOND  for helping with the English. Finally, this paperwassubstantiallyimproved during reviewing, and we would like to address ourgratitude to the two anonymous referees for their crucial help.
ANIMALS AND HOUSING CONDITIONS Forty male and thirty-five female rhesus monkeys (Macaca mulatta) and forty male and eighty female cynomolgus monkeys (Macaca fascicularis) living in three distinct Chinese breeding farms were included in this study. Data from the cynomolgus populations have been previously published in  Camus et al. (2013a,b)  whereas data from the rhesus monkeys are newly presented here. The animal information and housing conditions are described in Table  1. Toys and fruits were provided as environmental enrichment, as well as swings in the social groups. A schematic plan of a common social-housing cage is available in our previous paper (see Figure  S1  in  Camus et al. , 2013b).
Single-housing (Camus et al. , 2013a) For each room, data were collected in a randomized order at three time points (morning, noon, and afternoon), on 2 nonconsecutive days (6 sessions per individual). The observer was facing the door or window at all times rather than the cages and looked at the camera screen rather than at the individuals (a gaze directed to a macaque's eyes being interpreted as a threat).