dataesr/scientific-paragraphs-categorization huggingface dataset:
Meta-analysis and statistics For meta-analysis, we selected studies for which data were available for repeats of size 24 or more  [17]. Association between the number of repeats and disease status was analysed by chi-square and Fisher's exact tests, with the samples split into two categories: samples with an intermediate expansion of 24-30 repeats and the remainder including those with repeat length > 30. Heterogeneity across studies was examined using the I-squared statistic and chi-square-based Q-tests. The Q-test power was calculated using the Jackson method  [25]. Both fixed-effect and random-effects models were used. Sensitivity analyses were conducted to evaluate the influence of individual results on the pooled estimate after sequential removal of each study. Potential publication bias was assessed by both Begg and Egger tests.
Alzheimer's Disease Neuroimaging Initiative Data used in the preparation of this article were obtained from the Alzheimer's Disease Neuroimaging Initiative (ADNI) database (adni. loni. usc. edu). The ADNI was launched in 2003 as a public-private partnership, led by Principal Investigator Michael W. Weiner, MD. The primary goal of ADNI has been to test whether serial magnetic resonance imaging (MRI), positron emission tomography (PET), other biological markers, and clinical and neuropsychological assessment can be combined to measure the progression of mild cognitive impairment (MCI) and early Alzheimer's disease (AD).
C9orf72 repeat expansion distribution in the new cohort For the British dataset, the median age of ALS onset was 62. 4 years and for controls, 60. 1 years, with a malefemale ratio of 62:38. In the ADNI dataset the median age was 73. 2 and the male-female ratio 49:51. We identified no difference in the distributions of repeat size between ALS and controls in the 1-23 repeat interval (Kolmogorov-Smirnov test, p-value = 0. 96). Nine cases and one control were found to carry 24-30 repeats. Five controls were found to carry expansions whose length is rarely seen in unaffected controls and among which a few were hundreds of repeats long (Fig. 1  and Additional file 1: Table  S1 ). Specifically, five British controls carried large expansions up to hundreds of repeats (sizes 60, 357, 406, 413 and 484) and one British control carried an intermediate repeat expansion of 26 repeats.
Meta-analysis of the 24-30 repeat expansion A literature review  [17]  identified four studies which provided original data on the number of patients carrying 24-30 repeats, including one study from North America  [23] , one study from Italy  [16] , one study from France  [15] , and one study from Spain  [13]. We extended the review to all articles published up to January 2019 and identified one further study  [28]  providing data suitable for our meta-analysis. However, in this additional study, most of the ALS cases overlapped with our British dataset, and we therefore excluded it from our analysis. Therefore, after reproducing the previously published meta-analysis, we performed two new meta-analyses. In the first we added our British samples (1295 cases and 340 controls) and in the second we added our British/ADNI dataset (1295 cases and 613 controls).
The final dataset used in this study comprised 5071 cases and 3747 controls. We were able to reproduce the association between the C9orf72 24-30 repeat expansion and ALS using the previously used four datasets (fixed-effect model odds ratio = 5. 13, 95% confidence interval = 1. 28-22. 24, p-value = 0. 03) (Table  1A ). The association was replicated after including our cohorts, both when the British cohort was added (fixed-effect model odds ratio = 4. 01, 95% confidence interval = 1. 21-13. 26 p-value = 0. 02) (Table  1B ) and when the British/ ADNI dataset was added (fixed-effect model odds ratio = 4. 82, 95% confidence interval = 1. 45-15. 96, p-value = 0. 01) (Table  1C ). The use of the fixed-effect model was suggested by the chi-square Q-test which did not show heterogeneity (p-value = 0.
61), but this was underpowered to detect heterogeneity (power = 12. 07%), suggesting that a randomeffects model is a more appropriate model to use. Under a random-effects model, we did not show association using the four studies from the previous meta-analysis only (p-value = 0. 07). Using the UK/ADNI dataset we were able to show this association under the random-effects model hypothesis also (odds ratio = 4. 2, 95% confidence interval = 1. 23-14. 35, p-value = 0. 02) (Table  1C ). No evidence of publication bias from either the Egger (p-value = 0. 69) or Begg (p-value = 0. 14) test was observed (Additional file 1: Figure  S1 ). Sensitivity testing, recalculating the pooled odds ratios by omitting one study per iteration (Additional file 1: Table  S2 ) showed that results were consistent, without major fluctuations.
Association between C9orf72 intermediate expansion on the aggregated dataset We also investigated the association between the intermediate repeats and ALS in a final dataset generated from the aggregation of the 5 datasets used for the Heterogeneity chi-squared Q = 1. 80 (d. f. = 4), p = 0. 77, power = 13. 01% I-squared (variation in OR attributable to heterogeneity) = 0. 0% (0. 0-53. 7%) Fixed-effect Test of OR = 4. 82: z = 2. 57 p = 0. 01 Random-effects Test of OR = 4. 20: z = 2. 29 p = 0. 02 meta-analysis (5071 cases and 3747 controls). The association was demonstrated both with the chi-squared test (p-value = 5 × 10 -4 , odds ratio = 14. 83, 95% confidence interval 1. 99-110. 57) and the Fisher's exact test (p-value = 2 × 10 -4 ). Considering that the frequency of the C9orf72 expanded repeat greatly varies across European populations  [5, 7, [13] [14] [15] 31] , we attempted to investigate whether this is also true for the intermediate repeats.
To do so we split the aggregated dataset cases in two groups, 3015 northern European patients (Americans, French and British) and 2056 southern European patients (Italians and Spanish). Both the chi-squared test (p-value = 5 × 10 -3 , odds ratio = 6. 17, 95% confidence interval 1. 43-26. 61) and the Fisher's exact test (p-value = 5 × 10 -3 ) showed a significant difference between the two groups. On the basis of intermediate repeat heterogeneity between the two groups, we tested the association between the intermediate repeats and ALS in the two groups separately. For the northern European dataset (3015 cases and 2637 controls) the resulting odds ratio was 15. 83 (95% confidence interval 2. 11-118. 67, chi-squared p-value = 3 × 10 -4 and Fisher's exact test p-value = 10 -4 ). In the southern European dataset (2056 cases and 1110 controls) the association was not significant (Fisher's exact test p-value = 0.
Discussion We have shown that intermediate repeat expansions in the C9orf72 gene are associated with ALS. An expanded hexanucleotide repeat in C9orf72 is the most common cause of ALS. While expansions of hundreds of repeats in size are clearly associated with the disease, a lower limit has not yet been identified. A few studies have suggested that 24-30 repeats might be sufficient to cause the disease  [12, 17]. However, considering that the statistical power of the datasets used was limited, further investigation was essential. We have added 1295 cases and 613 controls to the meta-analysis and demonstrated the association of such intermediate expansions with ALS both under a fixed-effect (p-value = 0. 01) and a random-effects model (p-value = 0. 02). The latter tells us that the association between 24 and 30 repeats and the disease exists without assuming that the effect of this mutation is homogenous across the patient groups.
This up-todate figure obtained from a meta-analysis on 5071 ALS patients and 3747 controls from 5 populations (UK, USA, Spain, Italy and France) provides definitive evidence for the association between intermediate repeats and ALS (p-value = 2 × 10 -4 ) and expands the role of C9orf72 in the development of ALS by enlarging the number of cases that can be explained, with direct implications for research and clinical practice. Indeed, we suggest the threshold to discriminate between neutral and pathogenic C9orf72 expansions should be lowered to 23, considering as pathogenic, expansions of 24 or more repeats, in agreement with previously reported observations  [5, 17]. Expansions of 20-23 repeats have contradictory evidence of association with ALS. This suggests that if such an association exists, their contribution to disease risk is weaker than for longer expansions, and we need larger datasets to investigate it.
Acknowledgments This is an  EU   Joint Programme -Neurodegenerative Disease Research (JPND)  project. The project is supported through the following funding organisations under the aegis of JPND -www. jpnd. eu( United Kingdom ,  Medical Research Council  ( MR/L501529/1 , AAC PI; MR/R024804/1, AAC PI) and  Economic and Social Research Council  ( ES/L008238/1 , AAC co-PI)) and through the  Motor Neurone Disease Association. This study represents independent research part funded by the  National Institute for Health Research (NIHR) Biomedical Research Centre at South London and Maudsley NHS Foundation Trust and King's College London. The work leading up to this publication was funded by the  European Community 's  Horizon 2020 Programme  ( H2020-PHC-2014 -two-stage; grant agreement number  633413 ).
16 0. 90 19. 14 100. 0 Heterogeneity chi-squared Q = 1. 82 (d. f. = 3) p = 0. 61, power = 12. 07% I-squared (variation in OR attributable to heterogeneity) = 0. 0% (0. 0-74. 8%) Fixed-effect Test of OR = 5. 13: z = 2. 18 p = 0. 03 Random-effects Test of OR = 4. 16: z = 1. 83 p = 0. 07 B -Y. Chen et al. meta-analysis datasets + our British dataset Garcia-Redondo et al. 1/0 781/248 0. 95 0. 04 23. 52 19. 5/14. 7 Ratti et al. 1/0 1275/862 2. 03 0. 08 49. 89 15. 3/14. 7 Rutherford et al. 3/0 995/1444 10. 19 0. 53 197. 45 10. 5/17. 2 Millecamps et al. 6/0 725/580 10. 49 0. 59 186. 57 14. 2/18. 2 British dataset 9/1 1295/340 2. 37 0. 30 18. 79 40. 5/35. 2 M-H pooled (fixed-effect) 20/1 5071/3474 4. 01 1. 21 13. 26 100. 0 M-H pooled (random-effects) 20/1 5071/3474 3. 41 1. 00 11. 66 100. 0 Heterogeneity chi-squared Q = 2.
00 (d. f. = 4), p = 0. 74, power = 12. 83% I-squared (variation in OR attributable to heterogeneity) = 0. 0% (0. 0-58. 4% Fixed-effect Test of OR = 4. 01: z = 2. 28 p = 0. 02 Random-effects Test of OR = 3. 41: z = 1. 96 p = 0. 05 C -Y. Chen et al. meta-analysis datasets + our British/ADNI dataset Garcia-Redondo et al. 1/0 781/248 0. 95 0. 04 23. 52 20. 7/14. 7 Ratti et al. 1/0 1275/862 2. 03 0. 08 49. 89 16. 3/14. 7 Rutherford et al. 3/0 995/1444 10. 19 0. 53 197. 45 11. 1/17. 2 Millecamps et al. 6/0 725/580 10. 49 0. 59 186. 57 15. 0/18. 2 British/ADNI dataset 9/1 1295/613 4. 28 0. 54 33. 88 36. 9/35. 2 M-H pooled (fixed-effect) 20/1 5071/3747 4. 82 1. 45 15. 96 100. 0 M-H pooled (random-effects) 20/1 5071/3747 4. 20 1. 23 14. 35 100. 0.
The test set-up is illustrated in Fig. 3. The impact hammer and the accelerometer were connected to a signal conditioner (PCB model 482C15) itself connected to a data acquisition device (NI USB-6356) connected to a computer. The applied force and the acceleration were recorded with a sampling frequency of 1 MHz by using a MATLAB application which was specifically developed for this test. The test was repeated at five temperatures (-20, 0, 15, 35 and 50 °C) and five impacts were applied at each temperature. Fig. 4  shows an example of the time domain signals for the five hits at 15°C. The experimental data in time domain were then converted in frequency domain with a 1 Hz resolution using the Fast Fourier Transform (FFT). The FRFs were calculated from the frequency domain signals as follow:     ( ) ( ).
Conclusion In this paper, conventional cyclic tension-compression tests and dynamic tests were performed to characterize the LVE behaviour of an AM specimen. The experimental measurements from dynamic tests were back-analysed with a simple procedure to determine one value of the complex modulus at the first resonance frequency for each tested temperature. The complex modulus values determined from dynamic tests were then compared to the complex modulus data from cyclic tension-compression tests, assuming the same shift factors are valid for both tests. The norm of the complex modulus obtained from dynamic tests is slightly higher than the one determined with the conventional approach using cyclic tests. The differences observed between the two tests tend to increase with temperature but remain less than about 15% over the whole reduced frequency range.
Second, each process receives only for the current or a higher timestamp, e. g. , received messages are stored, e. g. , in line 30, only if they carry the current or a future (ballot, label) pair; cf. line 5. We introduce a tag annotation in which the programmer can provide us with the variables and parts of the messages that are supposed to encode abstract time and timestamps. Using these tags, the protocol is annotated with verification conditions stating that (1) abstract time is monotonically increasing (line 24), which use the predicate tag_leq in Fig. 10  (page 22), (2) each process sends only for the current timestamp (line 26), and (3) each process receives messages from the current or a higher timestamp (line 32). Given an annotated asynchronous protocol, we check the validity of these assertions using the static verifier Verifast  [23].
Using Verifast was extremely useful to prove the conditions on the content of the mailbox, which is an unbounded list (lines 2 and 47). For example, the assert at line 47 states that all messages in the mailbox have their  ballot and label fields equal with the local variable ballot and label. The other predicates are given Fig. 10. If these checks are successful, the existence of a canonical form is guaranteed by a new reduction theorem proven in Section 6. Our reduction uses ideas from  [17]  where the notion of communication closure was introduced in CSP. Our notion of communication closure is more permissive than the original form, as we allow to react to a message that is timestamped with higher value (b , ) than the current local abstract time (b, ), provided that the code immediately "jumps forward in time" to (b , ).
Protocols with nested loops. Let us consider a protocol P without reception loops. The rewriting algorithm proceeds bottom-up: it starts rewriting the most inner loop using the procedure above. For each outer loop it first replaces the nested loop with a call to the computed CompHO protocol, and then applies the same rewriting procedure. Since we considered passing by value procedure calls in the CompHO semantics, all local variables are input parameters. Inner loops appearing on different branches may belong to the same subprotocol; in other words these different loops exchange messages. If tags associates different synchronization variables to different loops then the rewriting Here is where we used the strength of Verifast, to reason about dynamically allocated data structures. The size of the program annotated with the proofs for the asserts is given in LoC in the column "Annot.
The phase number is the view number (like in Paxos), that is, two processes either agree on the identity of a leader in a view or they know of no leader. Normal-Op is the sub-protocol used in ViewStamped to implement the broadcasting of new commands by a stable leader. Multi-Paxos is described in Sec 4. It is Paxos from  [27]  over sequences, without fast paths, where the classic path is repeated as long as the leader is stable. In Paxos parlance, the tags for leader election (outer protocol) are Phase1a, Phase1b, Phase1aStart (in this order). The rounds of the sub-protocol are called Phase2aClassic, Phase2bClassic, learn. We considers that acceptors and leaders play also the role of learners. Our tool has rewritten the protocols from Fig. 11. The implementation uses pycparser  [3] , a parser for the C language written in pure Python, to obtain the abstract syntax tree of the input protocol.
Cancer Incidence and Vital Status Follow-Up Cancer incidence was determined through record linkages with regional cancer registries (Denmark/Italy/the Netherlands/Norway/Spain/Sweden/United Kingdom; complete up to December 2006) or via a combination of methods, including the use of health insurance records, contacts with cancer and pathology registries, and active follow-up through study subjects and their next-of-kin (France/Germany/Naples/Greece; complete up to June 2010). Vital status follow-up (98. 5% complete) was collected by record linkage with regional and/or national mortality registries in all countries except France, Germany, and Greece, where data are collected through an active follow-up. Censoring dates for complete follow-up were between June 2005 and June 2009 in Denmark, the Netherlands, Spain, the United Kingdom, Sweden, Norway, and Italy.
The custom GoldenGate assay was designed using the Illumina online Assay Design Tool in May 2012. SNP genotype dataset for CEU population (Utah residents with Northern and Western European ancestry; HapMap Data Rel 28 Phase II + III, August 10, on NCBI B36 assembly, dbSNP b126) were loaded in the Haploview program (Broad Institute, MIT and Harvard, Cambridge, MA, USA) and SNPs with minor allele frequencies (MAFs) greater than 5% and the r 2 linkage disequilibrium (LD) statistic of 0. 8 were selected as tagging SNPs (tagSNPs). Additionally, we searched published literature for previously reported functional and regulatory SNPs in the genes of interest and included them in genotyping irrespective of MAFs or r 2 with other SNPs. Genotyping was performed by the Genetics Laboratory at Imperial College London.
Further adjustment for body mass index (BMI; continuous), smoking status (never, former, current smokers, missing), physical activity (active, moderately active, moderately inactive, and inactive), alcohol intake (continuous), hormone therapy, and menopausal status did not substantially change the results, and thus these variables were not included in the final statistical model. Subgroup analyses were conducted by sex and tumor location (colon vs. rectum). To examine the associations between genes (a combination of SNPs) and genetic pathways (a combination of genes) and CRC risk, we used the Adaptive Rank Truncated Product (ARTP) method  [25]  as implemented in the first step (no interaction) of the R package PIGE (http://cran. r-project. org/web/packages/PIGE/index. html). This method can combine associations of SNPs in each gene (or from the genes in a pathway) to provide a P-value at the gene or pathway level, respectively.
Genetic markers in high LD (r 2 ≥ 0. 8) were excluded using the AdaJoint R package (https://cran. r-project. org/web/packages/ARTP2). To investigate the multiplicative interaction between the genes and genetic pathways with 25(OH)D on CRC risk, we used the modified ARTP method as implemented in the R package PIGE. The P-values at the SNP and the gene levels were corrected for multiple testing for the number of SNPs and for the number of genes, respectively, using the false discovery rate (Benjamini-Hochberg or BH) method  [26]. Furthermore, we used traditional methods to assess potential interactions between SNPs and 25(OH)D stratifying by categories of 25(OH)D concentrations and assuming a log-additive model for genetic markers. Also, we assessed the association of 25(OH)D (per 24. 96 nmoL = 10 ng/mL) with CRC risk by genotype.
Baseline Characteristics of Cases and Controls Selected baseline characteristics of the CRC cases and matched controls are shown in Table  1. The mean age at blood donation of cases and controls was 58 years. On average, CRC cases had 4 years between blood donation and the time of diagnosis. The dataset included 520 rectal cancer cases and 900 colon cancer cases. a Percent missing is not shown. Therefore the total percentages do not add up to 100%. b Season standardized using the sine-curve method  [25].
The somatostatin (SST) family members are a group of neuropeptides that are best known for their role in the regulation of growth, development and metabolism. The occurrence of six paralogous SST genes named SST1, SST2, SST3, SST4, SST5 and SST6 has been reported in vertebrates. It has been proposed that SST1, SST2 and SST5 arose in 2R from a common ancestral gene. SST3 and SST6 would have been subsequently generated by tandem duplications of the SST1 and SST2 genes respectively, at the base of the actinopterygian lineage. SST4 is thought to have appeared more recently from SST1, in teleost-specific 3R. In order to gain more insights into the SST gene family in vertebrates, we sought to identify which paralogs of this family are present in cartilaginous fish. For this purpose, we first searched the recently available genome and transcriptome databases from the catshark Scyliorhinus canicula.
Identification of SSTd and SSTe sequences from cartilaginous fish S. canicula SSTd and SSTe genes were sought by TBLASTN (Altschul et al. 1990) using the C. milii and L. erinacea SST2-like sequences (AGCKLFFWKTFSHC and APCKLFFWKTFSHC) as queries against the S. canicula genome draft assembly (version 1, unpublished). S. canicula full-length SSTd cDNA (but not SSTe cDNA) was also retrieved from S. canicula Illumina transcriptome (version 1, unpublished). The SST repertoire of SST genes in R. typus and C. milii was sought by TBLASTN using the S. canicula SST cDNA sequences as queries against the R. typus and C. milii genome assembly databases  (Read et al. 2017; Venkatesh et al. 2014). Protein sequences were predicted by using consensus splice donor and acceptor site and by sequence identity comparison with related SST genes.
Synteny analysis To generate synteny maps, genes flanking C. milii and R. typus SST genes were obtained from the NCBI database. Non-protein coding genes and genes coding unknown proteins were excluded from further analysis. Note that SSTb was not included in the study since it is absent in both C. milii and R. typus, while in S. canicula, it is located on a very short scaffold that did not contain any other genes. Genes flanking the SST genes in eight additional osteichthyan species, namely human, chicken (G. gallus), green anole lizard (A. carolinensis), Western clawed frog (X. tropicalis), coelacanth (L. chalumnae), spotted gar (L. oculatus), zebrafish (Danio rerio) and medaka (Oryzias latipes), were obtained from Genomicus (version 92. 01)  (Muffato et al. 2010 ).
Phylogeny analysis A set of 147 vertebrate SST precursor sequences was primarily collected from Yun et al. (2015) and Tostivint et al. (2016) and supplemented by the chondrichthyan sequences characterized in the present study. All these sequences were aligned using the Muscle algorithm (Edgar 2004) then manually adjusted. The phylogeny of SST precursors was inferred by Maximum Likelihood using PhyML (Guindon & Gascuel 2003) with the LG model of substitution combined to the neighbor-joining interchange (NNI) method on Seaview browser (Gouy et al. 2010). The robustness of the tree was assessed by the approximate likelihood test (aLRT) SHlike branch support and by bootstrapping procedure from 500 data set replicates. The Genbank accession numbers for all sequences used in the analysis are listed in Figure S1.
2013). Surprisingly, none of these genes appeared as the apparent ortholog of the atypical SST-like gene found in C. milii and L. erinacea. Nevertheless, these findings were regarded as in good agreement with the model described above since they support the existence of these three SST genes in the gnathostome ancestor  (Liu et al. 2010 , Tostivint et al. 2013 , 2014). However, due to the lack of a fully sequenced and assembled S. canicula genome, this view could not be confirmed  (Quan et al. 2013). The first goal of the present study was to determine the complete repertoire of the SST gene family in cartilaginous fish. For this purpose, we searched the newly available S. canicula genome and transcriptome databases (unpublished data). We also took the opportunity to explore or reexplore the genome and transcriptome databases from two additional species, the whale shark, Rhincodon typus  (Read et al.