dataesr/scientific-paragraphs-categorization huggingface dataset:
Conclusions The present study shows that MALDI-TOF MS can be used to reliably identify tick species infected or not by Rickettsia spp without the use of a molecular method requiring DNA sequence information. It is important to note that no Rickettsia spp spectrum is available in the Bruker reference database and that this is the first analysis of Rickettsia strain by MALDI-TOF MS. This work also demonstrated that MALDI-TOF MS could be applied for the rapid detection of Rickettsia spp in Ticks removed from patients. The rapid determination of a tick's identity and it infectious status should guide decisions related to specific patient monitoring or the administration of preventive treatment. Additionally, the low consumable costs, minimal time required for sample preparation and rapid availability of the results of MALDI-TOF MS could be useful for epidemiological studies and the monitoring of tick-borne diseases via the dual identification of vectors and their borne pathogen in one step.
Parasite screening A total of 446 individuals were screened for the presence of haemosporidian parasites (genus Plasmodium and Haemoproteus) with a nested PCR. This reaction amplified a 524-bp-long fragment of the cytochrome b gene of the two parasite genera  (Waldenstrom et al. 2004 ). The primers used, HEAMF (5 0 -ATGGTGCTTTCGATATATG CATG-3 0 ) and HAEMR2 (5 0 -GCATTATCTGGATGTGA TAATGGT-3 0 ), were defined in  Bensch et al. 2000. This method is highly repeatable, with a lower limit of detection of one infected blood cell per 100,000. Samples from negative individuals were analyzed twice to avoid false negatives. We identified the parasite strain by sequencing PCR products and comparing these sequences to those on GenBank using the algorithm BLASTN (http://blast. ncbi. nlm. nih. gov/Blast.
Among We tested each population-locus combination for deviations from Hardy-Weinberg equilibrium and for linkage disequilibrium between pairs of microsatellite loci, using exact tests implemented in GenePop 4. 0  (Rousset 2008). Global deviation from Hardy-Weinberg equilibrium was investigated for each population using Fisher's exact tests. The assumption of neutrality of microsatellite loci was tested using the method implemented by  Beaumont and Nichols (1996)  in LOSITAN  (Antao et al. 2008). Withinpopulation genetic diversity of microsatellites was assessed by computing the expected heterozygosity (H E ) using the software Genetix  (Belkhir et al. 2004 ). Microsatellite allelic richness (A) was computed using a rarefaction index for the smallest sample size for one locus in one population (17 individuals in St Elven) using F STAT 2.
9. 3  (Goudet 1995). H E and allelic richness for microsatellites were compared between insular and mainland populations with Mann-Whitney tests. MHC allelic frequencies were calculated using Arlequin 3. 5  (Excoffier and Lischer 2010)  as the number of individuals carrying a certain allele divided by the total allele count observed in the population. Total allele count is defined as the sum of alleles found per individual in a population  (Loiseau et al. 2009). We note that this method of determining allele frequencies may underestimate the frequency of common alleles and overestimate the frequency of rare alleles  (Ekblom et al. 2007 ). Genetic diversity for MHC was calculated by the number of different alleles found in each population and was estimated using richness cumulative curves (index of allelic richness, R) with the software ESTIMATES 7.
Population differentiation Genetic differentiation between populations for microsatellite markers was measured using F ST and tested for each pair of populations with exact tests implemented in Gene-Pop 4. 0  (Rousset 2008). Genetic differentiation between populations based on MHC class I genes was assessed using the software Arlequin 3. 5  (Excoffier and Lischer 2010). We identified 95% confidence intervals of overall pairwise F ST for microsatellites by bootstrapping over loci in F STAT 2. 9. 3 (Goudet 1995). We assessed how many MHC pairwise F ST confidence limits overlapped and tested the significance of this distribution with a chi-square test. Although F ST is arguably the most reported statistic in population genetics, its use is still debated  (Hedrick 2005; Jost 2008; Meirmans and Hedrick 2011; Whitlock 2011 ), so we also calculated D values (D EST , Jost 2008) for both neutral and adaptive markers.
We calculated D EST using the online program SMOGD v. 1. 2. 5 (1000 bootstraps, Crawford 2010) and for MHC genes using SPADE (Chao and Shen 2010) (10. 000 bootstraps). The pairwise F ST and D EST values for both markers were highly significantly correlated (Mantel tests; respectively r = 0. 98, P < 0. 0001 for microsatellites and r = 0. 99, P < 0. 0001 for MHC). We therefore report only F ST in the rest of the manuscript, but supply D EST values in Table  S2. For each marker type, we tested for isolation by distance using a Mantel regression of F ST /(1-F ST ) against the natural logarithm of the geographical distance between populations  (Rousset 1997) , with statistical significance inferred in XLSTAT. A Mantel test was also used to characterize the relationship between pairwise microsatellite and MHC population differentiation (F ST ).
Acknowledgments We thank the volunteers for their help on the field:  M. Brenier ,  E. Cellier-Holzem , and  R. Guerreiro. We are very grateful to all people in the  KLIVV genetics laboratory  for their welcoming and especially  H. Brandl ,  G. Munimanda ,  W. van Dongen ,  P. Vegh , and  J. White  for their help. Financial support was provided by the  Burgundy University , the  ANR IRMAL , and the  CNRS program MIE. Permit  to band the birds and take blood samples was given by the  Museum National d'Histoire Naturelle  and the  Pr efecture du Morbihan et du Finist ere.
The ERK inhibitor did not affect VacA-induced vacuolating activity (Supplementary Figure  S1b ). Furthermore, ERK knockdown inhibited VacA-induced Cx43 increase and LC3-II generation (Figure  4g ). ERK knockdown, however, did not affect Cx43 mRNA level (Figure  4h ). We also monitored by confocal microscopy the effect of ERK knockdown on VacA-increased Cx43. VacA increased accumulation of Cx43 in control cells, whereas VacA did not increase accumulation of Cx43 puncta in ERKknockdown cells (Figure  4i ). More interestingly, ERK knockdown significantly suppressed VacA-induced cleavages of PARP and caspase-9 (Figure  4j ). These data indicated that GSH level controls ERK activation, and thereby regulates VacA-increased Cx43 and LC3-II generation as well as apoptosis. Rac1 signaling pathway regulates VacA-induced Cx43 increase Hotchin et al.
ACKNOWLEDGEMENTS We thank  K. Maeda  and  K. Hirano  for skillful assistance. This work was supported by  Grants-in-Aid for Scientific Research  from the  Ministry of Education, Culture, Sports, Science and Technology of Japan ,  Improvement of Research Environment for Young Researchers  from the  Japan Science and Technology Agency  and the  Cooperative Research Grant  of the  Institute of Tropical Medicine, Nagasaki University , 2014. JM and MN were supported by the  Intramural Research Program, National Institutes of Health, National Heart, Lung, and Blood Institute , and by  Takeda Science Foundation , respectively.
Samples were then submitted to 45 cycles (15 s at 95 °C and 45 s at 62 °C). Real time fluorescence was measured at the end of the extension step, every cycle and analyzed by the Light Cycler 480 software. Each extract was analyzed undiluted, log diluted and twice log-diluted (5 µL in 45 µL of DEPC water), in triplicates. Determined concentrations were expressed in genomic units (GU) per filtered volume. The detection limit (LOD) of the PCR was 10 GU per reaction.
For the q-PCR step, 5 µL of extracted DNA were added to 45 µL of Probes Master Mix Roche (Roche Diagnostics) containing 2× reaction buffer, 300 nM of each primer and 100 nM of probe. DEPC water was used as negative control. The PCR assay was performed with a Light Cycler 480. The activity of the DNA polymerase was released by heating at 95 °C for 10 min. Samples were then submitted to 45 cycles (15 s at 95 °C and 1 min at 60 °C). Real time fluorescence was measured at the end of the extension step, every cycle and analyzed by the Light Cycler 480 software. Each extract was analyzed undiluted and log-diluted (5 µL in 45 µL of DEPC water), in triplicates. Determined concentrations were expressed in GU per filtered volume. The LOD of the PCR was 2 GU per reaction.
For the q-PCR step, 5 µL of extracted DNA were added to 20 µL of Probes Master Mix Roche (Roche Diagnostics) containing 2X reaction buffer, 900 nM of each primer, 200 nM of probe and 0. 025 µg/µL of T4 Gene 32 Protein (10972983001, Roche Diagnostics). DEPC water was used as negative control. The PCR assay was performed with a Light Cycler 480. The activity of the DNA polymerase was released by heating at 95 °C for 5 min. Samples were then submitted to 45 cycles (15 s at 95 °C and 1 min at 60 °C). Real time fluorescence was measured at the end of the extension step, every cycle and analyzed by the Light Cycler 480 software. Each extract was analyzed undiluted and log-diluted (5 µL in 45 µL of DEPC water), in triplicates. Determined concentrations were expressed in GU per filtered volume. The LOD of the PCR was 0.
The present study has some limitations that have to be discussed. First, the cost-effectiveness analysis incorporated some international data when local data were missing. Due to transferring international data to the German health care setting, uncertainty surrounding those parameters arises. However, data were retrieved from countries with similar settings to Germany, and only in case no local data source could be identified. Second, cost data were based on the ASHIP data from the AOK Hesse (Ultsch et al. 21 ), which reflects a limited regional sample and therefore may not be representative for total Germany. Additionally, no cost data regarding different pain splits could be identified. Nevertheless, it is expected that these uncertainties surrounding the cost inputs were of minor impact on the results as demonstrated by the deterministic sensitivity analysis.
Epidemiological data Age-specific incidence rates of 1st episode of HZ were retrieved from a recent German study that analyzed data from 2005 to 2009 from the German Pharmacoepidemiological Research Database (GePaRD). 28 The authors conducted a retrospective cohort study analyzing data from about 7 million SHI insured individuals in Germany. Observed HZ incidence rates from the GePaRD ranged between 10. 4 to 12. 5 per 1,000 person-years. 28 Incidence rates from the GePaRD were reported separately for each year, therefore, for insertion in the model, a cumulative incidence was calculated for the period from 2005 to 2009. As shown in Table  3 , the resulting annual HZ incidence rates varied between 0. 66% for the age group 50-54 y and 1. 43% for individuals aged 100 y. Alternative data on HZ incidence rates from the Association of Statutory Health Insurance Physicians (ASHIP)  37  were used for sensitivity analyses.
37 The alternative annual HZ incidence rates varied between 0. 66% for age group 50-54 y and 1. 28% for individuals aged 100 y. 37  Recurrent rate of HZ episodes was set to zero in the model, as HZ cases which appeared recurrently more than one year later were considered as another incident case in the GePaRD incidence data. 28 Alternative values based on a matched cohort study in the USA  38  were used for sensitivity analyses, resulting in 0. 10% rate of HZ recurrence for age group 50-54 y and 0. 27% for individuals aged 100 y. 38  The proportion of HZ patients that developed PHN was based on GePaRD data as well. 28 As for the HZ incidence rates, an average PHN cumulative incidence per 1,000 person-years was calculated for the years 2005 to 2009. The model using a 1month cycle length is based on the 1-month definition of PHN to formulate its calculations.
37  based on the Association of Statutory Health Insurance Physicians (ASHIP) database of 2007/2008. 37 The authors reported a female proportion of 65% of HZ cases in the ASHIP sample, covering the AOK Hessen. 37 ender split for PHN of 81% females was identified in Weinke et al. 39  who collected data through telephone interviews with 11,009 respondents. The split between the different HZ and PHN pain states were obtained from the SPS pivotal trial study. 5 The latter provided information on pain severity levels using a questionnaire specifically developed to assess HZ and PHN associated pain (the Zoster Brief Pain Inventory, ZBPI). 5 Reported pain splits by age group are shown in Table  3. Further data for HZ and PHN pain split from published literature were considered in sensitivity analyses.
29, 40 Z and PHN duration information provided in the SPS study  5  was used in base case analysis. The SPS study confirmed that most patients experience a 30-day duration of HZ, as reported in the literature, and a 9 month average duration of PHN episode, respectively. 5 mographic data and mortality Demographic data were gathered from data of the German Federal Statistical Office. 41  Corresponding general population mortality (background mortality) was based on mortality tables from the German Federal Statistical Office. 42  Mortality due to HZ was taken from Ultsch et al. 37  based on the ASHIP database. Mortality rate based on the ASHIP database ranged between 0. 02 per 100,000 person-years for age group 50-54 and 3. 86 per 100,000 person-years for individuals aged 90 y No mortality was deemed to be associated with PHN.
Vaccine characteristics Vaccine efficacy data came from 2 pivotal clinical trials, the ZEST and SPS study. 5, 15 Using vaccine efficacy data from the 2 pivotal clinical trials-the ZEST and the SPS studies  5, 15  -both direct and indirect effects of the vaccine were included in the model. The direct effect of the vaccine relates to the to the decrease in both the incidence of HZ and the proportion of PHN per HZ case. The indirect effect relates to the number of PHN cases avoided through the decrease in the number of HZ cases. The values inserted in the model for direct and indirect vaccine effects on HZ and PHN are shown in Table  3. Furthermore, the duration of PHN is reduced through vaccination, and this ultimately affects the pain severity experienced by the patients, as they spend a shorter period of time in each painful PHN health state.
Utility values Considering the model structure, utility values were collected for all levels of HZ and PHN pain. In the absence of local disease specific data, international studies were screened to identify the more reliable sources for disease-specific utilities. 33, 46, 47 Therefore, data from a US-study analyzing the relationship of pain and quality of life in 385 individuals aged 65 y with pain caused by PHN  13  were used. Data in this survey was collected by a 15-items questionnaire, which covered pain intensity, pain interference and health-related quality of life. Quality of life was measured via the EuroQoL (EQ-5D) survey, which consists of 5 items related to mobility, self-care, usual activities, pain/discomfort and anxiety/depression. 13 Utility values according to the level of pain were retrieved from an observational survey, including 84 European patients with PHN.
20 Data in this study were collected with a questionnaire, which included among others, pain severity, measured by the modified short form brief pain inventory (mBPI-SF). Quality of life was assessed via the EuroQoL (EQ-5D) survey as well. 20 he disease specific utilities were used to calculate the proportion of the baseline utility (in this case, HZ with no pain) that needs to be subtracted to obtain the utility value for a given health state. These decrements were then applied to the age-specific utility for the cohort. The upper input parameters for burden of disease data, mortality, vaccination efficacy and utility values are summarized in Table  3. In the following, the cost parameters inserted in the Markov Model are described.
21 assessed hospitalization costs, using database insurance records reporting inpatient diagnosis and length of stay in hospital. The mean annual hospitalization costs per user were calculated by means of German Diagnosis Related Groups (G-DRG), amounting to €3,080. 85 and €3,890. 62 respectively for HZ and PHN patients. 21 Hospitalization rate was reported to be 3. 2% for HZ patients and 14. 6% for PHN patients aged 50 y  21. The mean annual medication costs per user for HZ patients aged 50 years, were estimated to be €55. 72 and for those with PHN €219. 79. 21 The costs included consideration of prescribed medication in relation to a German dermatological guideline. 50 edication groups considered were virostatics, specific immunoglobulin, topical analgesics, non-opioids, opioids, antidepressants and anticonvulsants.
As a result, the DNA sequence dataset we analyzed contains a large fraction of non-productive sequences, which are either out-of-frame or contain a stop codon. These sequences experienced no selection and owe their survival to the receptor expressed by the other chromosome. For this reason they provide us with the raw, unselected product of the generation process. We used such out-of-frame sequences from the naive subsample to infer the statistics of the VDJ recombination process, and the out-offrame sequences from the memory subsample to learn the statistics of hypermutations. McCoy et al  [19]  previously exploited these differences between in-and out-of-frame sequences in human BCR memory repertoire analysis. The naive productive sequences (in frame and with no stop codon) are expected to have passed a selection process before being admitted to the periphery (henceforth called initial selection, to distinguish it from selection following a recognition event).
Our inference of recombination scenarios for individual reads requires accurate knowledge of the germline sequence of all the V, D, and J genes. These genes have several alleles, and an accurate accounting of sequencing errors and somatic hypermutation events requires knowing which alleles are present on the two chromosomes of each individual. The existing databases do not provide such information, but list all alleles that have been detected, together with an estimate of the population frequency of these alleles  [28]. To address this problem, we have developed a method for identifying alleles directly from the sequence data for each individual. By accumulating patterns of mismatches between reads and database gene sequences that occur too often to be attributed to chance errors, we can detect the presence of alleles for each gene.
A. Distribution of recombination events The raw distribution of recombination events P pre before any selection takes place was inferred from the naive, non-productive sequence dataset, as explained above show the distribution of numbers of nucleotide insertions in the DJ and VD joints. These distributions are independent of the identities of the genes on either side of the junction, and the VD and DJ insertions are very similar. The selection process that acts on going from the primitively generated to the naive repertoire causes the mean number of insertions to decrease significantly. The bottom two panels (c,d) show the distribution of deletions from the V and J genes (negative deletions account for palindromic insertions). Deletions are gene-dependent and the plots show the deletion profile averaged over all genes (gene-dependent profiles are shown in Fig.
E. Hypermutations Upon recognition of an antigen, B-cell receptors undergo an affinity maturation process, by which their binding strength to the antigen is increased through the combination of random somatic hypermutations and selection. Thus, receptor sequences from antigen-experienced cells, such as memory cells, are expected to show the effect of somatic hypermutations, and we can use these sequences from our dataset to learn their statistics. Hypermutations appear in our sequence reads as mismatches with the genomic sequence. However, because the survival of a sequence in the memory repertoire depends on its affinity for a particular antigen, the statistics of its hypermutations should reflect other factors than just the hypermutation process itself. To overcome this issue, we make the assumption (as in  [19] ) that when the hypermutation machinery is activated in a cell, it acts on both chromosomes indifferently.
For all properties the actual numerical values used to calculate the correlations are listed in the inset tables. (h) charge A 0 L 0 R 1 K 1 N 0 M 0 D -1 F 0 C 0 P 0 Q 0 S 0 E -1 T 0 G 0 W 0 H 0 Y 0 I 0 V 0 (i ) pH A 0 L 0 R 2 K 2 N 0 M 0 D -2 F 0 C -2 P 0 Q 1 S -1 E -2 T -1 G 0 W 1 H 1 Y -1 I 0 V 0 -0. 5 0 0. 5 (j ) polar A 0 L 0 R 1 K 1 N 1 M 0 D 1 F 0 C 0 P 0 Q 1 S 1 E 1 T 0 G 0 W 1 H 1 Y 1 I 0 V 0 -0. advances. Thanks to our stochastic framework, we were able to identify heterozygous genes as well as their alleles, even when these were not present in existing databases. We could then reconstruct the partition of these alleles  into two chromosomes. We found no significant bias in chromosome usage. Perhaps the most important difference between B-and T-cell receptors is the existence of hypermutations accumulated during affinity maturation.
Acknowledgements. We thank  Harlan Robins  for the BCR data without which this work would not be possible. We also thank  Mr. Shai Chester  for assistance in developing our allele-finding algorithms. The work of YE, QM, TM and AW was supported in part by grant ERCStG n. 306312. The work of ZS and CC was supported in part by  NSF  grants  PHY-0957573  and  PHY-1305525.
Although several proteins involved in this process have been identified, the detailed mechanisms triggering the DNA ejection and the precise role of UL36 are still unknown  (22). Finally, UL36 has been involved in the last stages of viral assembly. A null mutation in the UL36 gene leads to an accumulation of unenveloped DNA-filled capsids in the cytoplasm and a massive production of defective "light particles" lacking capsid, thus suggesting that UL36 is necessary for associating the preassembled outer viral layers to the capsids  (25, 26). The outer tegument protein UL48 has been shown to interact with The atomic coordinates and structure factors (codes 4TT0 and 4TT1) have been deposited in the Protein Data Bank (http://wwpdb. org/). 1 To whom correspondence may be addressed. E-mail: stephane.
EXPERIMENTAL PROCEDURES Sequence Analyses-The protein sequence alignments of HSV-1, HSV-2, PRV,  3  VZV, GHV-2, and CHV-1 UL36 were performed using ClustalW and were displayed with ESPRIPT. The coiled coils were predicted with PCOILS. The secondary structure predictions were performed using SOPMA. Cloning, Expression, and Purification of UL36 Fragments-Amino acids 1600 -1733 and 760 -1733 of strain 17 HSV-1 UL36 (accession number FJ593289. 1) were cloned between the BamHI and EcoRI restriction sites of the pGEX-6P1 (GE-Healthcare) expression vector, resulting in fusions of these fragments with an N-terminal cleavable glutathione S-transferase. Both fragments were produced in Escherichia coli BL21 (DE3) cells (Stratagene). Freshly transformed cells were incubated overnight in 5 ml of LB supplemented by ampicillin at 37 °C.
Crystals were grown by vapor diffusion at 20 °C in hanging drops using 0. 1 l of protein and 0. 1 l of the well solution containing 0. 2 M (NH 4 ) 2 HPO 4 , 20% PEG 3350 (native dataset), or 100 mM HEPES, pH 7. 5, and 2 M ammonium formate (KI soak dataset). In both cases, small hexagonal crystals grew within 1-4 days and were harvested 2 weeks later. The native crystals were flash-cooled in 0. 2 M (NH 4 ) 2 HPO 4 , 20% PEG 3350, 20% glycerol, although the derivative crystals were incubated for 10 min in HEPES, pH 7. 5, 1 M ammonium formate, 0. 8 M potassium iodide, 25% glycerol before flash-freezing. A native dataset was collected at 100 K on a single crystal at the PROXIMA 1 beamline at the Soleil synchrotron. A highly redundant iodine SAD dataset was collected at 100 K on a single crystal at a wavelength of 1.