Background: Although many studies indicate a positive correlation between GHRL gene Leu72Met polymorphism and an increased susceptibility to type 2 diabetes mellitus (T2DM), inconsistencies between independent studies still remain.Objective: Considering the inconsistencies between them, we have performed the current meta-analysis study. The objective of this study is to better examine the correlation of the GHRL gene Leu72Met polymorphism and T2DM.Methods: The current meta-analysis, involving 8,194 participants from 11 independent studies, was performed. A fixed effect model was used to evaluate the pooled odds ratios (ORs) and the corresponding 95% confidence intervals (95% CIs).Results: A significant association was found between T2DM and GHRL gene Leu72Met polymorphism under recessive , and homozygous genetic models  in the whole population. The correlation was more distinct in our subgroup analysis of the Chinese population under recessive , dominant , additive , and homozygous genetic models .Conclusions: In short, GHRL gene Leu72Met polymorphism was significantly correlated with increased T2DM risk, particularly in the Chinese population. Individuals carrying the Met72 allele of GHRL Leu72Met gene polymorphism, particularly those of Chinese ancestry, may be more susceptible to developing T2DM disease. ATP binding cassette transporter 1 (ABCA 1) (-defensin 1 (5- lipoxygenase (ALOX5) (solute carrier family 30 member 8 (SLC30A8)  (interleukin-10 (IL-10) (paraoxonase 2 (PON2) gene (GHRL (Ghrelin) (Type 2 diabetes mellitus (T2DM) is characterized by hyperglycemia in the context of insulin resistance and impaired insulin secretion. Its long-term complications include abnormal protein and lipid metabolism and multisystem damage . T2DM isfensin 1 , 5- lipo (ALOX5) , solute , TCF7L2 , superoxN2) gene , and GHRGHRL (Ghrelin) is a promising susceptibility gene for T2DM. First discovered by Kojima in 1999, the GHRL gene encodes for a 28 amino acid brain-gut peptide hormone, ghrelin  (GHRL gene Leu72Met polymorphism was not found to be significantly associated with susceptibility to T2DM in the Korean population (Though many studies examining the association between the  to T2DM . In 2012 to T2DM . On the = 0.026) . In 2006GHRL Leu72Met gene polymorphism  fasting plasma glucose (FPG) is used in the diagnosis of T2DM as proposed by the American Diabetes Association (2005). The FPG level ≥7.0 mmol/L or the 2 h plasma glucose of oral glucose tolerance test ≥11.1 mmol/L. No genetic relationships are present between the participants of the study. (c) Adopted data is derived from case-control or cohort studies published in the official journals or postgraduate dissertations. (d) In the individual studies, the control group genotype member should follow the Hardy-Weinberg equilibrium (HWE).The selected studies had to conform to the following inclusion criteria: (a) assessment of the association between T2DM and The data present in the studies was extracted according to classic protocol. Several investigators performed the meta-analysis. Two investigators retrieved the studies in duplicate and a third served as the arbitrator to settle differences between the two investigators and helped form a consensus. The excluded researches included those that failed to satisfy inclusion criteria, were repeated in other publications present in the analysis, or supplied inadequate information. Same data presented in different papers, but derived from the same group of authors was used only once. Items, such as the name of the first author, publication date, research region, genotypes numbers, and a whole number of subjects were included in the extracted data.GHRL gene Leu72Met polymorphism and T2DM was analyzed using odds ratios (ORs) and their corresponding 95% confidence intervals (CIs). The heterogeneity among the studies was calculated by the chi-square-based Q-test (P < 0.05) (Z-test (P < 0.05).Four genetic models were used in the meta-analysis: recessive (MM vs. LL+LM), dominant (LM+MM vs. LL), homozygous (MM vs. LL), and additive (M vs. L) genetic models. The association of  < 0.05) . If hete < 0.05) . But in  < 0.05) . The pooP < 0.05) and the funnel plot, to estimate potential publication bias. Egger's linear regression test on the natural logarithm scale of the OR was used to evaluate the funnel plot symmetry (P < 0.05) (The Fisher's exact test was used to assess the HWE ( < 0.05) . The StaOf the 19 manuscripts produced by the search 11 satisfied the inclusion criteria. Among the eight discarded studies, three papers were repeated publications, two were reviews, and the remaining three were not related to the topic in question. None of the studies deviated from the HWE. The data listed in GHRL Leu72Met gene polymorphism and T2DM under recessive  and homozygous genetic models . No significant correlation was found under dominant  or additive genetic models . In our subgroup analysis of the non-Chinese population, we found no significant correlation was detected under recessive , dominant , homozygous , and additive genetic models , dominant , additive , and homozygous .In our analysis of the whole population, we found a significant correlation between the P = 0.563).We used the funnel plot and Egger's test to assess for publication bias among the individual studies. The symmetry of the funnel plot  and the GHRL gene Leu72Met polymorphism and T2DM under recessive (OR: 1.33) and homozygous genetic models (OR: 1.34). While correlation in the non-Chinese population was not significant under any of the genetic models (P > 0.05), correlation in the Chinese population was significant under all four genetic models—recessive (OR: 1.52), dominant (OR: 1.70), homozygous (OR: 1.54), and additive genetic models (OR: 1.16). Though the Met72 allele of GHRL Leu72Met gene polymorphism has no significant effect on the T2DM susceptibility in the non-Chinese population, data suggested that the mutation might be strongly associated with the increased T2DM risk within the Chinese people. The significant difference between the two populations might be attributable to the difference in region and ethnicity.In the current meta-analysis, we observed a significant correlation in the whole population between the Ghrelin has been shown to takes part in glucose metabolism which could promote insulin production in rats . GhrelinGHRL gene Leu72Met polymorphism does not alter the sequence of mature ghrelin, changes to mRNA stability or protein synthesis may contribute to T2DM development by affecting ghrelin secretion and activity. In 2008, Zavarella et al. observed a significant decreasing trend from Leu/Leu to Leu/Met and to Met/Met homozygous subjects in fasting insulin levels, triglycerides, and HOMA-IR index and an increase trend in ghrelin levels. Their observations suggested that the Met72 variant might play a protective role in modulating insulin resistance. They also speculated that an unknown Met72 functional variant in linkage disequilibrium could increase the ghrelin levels and probably improve insulin sensitivity , do not have a relevant impact in the secretion of total and acylated ghrelin (Ghrelin is present in the bloodstream in two major forms: desacyl and acylated ghrelin . Under pin vitro . In 2004 ghrelin .GHRL gene Leu72Met polymorphism with T2DM susceptibility was simply undetected by Liu's individual study. In 2012, Wang et al. also performed a similar meta-analysis of the Chinese population and came to a similar conclusion as the current meta-analysis (Liao et al. in 2013 reported a protective effect from T2DM in the Caucasians population, but no correlation between the genetic mutation and T2DM in the overall population . As onlyanalysis . BecauseGHRL gene Leu72Met polymorphism and T2DM. The ghrelin level was influenced not only by the GHRL gene Leu72Met polymorphism, but also by other gene polymorphism as Arg52Gln, and unhealthy dietary habits (There are inevitably some limitations in the current meta-analysis. The current meta-analysis still lacks large-scale studies on the correlation between the y habits . As the y habits .GHRL gene Leu72Met polymorphism was significantly associated with the increased T2DM risk, especially among the Chinese people. Individuals carrying the Met72 allele of GHRL Leu72Met gene polymorphism, particularly those of Chinese ancestry, may be more susceptible to developing T2DM disease. The above conclusion may open a novel window in guiding us to formulate a new T2DM therapy, especially in the Chinese T2DM population. Considering the limitations of this study, more large-scale researches are still necessary to clarify the conclusion in the long run.Finally, The authors report no relationships that could be construed as a conflict of interest.YL: conceived and designed the experiments, statistical analyses, and paper writing. YL, GG, HG, and HW: performed the experiments. YL, XY, and XL: analyzed the data. YL and HW: contributed reagents, material, and analysis tools. YL and HK: wrote the manuscript. YL and XL: reference collection and data management. YL, YZ, and ZY: study design.The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.


ICAM-1) rs5498 polymorphism and diabetic retinopathy (DR) risk in type 2 diabetic patients. We aimed to perform a systematic literature search and conduct random-effects meta-analysis to provide a quantitative evaluation.Genetic studies have reported contradictory results on the association between the intercellular adhesion molecule-1  and 95% confidence intervals (CI) were calculated using a random effects model.P = 0.987), the recessive model , the GG versus AA contrast , and the G allele versus A allele contrast . Subgroup analysis by ethnicity showed no association in Asian populations . Subgroup analysis by DR subtype also did not reveal any association of rs5498 with proliferative DR  and non-proliferative DR .Nine studies involving a total of 1792 cases and 1400 controls met our inclusion criteria. We did not find any significant association between rs5498 and DR risk at the dominant model  contains supplementary material, which is available to authorized users. Diabetic retinopathy (DR) is the single most common complication of diabetes mellitus and the leading cause of blindness in working-aged adults worldwide . DespiteIn recent years, emerging evidence has highlighted the potential role of intercellular adhesion molecule-1 (ICAM-1) in the development of DR. ICAM-1 is a immunoglobulin-(Ig)-like transmembrane glycoprotein expressed on the surface of leukocytes, endothelial cells, and epithelial cells . It inflICAM-1 is related to DR risk. More than 100 single-nucleotide polymorphisms (SNPs) were identified in the ICAM-1 gene. The best studied SNP is a G/A polymorphism in exon6 at codon 469 (rs5498), resulting in a lysine (Lys) to glutamine (Glu) substitution in Ig-like domain 5 that is essential for dimerisation, surface presentation and solubilisation of the protein [ICAM-1 rs5498 polymorphism.Because ICAM-1 has been implicated in DR development, multiple studies have investigated how genetic variation at 469 rs549, resulti469 rs549, resultiThe search strategy for this meta-analysis was comprehensive, aiming to retrieve the largest possible number of relevant studies. We systematically screened 5 electronic databases including Pubmed  published in English or Chinese; and 3) published prior to January 2018 unless an online version of the study had been released prior to this date; and 4) reported odds ratios (ORs) and 95% confidence intervals (95% CIs) or data to calculate them. Case-only and case series studies with no control population were excluded, as well as studies based only on phenotypic tests, reviews, meta-analysis. We also excluded unpublished studies or gray literature because we expected them to contain insufficient reporting for our analysis.One reviewer performed the initial screen of all papers identified by the electronic searches. Studies were excluded when the title clearly indicated that it did not meet the inclusion criteria. Where a title/abstract could not be rejected with certainty, the full text of the publications was obtained for assessment. Studies were considered eligible if they met the following criteria: 1) evaluated the frequency of the Data extraction was performed by the first author and entered into predesigned electronic tables. The second author checked the extracted data. Disagreements were resolved by discussion between the two authors. The following items were considered: first author, year of publication, location of the study, ethnicity, number of cases and controls, diagnostic criteria, allele or genotype frequency, Hardy-Weinberg equilibrium (HWE) status, and genotyping method. The methodological quality of each study was assessed by the Newcastle-Ottawa Scale, which was used for its simplicity in comparing observational studies. Studies were evaluated based on cohort selection, comparability and ascertainment of exposure using nine multiple-choice questions. Studies were deemed of low quality if the total score was 5 or lower .http://ihg.gsf.de/cgi-bin/hw/hwa1.pl). For our main analysis, we compared allele frequencies  between cases and controls. We also evaluated a dominant model (GG + AG versus AA) and a recessive model (GG versus AG + AA) for the G allele. All associations were presented as ORs with their corresponding 95% CIs. Heterogeneity was evaluated by Cochran’s Q and the I2 statistic. When heterogeneity was high , a pooled analysis was conducted using DerSimonian and Laird random effects models [Statistical analysis was performed using STATA 11 . To assess HWE status, we used a publicly available program (n = 5), India (n = 2), Japan (n = 1), and Slovenia (n = 1). Study characteristics and methodological quality of included studies are shown in Table Our search yielded 476 records, with 14 articles being possibly eligible after review on abstract level. After full-text review, we excluded 5 studies. Overall, 9 studies involving 1792 cases and 1400 controls met the selection criteria and could be used for meta-analysis –20. FiguICAM-1 rs5498 polymorphism varied from 16.7 to 54.8%. The pooled effect estimates among all studies did not find statistically significant associations between the ICAM-1 rs5498 polymorphism and retinopathy in type 2 diabetes at the dominant model , the recessive model , the GG versus AA contrast , and the G allele versus A allele contrast   (Table P = 0.014), the GG versus AA contrast  and the G allele versus A allele contrast   for the 2) Table . Among t2 ranged from 65.3–84.0%). The funnel plot did not demonstrate apparent asymmetry  with a glutamic acid (E). It is thought that the SNP affects mRNA splicing patterns that modify cell-cell interactions and influence inflammatory response [The response . Howeverresponse , 19. We ICAM-1 rs5498 polymorphism and retinopathy in type 2 diabetes. Previous meta-analyses on the same topic included fewer studies  [Combining published data from nine studies involving 3192 participants, this is the largest meta-analysis on the relationship between the , n = 7) –23. Resu, n = 7) , we incl, n = 7) , 23. Sunn = 8), it became evident from this meta-analysis that further studies should include larger non-Asian populations to evaluate race-specific effects of rs5498, such as African and Caucasian populations. The study by Petrovic et al. was the only one performed in Caucasians [ICAM-1 gene. We could not exclude the possibility that other ICAM-1 polymorphisms played a role in the susceptibility to type 2 DR. The study by Simões et al. identified a significant association between the ICAM-1 rs1801714 polymorphsim and non-proliferative DR in type 2 diabetes [ICAM-1 gene [ICAM-1polymorphisms was very small at the time of planning our meta-analysis and consequently we chose to focus our analysis only on the rs5498 polymorphism. It would be advisable that in the future attention should be paid to the relationship between other ICAM-1 polymorphisms and DR in type 2 diabetes.Since most of the included studies were conducted on Asian populations  levels with DR because of limited published data. There was some evidence suggesting that circulating sICAM levels were positively associated with DR prevalence in type 2 diabetes , 26. It ICAM-1 rs5498 polymorphism with DR in type 2 diabetes. Further investigation involving non-Asian populations is warranted on the association between this polymorphism and DR, particularly studies with larger sample size that adjust for confounding variables.In conclusion, Our meta-analysis has shown that there is no significant association of the Additional file 1:Search strategy used in Pubmed. (DOC 27 kb)Additional file 2:PRISMA Checklist. (DOC 57 kb)


A meta-analysis using data from seven German population-based cohorts was performed by the German Epidemiological consortium of Peripheral Arterial Disease (GEPArD) to investigate whether one question about claudication is more efficient for PAD screening than established questionnaires. Claudication was defined on the basis of the answer to one question asking for pain in the leg during normal walking. This simple question was compared with established questionnaires, including the Edinburgh questionnaire. The associations of claudication with continuous ABI values and decreased ABI were analyzed by linear and logistic regression analysis, respectively. The results of the studies were pooled in a random effect meta-analysis, which included data from 27,945 individuals .Meta-analysis revealed a significant negative association between claudication and ABI, which was stronger in men  than in women . Likewise, the presence of claudication symptoms was related to an increased odds of a decreased ABI in both men  and women .Asking only one question about claudication was able to identify many individuals with a high likelihood of a reduced ABI with markedly higher sensitivity and only slightly reduced specificity compared to more complex questionnaires. At least in men, this question should be established as first screening step. Peripheral arterial disease (PAD) is an underdiagnosed disease with high cardiovascular and cerebrovascular morbidity and mortality, 2. It aIn a study with 18500 French general practitioners, intermittent claudication was well known, but knowledge of recommendations for further diagnostic workup and treatment of PAD was poor.In order to diagnose most individuals with PAD, regular measurements of the ankle-brachial index (ABI) in the whole population starting at an age of about 40 years seem to be useful. Currently, general practitioners usually do not screen for PAD. ABI measurements need training, equipment and are time- consuming. No established simple and fast screening strategies were available.While PAD is sufficiently prevalent for screening, increases all-cause and cardiovascular mortality by 2–3 fold and an accurate test (ABI measurement) is available, adequately powered studies on morbidity and mortality reduction and cost-effectiveness of screening have been lacking. HoweverThe most important criterion for the definition of peripheral arterial disease (PAD) is an ankle-brachial index (ABI) below 0.90. Low ABI is associated with increased mortality and morbidity. TherebyClaudication as a major clinical symptom of PAD could also be identified as an independent predictor of increased mortality based on data from the population-based Study of Health in Pomerania (SHIP). Of noteLarge studies analyzing the association between claudication and ABI are missing. Therefore, we performed a meta-analysis within the German Epidemiological Peripheral Arterial Disease (GEPArD) consortium using cross-sectional data from seven population-based cohort studies to investigate the association between claudication and ABI in both sexes, using different definitions of claudication, ranging from one single question to the Edinburgh Questionnaire. Data from seven well-established cohort studies, which cover different regions of Germany, were analyzed.This analysis was performed, because PAD awareness would be significantly improved if many patients could be identified by a single question of a general practitioner. As individuals with severe PAD are more likely to actively seek physician’s advice, it would especially be of great help to identify those with mild PAD.This consortium was founded to jointly perform analyzes on PAD-related research questions using epidemiological data from different German regions. For the current analyses we included data from all recent German population-based cohorts with data on claudication and ABI, we are aware of. Cross-sectional data from seven population-based cohorts were available: Cardiovascular disease, living and ageing in Halle (CARLA), GutenbeThe above mentioned seven population-based cohorts included men and women who were randomly selected from regional population registries, if they followed the invitation to take part. For this analysis, all participants of these studies with valid data on claudication, ABI and covariates were included. Participants with prior percutaneous transluminal angioplasty or bypass surgery of peripheral arteries, who were already diagnosed with PAD, were excluded from this analysis.Ankle and brachial blood pressure values were obtained from the first examination of each participating study. In all studies, the highest ankle pressure of each leg was documented and then the leg with the lower ankle pressure was used for the calculations of this analysis. An ABI < 0.90 was defined as low ABI. Claudication was analyzed according to different questionnaires or a single question . The assInformation on covariates was derived from interviews or by standardized measurements. Smoking status was classified as never/ex-smokers or current smokers. Hypertension was defined by either self-reported use of antihypertensive medication or a systolic BP ≥140 mmHg and/or a diastolic value ≥ 90 mmHg . Diabetes mellitus was defined by self-reported physician’s diagnosis or use of antidiabetic medication (ATC code A10) or by HbA1c≥6.5%. Use of lipid modifying medication was assessed based on ATC coding (code C10). Waist circumference (WC) was assessed to the nearest 0.1 cm using an inelastic tape measure. Serum levels of low density lipoprotein (LDL) and high density lipoprotein (HDL) cholesterol were assessed by the methods described for each cohort previously.2 pointed towards heterogeneity of the estimates across the studies assuming that the true effect sizes are significantly different between the studies as indicated by the p-value for heterogeneity provided in the graphics. Meta-analysis was conducted using the metan command in STATA 13.1 . Based on data from the largest cohort, GHS, additionally sensitivity and specificity of each of the three definitions to detect low ABI was calculated for two different age-groups . The methods were carried out in accordance with the relevant guidelines and regulations.The association between claudication with continuous ABI values was analyzed in a linear regression analysis for each cohort separately. Likewise, the association between claudication with low ABI values was analyzed in a logistic regression analysis on cohort-level. The results of the population-based studies were then pooled in a meta-analysis. Study-specific effect sizes for the association between claudication and ABI were combined to a pooled effect size by random-effect meta-analysis. Random Overall data of 27,945 subjects  could be included into the respective analyses. Characteristics of the samples are shown in The proportion of participants complaining of claudication varies largely between the different cohorts with a minimum of 5.7% (HNR) and a maximum of 31% (CARLA) when definition 1   was usedThe unadjusted mean ABI reached from 1.04 to 1.19 among the participating studies, while the proportion of participants with low ABI varied between 1.2% and 6.2% .For claudication definition 1, the meta-analysis demonstrated a significant negative association between claudication and ABI, which was more pronounced in men. Specifically, men with claudication had a 0.07 units lower ABI than those without claudication  . This diFor claudication definitions 2 and 3, effect sizes and Odds Ratios increase with increasing complexity of the definitions, both in men and women. For definition 3, which is only available in both KORA studies and GHS, though, Odds Ratio for low ABI is 13.99  in men and 8.16  in women.Calculation of sensitivity and specificity of each of the three definitions to detect low ABI, based on data from the largest cohort, GHS, revealed a low sensitivity for all definitions which decreases markedly from definition 1 to 3. The specificity is very high (>90%) for all claudication definitions with a slightly lower value for definition 1 as compared to definitions 2 and 3 .Focusing on cardiovascular and cerebrovascular event prevention, we need to identify early PAD stages in order to be able to reduce morbidity and mortality rates in these individuals. Our analysis demonstrates the advantages of a simple first step- screening strategy for PAD which markedly improves sensitivity accompanied by minor declines in specificity in comparison to a more complex questionnaire. Adopting this single question about pain or cramps in the legs during normal walking in history taking of general practitioners would help to achieve the guideline requirement to implement and support medical and public awareness of PAD. This woThe narrow 95% confidence intervals overall and especially in the largest cohort (GHS)  demonstrThis single, simple question for claudication should serve as first screening step in all patients of a general practitioner who are at least 40 years old, as PAD prevalence already reaches 5% in women and men at age 45–49 years in high- income countries. If indiIn diabetics, additionally an oscillography, a duplex ultrasound of the lower extremity arteries and/or toe-brachial index (TBI) measurements should be performed, as ABI measurements may not be correct in individuals with mediasclerosis of the leg arteries. In this analysis, we did not account for elevated resting ABI in diabetics. As we focus on detection of PAD at an early stage, we decided not to exclude subjects with known diabetes mellitus, because primary care physicians and patients may not know about diabetes mellitus at first presentation. It is reassuring to see significant results even with the inclusion of diabetics.If PAD is diagnosed, patients should be treated according to current guidelines. This should include statin therapy and a plIf patients answer “no” to this claudication question, this does not exclude the presence of asymptomatic PAD. Further work-up for PAD should especially be considered in patients with diabetes, hypertension, hypercholesterolemia, positive family history of atherosclerotic disease, impaired renal function and smokers. It has been demonstrated, that even mild-to-moderate chronic kidney disease is associated with an increased risk of incident PAD. PatientClaudication- definition 1 is unspecific and may include “claudication” from non-ischemic causes. Claudication- definition 3 (Edinburgh Questionnaire) markedly underestimates PAD according to ABI <0.90—definition. While the Edinburgh Questionnaire was found to be highly sensitive in comparison to a subjective physician diagnosis of intermittent claudication, in this study it is not very sensitive in comparison to the generally accepted PAD-diagnosis based on ABI-measurement. There seems to be a relevant number of individuals, whose complaints do not meet the Edinburgh questionnaire criteria, but who are PAD-patients according to ABI-measurement.It is important to keep in mind that most PAD patients are asymptomatic. But if we manage to motivate general physicians to ask their patients one question about claudication and thereby identify 20–30% of patients with mild PAD , this wiPAD-patients suffer from a bad prognosis. Therefore, early identification of these patients, which enables modification of risk factors and early pharmacological treatment, seems important. Consequentially, the higher sensitivity of an unspecific question seems more important than the slightly higher specificity of a complex questionnaire. In our opinion, it is about time to leave the traditional questionnaires in favor of one simple question. The data of our meta-analysis support to include this very simple one-item questionnaire into the daily clinical practice and into angiological and other vascular guidelines. Future claudication questionnaires should be evaluated in comparison with ABI measurements, not in comparison with the clinical evaluation by a physician. The analysis of the GHS data is in accordance with the results of the meta-analysis supporting the application of one question on claudication when screening for PAD.One question about claudication (definition 1) can identify individuals with a high likelihood of low ABI. The weaker association in women when compared to men could point to a different–so far unrecognized- clinical presentation of PAD in women. Already in 1982, a study from Finland demonstrated that the validity of “typical” claudication symptoms was poorer in women than in men and they were also less reliable predictors of death in women. The RosTo our knowledge this is the largest analysis of the association between claudication and ABI in population-based studies. Our data demonstrate that even one simple question about claudication can help to identify individuals with PAD. However, these results do not negate the advantages of a PAD screening by ABI measurement. An ABI measurement allows to identify the large group of individuals with asymptomatic PAD, especiaSome limitations have to be taken into account. The Edinburgh questionnaire for evaluation of claudication was not used in all studies of the present analysis. The number of participants varied widely among the studies. Rates of claudication and low ABI increase markedly at higher age, age groups were not homogenous in the seven population-based studies. The available data were not sufficient for calculation of the association between claudication and an alternative ABI calculation using the lowest leg pressure. The perAt least in men, one claudication question [asking for pain or cramps in the leg(s) during normal walking] can identify many individuals with a high likelihood of PAD and should be established as first screening step.S1 Data(DOCX)Click here for additional data file.


PRKAA1 polymorphisms conduce to the development of cancer. PRKAA1 gene encodes the AMP-activated protein kinase summit-α1, and plays an important role in cell metabolism. Thus, we performed a systematic review and meta-analysis of all enrolled eligible case-control studies to obtain a precise correlation between PRKAA1 polymorphism and cancer susceptibility.Currently, several studies have demonstrated that PRKAA1 expression correlated with cancer tissues and survival time.Extensive retrieve was performed in Web of Science, Google Scholar, PubMed, EMbase, CNKI and Wanfang databases up to August 26, 2018. Odds ratios (ORs) and 95% CIs were performed to evaluate the overall strength of the associations in five models, as well as in subgroup analyses, stratified by ethnicity, cancer type or source of control. Q-test, Egger’s test and Begg’s funnel plot were applied to evaluate the heterogeneity and publication bias. In-silico analysis was performed to demonstrate the relationship of PRKAA1 rs13361707 polymorphism is not significantly associated with cancer susceptibility under all five genetic models. For rs10074991, we revealed a significant decrease risk in allelic comparison model , heterozygote comparison model , and dominant genetic model . Evidence from TCGA database and GTEx projects indicated that the expression of PRKAA1 in gastric cancer tissue is higher, compared to normal stomach tissue, as well as it in breast cancer and esophageal squamous cell carcinoma. However, the Kaplan-Meier estimate showed that there is no significant difference of OS and RFS between the low and high PRKAA1 TPM groups in gastric cancer, breast cancer, and esophageal carcinoma.Twenty-two case-control studies from 14 publications were enrolled, with 17,068 cases and 20,871 controls for rs13361707, and 2514 cases and 3193 controls for rs10074991. Overall, we identified that the PRKAA1 rs13361707 polymorphism is not participant with the increased risk of cancer, while the A allele of PRKAA1 rs10074991 revealed a significant decrease risk.To sum up, The online version of this article (10.1186/s12881-018-0704-8) contains supplementary material, which is available to authorized users. PRKAA1) of α-subunit is located at chromosome 5p12.1 [It is well known that hereditary materials and environmental aspects could influence the risks and take a critical part in the tumorigenesis of numerous cancers , 2, nevee 5p12.1 , 4. Numee 5p12.1 . Secondle 5p12.1 . Thirdlye 5p12.1 .PRKAA1 polymorphisms focused on its significantly positive association with gastric cancer (GC) [PRKAA1 polymorphisms affect susceptibility of cancer.Most of the publications concerned about cer (GC) –11, as wcer (GC) , but thecer (GC) –16. ThesComprehensive literature search on Web of Science, Google Scholar, PubMed, EMbase, CNKI and Wanfang databases was conducted to draw out all eligible case-control studies, and the latest search date is August 26, 2018. The following are valid keyword search strings: (PRKAA1) AND  AND . What’s more, we also manually retrieved the references of reviews or original research on this issue to identify additional studies. For these republished and overlapping studies, we enrolled the most recently published articles or case-control studies with a maximum number of subjects.PRKAA1 and cancer susceptibility; 2) case-control studies; 3) demonstrate the frequency of genotypes of all cases and controls, or could obtain it via calculating. However, studies would be excluded if they meet the items: 1) animal studies, meta-analysis, comments, reviews or case reports; 2) no efficient data of the genotype frequency; 3) repetitive publications; 4) the research contents were concerned about other disorders instead of cancers.All the eligible studies were enrolled following the details: 1) assess the correlation between the polymorphisms in Extraction of data were independently completed by two authors , and all disagreements were finally obtained a consensus after discussion. The following items from the eligible case-control studies were extracted: SNP code, first author, year of publication, ethnicity, cancer type, genotyping method, source of control and genotype frequency. For the control group of each study, we definition it as population-based or hospital-based from whether it is collected from physical examination or just no-cancer patients, and ethnicity was distinguished as “Caucasian” or “Asian”.PRKAA1 polymorphisms  and cancer susceptibility, the P value was adjusted by Bonferroni corrections, PAdjust = PZ * 5 models [P value of the Q-test was > 0.1, we selected the Mantel-Haenszel method (fixed model) to evaluate the pooled OR estimate; On the contrast, the Der Simonian and Laird method (random-effect model) was preferred to evaluated the P value < 0.1 group [https://wpcalc.com/en/equilibrium-hardy-weinberg/), the P-value > 0.05 suggested a HWE balance for the control group. In order to appraise the stability of the meta-analyses, we managed sensitivity analyses by excluding each case-control study and observe whether it influence the pooled ORs and 95% CI [P<0.05 was regarded as statistically significant.We performed the meta-analyses in the pool, ORs with corresponding 95% CI was recorded to evaluated the strength of the correlation between 5 models . When th.1 group , 20. In d 95% CI . What’s d 95% CI , 23. Allhttp://gepia.cancer-pku.cn/), was applied to draw out it [To further explore whether the expression of PRKAA1 affect tumorigenesis, we tried to search some evidence from the public TCGA database and GTEx projects, and a newly developed interactive website, GEPIA , and the stratified analysis also indicated no relationships in the subgroups of cancer type, ethnicity and source of control , heterozygote comparison model , and dominant genetic model  .Table Sensitivity analysis by delete each signal study were performed to assess whether it influence the overall ORs results. The sensitivity analysis results conducted by all five genetic models were displayed in Fig. http://gepia.cancer-pku.cn/), was applied to draw out it.To further explore whether the expression of PRKAA1 affect tumorigenesis, we tried to search some evidence from the public TCGA database and GTEx projects, and a newly developed interactive website, GEPIA , as well as in breast cancer  and esophageal squamous cell carcinoma   and disease free survival (RFS) of these four cancers. The Kaplan-Meier estimate showed that there is no significant difference of OS and RFS between the low and high PRKAA1 TPM groups in gastric cancer , breast cancer , esophageal carcinoma , and lung adenocarcinoma .Evidence from TCGA database and GTEx projects indicated that the expression level of 01) Fig. . In addiPRKAA1 gene affect the occurrence and development of cancer, one of them may be the activated AMPK and phosphorylated p53 induced cell cycle G1/S arrest, as well as influenced the cell cycle checkpoint [PRKAA1 allele, and the patients would suffer from several epithelium disease [PRKAA1 polymorphisms conduce to the development of gastric cancer, Sun et al. [PRKAA1 expression, Li et al. [AMPK is a highly conserved heterotrimeric Ser/Thr kinase, it could phosphorylates many pivotal downstream targets through the increase ratio of AMP/ATP in environment, affect metabolic pathways of cell growth, cell cycle and autophagy , 34. In eckpoint . On the eckpoint , as welleckpoint .The lack disease . What’s  disease . Currentn et al.  revealedi et al.  indicatePRKAA1, and the variant sites include rs154268, rs3805486, rs461404, rs6882903, rs13361707 and rs10074991. After our comprehensive search and analyses, only rs13361707 and rs10074991 have 3 or more studies, so the other studies are excluded. Among these publications, the result is not consistent. Li et al. [PRKAA1 doesn’t affect the process of gastric cancer. On the meanwhile, the first three study of Shi et al. indicated that T allele of rs13361707 shows a significantly increasing risk of GC, however, Li et al., Kim et al. and Cai et al. revealed that the mutant allele of rs13361707  caused enhanced risk of GC. In another case-control study conducted by Dai et al. [There are several studies concerned about the polymorphisms of i et al. , they rei et al.  only repi et al.  revealedPRKAA1, within the perfect LD (R2 = 1.00). In a book named “Personalized Management of Gastric Cancer”, Zhu et al. [PRKAA1 gene, and a remarkable relationships between rs13361707 and these three genes were shown in the results from GTEx, so the polymorphism of rs13361707 might influence the expression of PRKAA1. In the current study, we enrolled all eligible case-control studies, from different race to different cancer types, aim to draw a systemic result for readers, and stratified analysis was also performed to avoid the bias. For example, the allele frequency of rs10074991, rs13361707 for all the enrolled Asian population based studies are ranged from 40 to 60%, while the Caucasian population based studies are ranged about 80% , aim to avoiding the false positive results. Finally, the stability of results was assessed by sensitivity analysis, and potential publication bias was eliminating from the results of Egger’s test and Begg’s funnel plot.There are several advantages of this meta-analysis. Initially, we enrolled all eligible studies focused on the relationships between PRKAA1 polymorphisms and overall cancer risks to conducted a comprehensive meta-analysis. Furthermore, NOS method was used to assess the quality of each accepted case-control study, and exceed the low quality studies to make sure the reliability of pooled result. Additionally, eligible studies were stratified and calculated in different subgroups to reduce the impact of heterogeneity, including subgroup of ethnicity, cancer type and source of control. Another point, the However, there are also several disadvantages. To begin with, the untrustworthy result may be obtained, because of lack of subjects in several genetic polymorphisms. What’s more, potential enrolled bias may be existed, due to that only publications written in English or Chinese were assessed. Then, most of the enrolled studies are concerned about GC, only 2 about ESCC, 1 about breast cancer, and 1 about lung cancer, therefore, the meta-analysis result might not be able to illustrated the impact of rs10074991 and rs13361707 in overall cancer risk. Last but not least, the striking level of heterogeneity between the enrolled publications might influence the result, although we conducted the meta-analysis with Der Simonian and Laird method.PRKAA1 rs13361707 polymorphism is not participant with increased risk of cancer, while the A allele of PRKAA1 rs10074991 revealed a significant decrease risk, especially in Asian population. In the future., larger sample size studies based on numerous cancer types should be conducted to confirm the exploration of this issue.Our data have successfully elaborated that Additional file 1:Table S1. Total sample sizes and allelic percentages of enrolled studies. Table S2. Methodological quality of the included studies according to the Newcastle-Ottawa Scale. Table S3. PRISMA 2009 Checklist. Table S4. Details of the sensitivity analyses for the polymorphisms in PRKAA1 and cancer risk. Table S5.P values of the Egger’s test for the polymorphisms in PRKAA1. Figure S1. Begg’s funnel plot for publication bias for PRKAA1 rs13361707 polymorphism (B vs. A). For Begg’s funnel plot, the x-axis is log (OR), and the y-axis is natural logarithm of OR. The horizontal line in the figure represents the overall estimated log (OR). The two diagonal lines indicate the pseudo 95% confidence limits of the effect estimate. Figure S2. Begg’s funnel plot for publication bias for PRKAA1 rs10074991 polymorphism (B vs. A). For Begg’s funnel plot, the x-axis is log (OR), and the y-axis is natural logarithm of OR. The horizontal line in the figure represents the overall estimated log (OR). The two diagonal lines indicate the pseudo 95% confidence limits of the effect estimate. Figure S3. Sensitivity analysis of Asian based group OR Co-efficient for PRKAA1 rs13361707 polymorphism (C vs. T). Results were calculated by omitting each study in turn. The two ends of the dotted lines represent the 95% CI. (PDF 1760 kb)


There is extensive debate regarding the protective effect of 3-hydroxy-3-methylglutaryl-coenzyme A reductase inhibitors (statins) on colorectal cancer (CRC). We aimed to assess the association between CRC risk and exposure to statins using a large cohort with prescription data. We carried out a case-control study in Catalonia using the System for Development of Primary Care Research (SIDIAP) database that recorded patient diseases history and linked data on reimbursed medication. The study included 25 811 cases with an incident diagnosis of CRC between 2010 and 2015 and 129 117 frequency-matched controls. Subjects were classified as exposed to statins if they had ever been dispensed statins. Analysis considering mean daily defined dose, cumulative duration and type of statin were performed. Overall, 66 372 subjects (43%) were exposed to statins. There was no significant decrease of CRC risk associated to any statin exposure . Only in the stratified analysis by location a reduction of risk for rectal cancer was observed associated to statin exposure . This study does not support an overall protective effect of statins in CRC, but a protective association with rectal cancer merits further research. Focus on primary prevention and screening is necessary in order to reduce the incidence and mortality of this cancer. Although lifestyle risk factors have been identified in CRC3, randomized trials have failed to show a reduction of adenomas recurrence with diet6 or dietary supplements8. A large body of evidence has shown that nonsteroidal anti-inflammatory drugs (NSAID), particularly acetylsalicylic acid (ASA), reduce the risk of colorectal neoplasia10 but with possible adverse events11. Indeed, a safe and effective CRC chemoprevention agent in average-risk population would help reducing the incidence of colorectal neoplasia.Colorectal cancer (CRC) is the third most common cancer worldwide13, with controversial results16. In addition to their main effect on cholesterol synthesis, statins may cause a number of other pleiotropic effects that may influence tumorigenesis, such as antioxidant activity, effects on cell adhesion, or angiogenesis17. In vitro, statins have shown anti-proliferative and pro-apoptotic effects on human CRC cell lines, and also in tumour xenograft models19. Studies analysing the effects of exposure to statins on the prevention or prognosis of colorectal neoplasia have shown controversial results, which have been proposed to be due to heterogeneity amongst drugs, or to effects restricted to some subgroup of patients20.Statins, inhibitors of 3-hydroxy-3-methyl-glutaryl-coenzyme A reductase, are a widely used and well-tolerated class of drugs for the treatment of hypercholesterolemia. Previous studies indicate their possible role in cancer chemopreventionIn this observational study we have analysed a population-based health records database aiming to examine the association between statins, their subtypes and pattern of use, and CRC risk.www.sidiap.org)21, which comprises clinical information routinely collected by primary care professionals of the Catalan Institute of Health. This database includes information from 5.8 million people in Catalonia  that have ever contacted the public health system since 2005. The data retrieved included routine clinical data, such as diagnoses and health measurements, and was linked to information on dispensed prescriptions generated by pharmacies’ claims for reimbursement by the Catalan Health System. Drugs were coded according to the Anatomical Therapeutic Chemical (ATC) classification system22, and the date and quantity of the drug withdrawn from the pharmacy were recorded. Irreversible encoding of patient identifiers ensured anonymization of the information in the SIDIAP study database. The quality of SIDIAP data has been previously documented, and it has been used to study the epidemiology of health outcomes23.Subjects were selected from the Information System for Development of Primary Care Research (SIDIAP) database  and sex and living in the same region defined by the primary healthcare centre catchment area. For cases, the disease onset date, defined as the earliest CRC diagnosis date registered, was set as the index date. For controls, the index date of their matched case was applied. Information regarding comorbidities and drug use was truncated to that recorded prior to the index date for cases and controls.24. The age-specific incidence rates of colon and rectum cancer estimated in the Tarragona and Girona cancer registries for 2012 were downloaded from https://ci5.iarc.fr. The age-specific rates were averaged over registries and summed for colon and rectum, then multiplied by the total Catalan population for the same age groups and period downloaded from https://www.idescat.cat. The population was corrected multiplying by 0.8 to account the average SIDIAP cumulative coverage on the studied period. Supplementary Fig. To assess if the codes used for case identification were reliable and exhaustive, and also to obtain an indirect measure of the external validity of our sample, we estimated the expected number of incident CRC cases in the population covered by the database according to cancer registries in Catalonia that contribute to the International Agency for Research on Cancer (IARC) publication Cancer Incidence in Five Continents (CI5) Volume XI22. The average dose for each of these duration categories was established, dividing the sum of DDD by the interval length. Finally, to measure the effect of timing of exposure, we compared non-users to the subjects exposed exclusively 1 month to 5 years before the index date (short exposure) and the subjects exposed throughout the period of more than 5 years before the index date (long exposure). We chose ≥5 years as a cut-off point of long term, according to the mean of follow-up and statin use in randomized controlled trials.Patients were classified as exposed to statins if they had retrieved at least one dispensation with an Anatomical Therapeutic Chemical code beginning with “C10AA”; otherwise, they were classified as unexposed. We also obtained exposure data for nonsteroidal anti-inflammatory drugs (NSAIDs) including aspirin  as they potentially could confound the association between statin use and cancer risk, and other lipid lowering drugs . Daily defined doses (DDD) for each dispensed prescription were calculated by multiplying the container pills by dose (in mg) and dividing by the World Health Organization defined DDD (in mg) for each individual drug25.In order to compare the results with previous studies, statins were classified as lipophilic  or hydrophilic (pravastatin and rosuvastatin) and, by effectiveness in lowering LDL cholesterol levels, as low-potency  and high-potency (atorvastatin and rosuvastatin)10. Socioeconomic status was evaluated using the MEDEA socioeconomic deprivation score26, which was divided into quintiles for the analysis. Chronic comorbidity conditions considered for multivariable adjustment included those associated with CRC in the data: hypertension, hyperuricemia, diabetes, osteoarthritis and spondyloarthropathy, chronic lower respiratory diseases, extrapyramidal and movement disorders, episodic and paroxysmal disorders, mental and behavioural disorders, chronic kidney disease, heart failure, cerebrovascular disease, liver disease, insomnia, osteoporosis, peptic ulcer, inflammatory bowel disease. Conditions clearly defined as indications of statins (dyslipidaemia and cardiovascular disease) were not considered for adjustment, since they are in the explored causal path.The potential confounders identified a priori for this analysis were age, sex, socioeconomic status, region, year, body mass index (BMI), tobacco, alcohol, comorbidity conditions and NSAID useOdds ratios (OR) and 95% confidence intervals (95% CI) were calculated from unconditional logistic regression models. We compared the effect of no use to any use of drug and assessed effects of dose (DDD) and duration of statin use. We also explored the effect of the type of the statin (potency and lipophilicity). Subgroup analyses were performed according to sex, age groups, NSAID use and cancer location (colon or rectum). Missing data for body mass index (BMI) was imputed using the prediction of a linear model according to age, sex and, outcome status (63% had complete data). To avoid models with many parameters, an adjustment score was built from the predictions of a logistic regression model for CRC that included all potential confounding covariates, without selection strategies. This adjustment score was efficient to render all potential confounders non-significantly associated to CRC. All logistic regression models included the adjustment score. P-values were derived from likelihood ratio tests. Statistical analysis was carried out using R statistical software .Characteristics of the study population according to statin use are presented in Table The most frequent statin used was simvastatin  followed by atorvastatin   Table . The anaNo differences were observed when statins were classified by their potency or lipophilicity. The analysis of specific statins did not show differential effects regarding CRC risk. Finally, as Supplementary Table We performed a stratified analysis to determine whether CRC location influenced the effect of statins Table . A statiA significantly increased risk of CRC was observed for subjects exposed to bile acid sequestrants , but no significant association with CRC was seen for exposures to neither fibrates nor nicotinic acid, independently on location, statin potency or dose , there was no decrease in risk associated with statin use. There were no consistent associations observed for duration and cumulated dose of statin exposure and rectal cancer, while the protective association was similar for diverse statin types. Besides, exposure to acid bile sequestrants showed an increase of risk of 33%.13. However, data from clinical trials have not confirmed the protective effect seen in observational studies14. Recently, two meta-analyses and one systematic review including 40, 42, and 59 individual studies, respectively20, reported a modest reduction in risk of CRC among statin users. In contrast, previous studies that linked pharmacy and cancer registry databases found no associations between statin use and CRC risk34. Our study adds further evidence to the lack of a relevant effect of statins on the risk of incident CRC, supporting that any effect, if present, is of marginal magnitude. Our observation of a significant 11% increase of risk related to statin exposures shorter than 5 years is isolated, not found in other analysis such as the one exploring tumour location, and thus is of uncertain value.Previous case-control and cohort studies have suggested that statins could play a role in cancer chemopreventionet al.15 showed in a stratified analyses a significant decreased association of risk in rectal cancer and for lipophilic statins, but this was limited to observational studies, and not when data was obtained from clinical trials. Our study has also observed that statin use may be selectively associated with reduced risk of rectal cancer. The reasons for this disparity in site association are unclear. Though colon and rectum are very similar at the molecular level35, environmental factors37 such as tobacco and physical activity38 differ in their role in their carcinogenesis. Moreover, there are clear differences among these two cancer locations regarding anatomic, embryologic, and physiologic differences40.Liu 41, they are less used than statins due to their poor tolerability (only 0.7% of our population was exposed), so that generally they are only prescribed to patients intolerant to statins or with severe dyslipidaemia. The role of bile acids on colorectal carcinogenesis has been widely studied42, and one trial had already reported a potential increase of CRC for long term use of cholestyramine back in 199243, but we have found no other references analysing the effect of these drugs on CRC, which may merit further research.The analysis of the other lipid-lowering agents was to rule out confusion by indication, and it detected that bile acid sequestrants increased the risk of CRC, but all other lipid-lowering agents were not associated with CRC, similarly to statins. This finding was unexpected; while bile acid sequestrants are mainly used for the treatment of dyslipidaemia as they reduce low-density lipoprotein cholesterol44. Regarding cancer location, we found a high proportion (72%) of cases classified as “colon not specified” (C18.9), and a lower number of rectal cancers than expected (18% observed vs 48% expected45). However, the total number of cases combining colon and rectum was consistent with those expected based on the incidence data published by the Catalan cancer registries (Supplementary Fig. 45. While it is plausible that some rectal cancers might be coded as “colon cancer not specified”, we can assume that the specificity of the rectal cancer location should be high. Misclassification was probably independent of statin use, and although might reduce statistical power for the analysis of rectal cancer, any bias, if existent, should be towards the null hypothesis. Regarding exposure, we used actually dispensed prescriptions at pharmacies. Though there is no way to prove that dispensed prescriptions were actually consumed, data on dispensed drugs is more reliable than electronic prescriptions, which may overestimate exposure at the expense of prescriptions that are never dispensed at pharmacies. Another limitation was that we could not adjust for some risk factors for CRC like physical activity, diet or family history of CRC, because these data were not available or did not reach usable quality. Finally, we have observed in our population that a lower BMI was associated with CRC. This is a typical finding in case-control studies, because the BMI data is registered close to the cancer diagnosis, in order to objectivize weight loss caused by the CRC as a part of the diagnostic procedures, while controls may have BMI recorded more often when obesity is requiring a clinical intervention by primary care physician.Potential limitations of this study, common to others using routinely collected data, include the lack of individual validation of exposure or cancer status. Nevertheless, SIDIAP has been widely used for other epidemiologic studies, and previous validation studies have shown that the collected information is reliable regarding disease codingcified” C8.9, and The strengths of this study include the large sample size, and the high representativeness of the population, since SIDIAP includes data on roughly 80%% of the Catalan population. Because SIDIAP contains data collected in routine practice conditions, the likelihood of observer bias is minimized. The use of electronic medical records and invoicing databases allowed us to overcome memory bias. Despite the study period was limited to 2010–2015, individual medication data was available from 2005 onwards, which allowed us to study a long period of exposure, ensuring a minimum of 5 years before the CRC diagnosis. This is of paramount importance when studying diseases with long latency such as cancer, and, in fact short exposure time is a major criticism to statin randomized trials.In conclusion, this study adds further evidence about the lack of a relevant association between statin utilization and risk of incident CRC. While we found no association between the use of statins and overall colorectal cancer risk, the suggestive evidence of a decrease in risk for rectal cancer requires further research.Supplementary Figure and Tables


This study aims to assess the effect of Balint training (BT) in gastroenterology intern nurse practitioners (GINP) systematically.This study will search EMBASE, MEDLINE, PsycINFO, Web of Science, Cochrane Library, Cumulative Index to Nursing and Allied Health Literature, Allied and Complementary Medicine Database, and China National Knowledge Infrastructure from inception to the September 30, 2019 with no language limitation. In addition, we will also search grey records, such as conference abstracts and dissertations. Study quality will be checked using Cochran risk of bias tool. Statistical analysis will be performed using RevMan 5.3 software.This study will systematically evaluate the effect of BT in GINP and will provide evidence to judge whether BT is effective for GINP clinically.The results of this study may provide helpful evidence of BT in GINP in the clinical training. This study will be published at a peer-reviewed journal or a conference presentation.2.22.2.1This study will include randomized controlled trials (RCTs) on assessing the effect of BT in GINP. Any other non-RCTs will be excluded.2.2.2The participants in the experimental must undergo BT alone.However, the participants in the control group can receive any interventions, except BT.2.2.3Participants of GINP who receive BT will be included without any limitations of their gender, age, and economic status.2.2.4The primary outcomes are psychological disorders, including depression and anxiety, as measured by Hamilton Depression Scale or Hamilton Anxiety Scale, or any other relevant scales.The secondary outcomes are clinical knowledge or skills, as measured using mini-Clinical Evaluation Exercise (mini-CEX) or any instruments; and satisfaction, as measured by any relevant surveys or scores; and examination scores.2.3We will comprehensively search for studies published in the following bibliographic databases: MEDLINE, EMBASE, PsycINFO, Web of Science, Cochrane Library, Cumulative Index to Nursing and Allied Health Literature, Allied and Complementary Medicine Database, and China National Knowledge Infrastructure, which cover records in the domains of health cares, medicine, nursing, allied health, health education and its training. Study searches through those databases will cover publication from their inception to the September 30, 2019 with no language limitation. The detailed search strategy for MEDLINE is presented in Table 2.4Two authors will carry out study selection independently and respectively. First, titles and abstracts of all identified records will be scanned to remove any irrelevant and duplicated studies. Second, we will collect full-text of all remaining articles to further judge if they meet all inclusion criteria. All excluded studies will be recorded with specific reasons. We will demonstrate the whole process of study selection in the flowchart.2.5Two authors will independently extract information from the included studies. Any differences and discrepancies will be reconciled with the help of a third author through discussion. The extracted information include first author, country of study, research study, research setting, participant characteristics, sample size and methods, research methods, interventions, controls, outcome measurements, and any other associated information.2.6If any missing or unclear data will be identified, we will contact primary authors to inquire those data. If we cannot get reply, we will analyze available data and will discuss its potential effects.2.7Two authors will independently evaluate research quality. Any discrepancies between two authors will be solved by consensus through arbitration by a third author. We will use Cochrane risk of bias tool to assess the study quality for each included study. We will assess each included study through 7 aspects, and each item will be assessed as “high,” “unclear,” or “low” risk of bias.2.8I2 statistic to determine the proportion of variation in effect size across included trials due to the heterogeneity regarding I2 ≤ 50% as acceptable heterogeneity, and I2 > 50% as obvious heterogeneity. If I2 ≤ 50%, we will use a fixed-effect model, and will conduct meta-analysis. If I2 > 50%, we will utilize a random-effect model, and will carry out subgroup analysis to investigate its possible reasons. If there is still obvious heterogeneity after subgroup analysis, we will report outcome results as a narrative summary.We will use RevMan 5.3 software to perform statistical analysis. Continuous data will be calculated as standardized mean difference and 95% confidence intervals (CIs). Dichotomous data will be exerted as risk ratio and 95% CIs. We will calculate 2.9We will perform subgroup analysis according to the different interventions, comparators, and outcomes.2.10Sensitivity analysis will be conducted to check robustness of synthesized results by excluding studies with high risk of bias.2.11If there are at lest 10 studies, we will conduct Funnel plot and Egger's regression test to identify any possible reporting bias.3–20,23 However, there are still not consistent results regarding this issue. Thus, considering several previous studies on investigating the effect of BT for GINP through measuring outcomes, such as mini-CEX, this study will systematically and comprehensively assess the effect of BT for GINP. The results of this study are expected to provide helpful evidence of BT for GINP.Previous studies have reported BT as a required component of the clinical practice. They hypothesized that BT has been used for the management of GINP.Conceptualization: Wan-lu Liu, Chen Chen, Jing Li, Olga Moreno.Data curation: Wan-lu Liu, Zheng-ri Zhu, Chen Chen, Jing Li.Formal analysis: Wan-lu Liu, Zheng-ri Zhu, Chen Chen.Funding acquisition: Wan-lu Liu.Investigation: Jing Li.Methodology: Wan-lu Liu, Zheng-ri Zhu, Chen Chen, Olga Moreno.Project administration: Jing Li.Resources: Wan-lu Liu, Zheng-ri Zhu, Chen Chen.Software: Wan-lu Liu, Zheng-ri Zhu.Supervision: Jing Li.Validation: Wan-lu Liu, Zheng-ri Zhu, Chen Chen, Olga Moreno.Visualization: Wan-lu Liu, Chen Chen, Jing Li.Writing – original draft: Wan-lu Liu, Zheng-ri Zhu, Jing Li.Writing – review & editing: Wan-lu Liu, Zheng-ri Zhu, Chen Chen, Jing Li, Olga Moreno.


To explore the effect of selective serotonin re-uptake inhibitors (SSRIs) on risk of type II diabetes mellitus (T2DM) and acute pancreatitis (AP), expecting to provide guidance for clinic. Literature was retrieved by searching Pubmed, Embase, Cochrane and Scopus and hand searching of reference lists of related articles. Stata 14.0 was utilized for processing and analysis, and adjusted odds ratios (aORs) were applied. Our study included 113898 T2DM patients and 284131 controls from nine studies and 17548 AP patients and 108108 controls from four studies. The pooled aORs of SSRIs on the risk of T2DM and AP were 1.38  = 1.24–1.54) and 1.26 (95% CI = 1.13–1.40), respectively. Study design, quality, ethnicity, follow-up, and sample size of patients were the resources of heterogeneity. Subgroup analysis showed that 2 weeks is a high-risk time for AP after SSRIs use, with 1.48-fold-times as much after it. This meta-analysis provides evidence of a significant positive association between SSRIs use and risks of T2DM or AP, and duration of 2 weeks of SSRIs use has higher risk of AP, which should be paid much attention to. Type II diabetes mellitus T2DM), an adult-onset disorder and characterized by insulin resistance, is increasing amongst children, adolescents, and young adults in recent years , and itsDM, an adSelective serotonin re-uptake inhibitors (SSRIs), as a new second-generation antidepressant, has become a first-line medication for depression, account for their safer and better tolerance than other types of antidepressants . The useSystematic search by retrieving PubMed, Embase, Cochrane, and Scopus to obtain relevance articles through December 2017. The following keywords were used for searching: ‘serotonin re-uptake inhibitors’ or ‘SSRIs’ or ‘sertraline’ or ‘paroxetine’ or ‘fluoxetine’ or ‘fluvoxamine’ or ‘citalopram’ or ‘escitalopram’ and ‘diabetic mellitus’ and ‘acute pancreatitis’. Cited references of the retrieved articles and reviews were also checked. Reference lists were screened to expect or obtain new articles. This systematic review and meta-analysis are reported in accordance with the Preferred Items for Systematic Reviews and Meta-analysis (PRISMA) Statement . We inclThe extracted data consisted of the following items: the first author’s name, publication year, population (ethnicity), methods, study design, matching criteria, sex, age (years), total number of cases and controls, adjusted OR or RR or HR estimates, and the corresponding 95% confidence interval (CI) for SSRIs use and adjusted confounding variables. Quality of studies was evaluated by the Newcastle–Ottawa scale, as recommended by the Cochrane Non-Randomized Studies Methods , a total2I statistic, P<0.10 and 2I > 50% indicated evidence of heterogeneity [P-value of Egger’s test <0.05 was considered significant [The end points included AP and T2DM. Meta-analysis was performed to calculate pooled ORs with 95% CIs by using Stata 14.0. We assumed there was similarity amongst OR, RR, and HR, because the rates of AP and diabetes mellitus events were less than 20% . Heterogogeneity . Random-ogeneity  was usednificant .The systematic search of PubMed, Embase, Cochrane, and Scopus provided a total of 787 citations, including 265 papers about AP and 522 papers about diabetes mellitus. After adjusting for duplicates and screening initial titles and abstracts, 749 were excluded. Twenty-six studies were potentially relevant studies, of which fifteen trials were excluded according to the exclusion criteria. Finally, eleven studies involved 17548 AP patients and 113898 T2DM patients were pooled for meta-analysis , includiP=0.031, 2I = 52.8%). Funnel plot was shown in P=0.666) suggested that there was no publication bias. Sensitive analysis exhibited that our result was stable (OR = 1.34 (1.15–1.57)), without change significantly by excluding Khoza et al. [As shown in a et al.  and Pan a et al.  NHSII.P=0.471, 2I = 0.0%, P=0.337) suggested no publication bias. Excluding Ljung et al. [Four studies providing data about risk of AP with taking SSRIs, the multivariate-adjusted pooled OR was 1.26 (95% CI = 1.13–1.40) with no heterogeneity . In subgroup analysis by methodologic quality, we discovered there was a significant result  with no heterogeneity  in high-quality studies (≥7 score), but this difference did not appear in low-quality studies (<7 score)  with increased heterogeneity . Outcomes of ethnicity exhibited that SSRIs could increase the risk of T2DM in both European  and American , but not Asian . To explore further the resource of heterogeneity, subgroups analyses about follow-up and sample size of patients were also conducted, and all of the results suggested that pooled estimates were significant. And heterogeneity decreased obviously when follow-up <10 years and sample size of patients <2000. To explore the clinical relationship between SSRIs use and AP, we evaluated ethnicity difference and time-point of AP appearing after SSRIs use. The result indicated that SSRIs indeed could increase the risk of AP of European , but may not of Asians . What was more, shorter duration of SSRIs use was more strongly associated with AP  than longer duration .On assessment of the risk of T2DM with taking SSRIs, subgroup analysis  about stOur study included nine observational studies to explore the effect of SSRIs on the risk of T2DM, the pooled adjusted OR showed that the risk increased 1.38 times with trying to calibrate as many other impact factors as possible, such as depression, hypertension and obesity, and so on. In addition, four studies in our study have for the first time identified evidence linking use of SSRIs to AP . RemarkaA case–control study of Anderson is included in our meta-analysis. In this large observational study, which included more than 160000 patients with depression treated with antidepressants, the long-term use of antidepressants in moderate to high doses in patients with diabetes increased the risk of 84%. This risk includes both tricyclic antidepressants and SSRIs. An average of 3.2 years of continuous use of antidepressants increased the risk by 2.60 times of diabetes (95% CI = 1.37–4.94) . A 10-yeFollowing are the possible mechanisms of diabetes caused by SSRIs. Boura-Halfon et al. have shown that SSRIs are potential inducers of insulin resistance, and its role may be to act as a direct inhibitor of the insulin signaling cascade in β-cells . The mecP=0.031, 2I = 52.8%) of pooled estimates of risk of T2DM with SSRIs taking was explored by subgroup analysis. We discovered that heterogeneity appeared obviously decreased when only nest case–control studies or high-quality studies were included. In addition, the similar results also appeared when follow-up <10 years or sample size of patients <2000. These results suggested that we not only found the resources of heterogeneity, but also knew that sample size had no impact on final results.The resources of heterogeneity , but may not of Asians . But because of small studies included, we could not be sure that there exists ethnicity difference. As to time-point of AP appearing, we discovered that SSRIs use could increase the risk of AP strongly when ≤14 days , with 1.48-fold-times of 14 days to 1 year, which we should pay great attention to risk of AP in 2 weeks after taking SSRIs.n=1540) of patients with type II diabetes had elevated levels of serum lipase (1.2% of which were more than three times higher). Amylase levels increased in 11.8% (n=1094) of the patients (0.2% of them more than three times higher) [Unfortunately, there has been no specific study to confirm the exact relationship between the occurrence of AP and the application of SSRIs. However, we can speculate that there may be some relationships amongst SSRIs, DM, and AP. In a cohort study of 337067 patients with T2DM, Noel et al.  showed t higher) . TherefoOur study concluded that SSRIs may increase the risk of T2DM and AP indeed with giving much more specific explanations about the relationship between SSRIs and T2DM; and for the first time identified evidence linking use of SSRIs to AP. However, there were some limits existing that we could not avoid. First, despite aORs performed, we failed to explain the source of heterogeneity amongst subgroup analysis completely, because not all the adjusted variables were exactly the same for each study. Second, subgroup analysis about type of SSRIs was not conducted limited to small number of studies, in this case, we explored other possible variability. Third, we could not obtain sufficient data to carry out subgroup analysis, such as dose, duration. Fourth, we only included articles with English language, so relevant studies in non-English might be missed, which lead to publication bias in a certain degree. Finally, only observational studies were included in the analysis, a cause-and-effect relationship could not be established.In conclusion, SSRIs could increase the risk of T2DM and AP, approximately 1.38-fold and 1.26-fold, respectively. In addition, the risk of AP was higher in the duration of ≤14 days with SSRIs use. Therefore, we should pay more attention to those taking SSRIs to prevent T2DM and AP. Although not so many studies have been included, sample size is large enough to support our conclusion. In the future, we hope that some high-quality studies about types, specific duration, intervals, and dose of SSRIs which trigger T2DM and AP are needed.


The results from meta-analysis of GWAS showing a p-value ranging between 0.05 and 5 × 10−8 were noteworthy under both Bayesian approaches . Re-analysis of observational studies and GWAS by Bayesian approaches revealed the noteworthiness of all significant associations observed by GWAS, but noteworthiness could not be confirmed for all associations found in observational studies.The 1858T allele in the protein tyrosine phosphatase non-receptor type 22 (PTPN22) locus shows one of the strongest and most consistent genetic associations with autoimmune diseases. We synthesized all meta-analyses reporting a genetic association of the PTPN22 1858T C/T polymorphism with autoimmune diseases. This work examined their validity to discover false positive results under Bayesian methods. We conducted a PubMed search to identify relevant publications and extracted the respective results, published until 30 November 2018. In observational studies, the associations of 1858 C/T genetic variant were noteworthy for 12 autoimmune or autoimmunity-related diseases -associated vasculitis, vitiligo, Graves’ disease, myasthenia gravis, Addison’s disease, giant cell arteritis, and endometriosis). In contrast, we could not confirm the noteworthiness for eight diseases . From the meta-analysis of genome-wide association studies (GWAS) with a  The 1858T allele in the protein tyrosine phosphatase non-receptor type 22 (PTPN22) locus has a strong and consistent genetic association with autoimmune diseases. This phosphatase is expressed in hematopoietic cells and in immune cells with highest levels found in neutrophils and natural killer cells [Many single nucleotide polymorphisms (SNPs) have been identified in PTPN22, but only one non-synonymous SNP has been intensively studied in relation to autoimmune diseases. The SNP rs2476601 is a change of cytosine to thymidine at nucleotide 1858 (C1858T) which results in an amino acid change from arginine to tryptophan at codon 620 (R620W). This codon is located in the polyproline binding motif P1 . The amiThere is no consensus whether 1858 C/T polymorphism is a gain- or loss-of-function variant. The C1858T has been reported as a susceptibility locus associated with several autoimmune diseases. It was first reported in type 1 diabetes mellitus (T1DM) . NumerouIn recent years, the advent of new genotyping and other molecular biology technologies has provided a huge increase in the quantities of data available for analysis. In epidemiology, candidate-gene and genome-wide association identified a large number of genes associated with diseases. Therefore, a general focus has been laid on the way genetic associations are reported. Meta-analysis has been widely used as a powerful approach to identify true-positive associated genes, but several limitations can alter the results. In most cases, results between overlapping meta-analyses on the same topic are inconsistent because of several confounding factors such as inclusion and exclusion criteria and number of included studies. p-value of <5 × 10−8 was set to determine the statistical significance in genome-wide association studies (GWAS) or its meta-analysis, a significance level of observational studies on genetic epidemiology  has not been changed. Therefore, judging the true significance between the suggested gene variant and disease is very important and Bayesian approaches can detect the claimed associations are genuine , which would be the reason why noteworthiness is so important. Bayesian approaches allow researchers to consider a much broader class of conceptual and mathematical models and permit to work on complex analytical problems, irrespective of the size of data [In addition, since the prior probabilities of genetic associations are low, there is a possibility that the number of false-positive associations by chance may be high, which could lead to an increased likelihood of finding false-positive associations. In the present study, we used Bayesian approaches ,11,12 beIn this review, we have synthesized all available data reporting the association of the PTPN22 1858 C/T polymorphism with autoimmunity retrieved from meta-analyses. Both meta-analyses of observational studies and GWAS were included and Bayesian approaches have been employed to estimate the noteworthiness of the evidence. We aimed to provide an overview to interpret the reported significant findings and discuss the genetic association of autoimmunity with the PTPN22 1858 C/T polymorphism.Studies were included if they satisfied the following criteria: (1) estimated the risk conferred by PTPN22 in autoimmune diseases using meta-analyses which reported odds ratio (OR) and 95% confidence interval (CI), and (2) published in English.Articles were excluded if (1) they did not report PTPN22 1858C/T polymorphism or autoimmune or autoimmunity-related diseases, (2) did not perform meta-analysis and (3) the original study did not report upper and lower confidence intervals, because these are necessary to calculate false-positive report probability (FPRP) and Bayesian false discovery probability (BFDP) , ethnicity of participants, the number of cases and controls, heterogeneity described with I2 or p-value and publication bias assessed as Egger’s p-value [We classified 53 articles into two categories depending on the methods of the respective studies, either meta-analysis of observational studies or GWAS. Data were extracted from each article, including the type of autoimmune disease, clinical symptoms, genetic variant, genotype comparison, odds ratio (OR), 95% confidence interval (CI),  p-value .p-value was lower than 0.05, equally described as 95% CI excluding 1.0 in meta-analyses of observational studies  1 − β  [To verify the noteworthiness of the reported genetic association, we used the Bayesian approaches, FPRP and BFDP. In our study, noteworthiness was used as a term that a genetic association of PTPN22 is associated with a disease in a significant manner. FPRP, proposed by Wacholder et al., tests the probability of no true association between the polymorphism and the disease . FPRP caeworthy) .−3 and 10−6) and two values of OR  which were thought to be valid for a noteworthy finding were chosen. FPRP was calculated by an Excel spreadsheet reported by Wacholder et al. [We used two pre-specified values for prior probabilities  and a BFDP value less than 0.8 was noteworthy. Each comparison of the genetic associations was regarded as noteworthy when FPRP of <0.2 or BFDP of <0.8 or both were fulfilled. We adopted these cut-offs of FPRP and BFDP, because these were set by the original authors  [We also used two pre-specified values for prior probabilities (10ctively) .p-value was adequately significant (p < 5 × 10−8) or was situated on the boundary of statistical significance (0.05 < p < 5 × 10−8).All the calculations to derive FPRP and BFDP were performed with the Excel spreadsheet released by Wacholder et al. and Wakefield et al. ,12. The n = 13), systemic lupus erythematosus  and type 1 diabetes mellitus , followed by five on juvenile idiopathic arthritis (JIA), 5 on Crohn’s disease (CD), 4 on anti-neutrophil cytoplasmic antibody (ANCA)-associated vasculitis, 3 on vitiligo, 3 on systemic sclerosis (SSc), 3 on Graves’ disease (GD), 3 on myasthenia gravis (MG), 3 on Addison’s disease (AD), 2 on psoriasis, 1 study on Behcet’s disease (BD), 1 on endometriosis, 1 on autoimmune thyroid disease (AITD), 1 on inflammatory bowel disease (IBD), 1 on giant cell arteritis (GCA), 1 on alopecia areata (AA), 1 on Sjögren’s syndrome (SS), and 1 study on ankylosing spondylitis (AS). The genetic association of the PTPN22 1858 C/T variant was evaluated in a total of 20 autoimmune or autoimmunity-related diseases. Most studies focused on rheumatoid arthritis . In contrast, the results did not show noteworthiness for eight diseases .−3 and 10−6 with a statistical power to detect an OR of 1.2. In addition, 13 and 11 comparisons were verified to be noteworthy at a prior probability of 10−3 and 10−6 with a statistical power to detect an OR of 1.5. With the use of BFDP, 32 and 30 comparisons showed noteworthiness at a prior probability of 10−3 and 10−6, respectively. In total, 32 (82.1%) of the 39 comparisons had noteworthy findings by FPRP or BFDP  using the FPRP estimation, at a prior probability of 10 or BFDP .−3 and 10−6 with a statistical power to detect an OR of 1.2, respectively. Moreover, 7 and 4 were noteworthy at a prior probability of 10−3 and 10−6 with a statistical power to detect an OR of 1.5, respectively. In terms of BFDP estimation, 8 and 6 comparisons showed worthiness at a prior probability of 10−3 and 10−6, respectively. In total, 9 (60%) of the 15 comparisons had noteworthy findings by FPRP or BFDP  of alopecia areata  using FPRP estimation, at a prior probability of 10−6 with a statistical power to detect an OR of 1.2 and 1.5. By means of BFDP, the four comparisons had noteworthy findings (<0.8) at a prior probability of 10−3 and 10−6 , findings were noteworthy for ANCA-associated vasculitis, T1DM, GCA and JIA. The four comparisons were verified to be noteworthy using FPRP estimation, at a prior probability of 10−6 with a statistical power to detect an OR of 1.2 and 1.5. In terms of BFDP, results were noteworthy at a prior probability of 10−3  for the genetic associations between the gene variant and disease. In recent years, Bayesian methods have been increasingly used because of their extreme flexibility as a major advantage. Bayesian methods can provide researchers with gains in performance of statistical estimation by incorporating prior information. BFDP allows the calculation of the recently proposed FPRP but uses more information ,12. ThesIn observational studies, the associations of the 1858 C/T genetic variant were noteworthy for 12 autoimmune or autoimmunity-related diseases  and the positive rate of true noteworthiness was different among diseases. The role of the pleiotropic 1858C/T of PTPN22 may suggest common and shared immune functions in various autoimmune diseases. More functional studies in the future may identify specific effects of this polymorphism in each autoimmune disease.However, the results did not show noteworthiness for the remainder, highlighting the need for further investigations. The non-significant association may implicate that the functional effect of C1858T on the Lyp protein is not a major contributing factor to study these autoimmune diseases or that the pathogenic inflammatory responses are not influenced or regulated by this pathway. In addition, the non-synonymous C1858T genetic variant is the only significant SNP out of many within the PTPN22 region, but we suggest that some other synonymous variants in these pathologies should be studied more intensively to elucidate their function. Furthermore, genetic association studies should be replicated in different populations with larger sizes.p-value <5 × 10−8 showed noteworthiness for PsA, MG, RA and JIA. Noteworthiness showing an association of four autoimmune diseases, namely ANCA-associated vasculitis, T1DM, GCA and JIA, could be observed in the GWAS non-significant findings when a p-value ranged between 5 × 10−8 and 0.05. It can be assumed that GWAS produces a more solid evidence than observational studies with a lower amount of false positive results, because of its stringent threshold to determine a significance. However, other genetic variants had a significant p-value  and found noteworthy by our Bayesian approaches and thus it may be concluded that results from GWAS with a p-value < 5 × 10−8could be identically replicated in observational studies. Some comparisons that could not be calculated using FPRP  could be computed using BFDP and were found to be noteworthy in our final analysis.All GWAS findings verified noteworthiness for all included diseases. Four studies with a PTPN22 encodes a protein tyrosine phosphatase that inhibits antigen-receptor signaling in T cells and promotes pattern-recognition receptor-induced type I interferon production by myeloid cells. Zheng et al.  proposedp-value per disease. Therefore, we could not consider other factors such as statistical models , ethnicity, and type of genotype comparison . In addition, despite our efforts, (1) some relevant articles may not have been included based on our search criteria as we considered publications limited to the PubMed database only; (2) some observational studies have been used for more than one meta-analysis, which raises the statistical issue of type 1 error inflation; and (3) we have not considered papers dealing with various permutations and interactions with other genes. Moreover, very few observational studies and GWAS involved African, Middle Eastern, and Asian populations and a population stratification analysis based on the same ethnic and geographic population may require further investigation.Although our report is the most updated and most analytical summary of available evidences on 1858 C/T polymorphism in autoimmune diseases, the review has some limitations. First, it should be noted that the lack of association might reflect the limited power of the studies including only a small number of patients with rare diseases. Secondly, we only included a single result of meta-analysis with the lowest In addition, FPRP often generates much smaller posterior null estimates than BFPD, because FPRP is a lower bound on the posterior probability relevant to the observed estimates ,12. HoweWe attempted to synthesize all meta-analyses on genetic associations of the PTPN22 1858 C>T polymorphism with autoimmune diseases and investigated their validity to discover false positive results under Bayesian methods. To verify results obtained from genetic analyses, both approaches may have advantages, and we were able to confirm significance in almost all autoimmune diseases within this borderline significance range. Therefore, PTPN22 is further confirmed as a candidate gene for further studies. Its promiscuous association with multiple autoimmune diseases might indicate a common mechanism underlying the development of autoimmune disease. Such a finding would have huge implications in our current understanding of autoimmunity and may be of therapeutic benefit and aid in devising preventative strategies. In addition, further studies should be performed to elucidate how PTPN22 1858 C/T could influence the pathogenesis of each autoimmune disease.


INK4a expression has been shown to be correlated with cancer-specific survival (CSS) in penile squamous cell carcinoma (SCC). Therefore, we performed this meta-analysis to clarify the prognostic value of P16INK4a for penile SCC. Recently, P16 A systematic search was performed in PubMed, Embase, and Web of Science to identify all relevant articles up to May 25, 2017. Hazard ratios (HRs) and corresponding 95% confidence intervals (CIs) of included studies were pooled to estimate the prognostic value. Trial sequential analysis (TSA) was performed to assess the quantity and strength of survival evidence.INK4a positive expression was significantly associated with a better CSS of penile SCC . No heterogeneity or publication bias was noted among the included studies. Furthermore, TSA demonstrated that the findings were based on sufficient evidence. Five retrospective studies consisting of 323 cases were finally included. We found that P16INK4a positive expression is independently associated with improved CSS for patients with penile SCC. P16 Penile cancer is an uncommon male cancer, with a very low incidence in Western countries (under 1/100 000 men in the United States and 0.1-0.9/100 000 men in Europe). However, its prevalence in the developing countries of Africa, Asia, and Latin America is greater (up to 3.7/100 000 men), and the cancer-related mortality remains significant in these areas . The difINK4a has been reported in HPV-related tumors such as the head and neck SCC and cervical cancer [INK4a detection and its prognostic significance in penile SCC. P16INK4a, known as a tumor suppressor, is shown to induce cell cycle arrest and prevent cell division by cyclin-dependent kinase 4 (CDK4) and cyclin-dependent kinase 6 (CDK6) inhibition, so as to control disease progression by mediating G1 arrest [INK4a is involved in retinoblastoma (Rb) inactivation by the viral E7 protein [INK4a expression in different populations with penile SCC, and P16INK4a expression has been shown to be correlated with an improved cancer-specific survival (CSS) in penile SCC [INK4a might be a prognostic marker for patients with penile SCC, which has not been conducted for a meta-analysis to date. Therefore, we searched relevant researches and performed this quantitative meta-analysis to clarify the prognostic value of P16INK4a expression in penile cancer.In recent years, overexpression of P161 arrest . Besides protein . Recent nile SCC –12. ThesThis meta-analysis was conducted according to the guidelines of the PreferredReporting Items for Systematic Reviews and Meta-Analyses (PRISMA) .INK4a'), and ‘penile' and (‘cancer' or ‘carcinoma' or ‘tumor' or ‘neoplasm'), and . The reference lists and bibliographies of all the eligible studies were, respectively, checked for other eligible investigations.A systematic literature search was performed in the electronic databases PubMed, Embase, and Web of Science to identify all relevant articles up to May 25, 2017. We used the combinations of the following keywords:  P16INK4a expression has been detected by immunohistochemistry (IHC); (4) studies reported cancer-specific survival (CSS) with hazard ratios (HRs) and 95% confidence intervals (CIs). Only research studies were considered eligible for the meta-analysis. Letter, brief communication, case report, and review were excluded according to our criteria. If more than one article on the same series of study subjects had been published, the most recent study with comprehensive data was selected. We also searched for relevant publications in Chinese (http://www.cnki.net/) to comprehensively understand the role of P16INK4a in penile SCC. Moreover, we used the Cochrane Risk of Bias tool and graded each potential source of bias as a low, high, or unclear risk  they are English publications; 2) the association of P16 they are the assoINK4a and cut-off definition, (5) follow-up time, and (6) HRs along with their 95% CIs and P values. If HRs and 95% CIs were not directly reported in publications, survival data were extracted from Kaplan-Meier (K-M) curves by using previously described methods [All eligible studies were reviewed according to the inclusion criteria by two independent reviewers (JYZ and HCZ), and uncertain data were reassessed by NHS. The following elements were extracted from each literature: (1) the first author and publication year, (2) characteristics of the studied population, (3) the number of patients, (4) detection method of P16 methods , 15. DatINK4a in penile SCC. A pooled HR of <1.0 represented an improved prognosis, and an HR of >1.0 correlated with poorer prognosis. Heterogeneity among studies was identified by the Cochran Q test and was quantified by the Higgins I2 statistic. Quantification of heterogeneity has been assigned low, moderate, and high to I2 values of 25%, 50%, and 75%. Besides, Galbraith plot and sensitivity analysis for included studies were, respectively, implemented to identify the source of heterogeneity. The random-effects model was used when significant heterogeneity was observed (P<0.10 or I2> 50%); otherwise, the fixed-effects model was applied [In the present meta-analysis, HRs and corresponding 95% CIs of included studies were combined to calculate the prognostic value of P16α spending function [β-spending function. When the cumulative Z-curve crosses the trial sequential monitoring boundary or enters the futility area, a sufficient level of evidence may have been reached, and no further trials are needed. If the cumulative Z-curve does not cross any of the boundaries, and the required information size has not been reached, there is insufficient evidence to reach a conclusion. TSA was carried out by the statistical software, TSA version 0.9 beta .Meta-analyses may result in type I errors due to random error from the studies included in the meta-analysis which had a small sample size, publication bias, and low quality, and studies whose conclusions tended to be changed by later studies with a larger sample size . We did function . BesidesAs shown in INK4a positive and P16INK4a negative. The mean or median age of included patients ranged from 60.0 to 69.0 years. All of the studies performed the detection of P16INK4a expression on tissue samples for immunohistochemical staining. Among the five studies, three directly reported the HR values [The main characteristics of the included studies are summarized in R values , and HRsR values , 12. The2 = 0%, P=0.764) .We found no heterogeneity among the included studies IP=0.764) , which wP=0.764)  and 6. S4 , whichINK4a positive patients  was 217 (A random-effects model (DL) was applied in the TSA for the five enrolled studies. The heterogeneity-adjusted required information size to demonstrate or reject a 54.3 % relative risk reduction  of P16 was 217 . The cumBegg's test and Egger's test were, respectively, conducted to evaluate the publication bias. As expected, the funnel plot of Begg's test exhibited symmetricalness with a P value of 0.462, and the P value of Egger's test was 0.249 . TherefoINK4a at an early stage of tumorigenesis [INK4a induces cell cycle arrest and prevents cell division by inhibition of CDK4 and CDK6 and inhibition of CDK-mediated phosphorylation of the Rb gene [INK4a expression has been described in penile carcinoma [INK4a in penile SCC requires further investigation.As a cancer of male sex organ, penile cancer and its treatment can seriously impact sexuality and intimacy, body image, urinary function, mental health, and life quality . Historiigenesis . P16INK4 Rb gene . Positivarcinoma , and thuINK4a was an independent prognostic factor for CSS in penile SCC. In the TSA, the pooled sample size was 323, which has surpassed the required information size of 217. Besides, the cumulative Z-curve has already crossed the conventional boundary and the α-spending monitoring boundary, suggesting that our meta-analysis was of sufficient evidence. In addition, there existed several advanced points in our meta-analysis. First, we strictly followed the inclusion criteria for articles, so as to ensure the quality of enrolled literatures. Second, the pooled sample size of our study was larger than that of any individual study, making the effect estimation more precise. Third, there was no statistical heterogeneity among the included studies , and thus a fixed-effects model was utilized. Fourth, the pooled HR of P16INK4a for CSS was 0.45, which was statistically significant (P<0.001) and was strong enough. Empirically, a predictive factor is considered to be strong when the value of HR is less than 0.5 or more than 2.0 [Our meta-analysis found that P16INK4a and penile SCC. Secondly, different researchers of included studies applied different cut-off values due to the lack of uniform standard for P16INK4a expression. This might affect the effectiveness of P16INK4a as a predictive factor for prognosis of penile SCC. Thirdly, although the pooled outcome of the five included studies indicated that P16INK4a positive expression was significantly associated with a superior CSS in penile SCC (P<0.001), individual results of three studies among them showed no statistical significance. Therefore, the prognostic value of P16INK4a for Penile SCC remained to be further investigated for confirmation [However, several limitations remained to be further refined. First of all, there were five eligible articles included in quantitative estimation, which led to the insufficiency of studies for subgroup analysis. Besides, there was no individual study on Asians, which might cause potential selection bias that hinders the comprehensive investigation of the association between P16irmation , 12. Fouirmation . HoweverINK4a positive expression is independently associated with improved CSS for patients with penile SCC, and TSA has been conducted for the first time to assess the quantity and strength of current evidence. In addition, a number of studies, which focus on the relationship between P16INK4a and penile cancer prognosis in Asian population, are required for further investigation in the future.In conclusion, P16