The carcinogenesis of hepatocellular carcinoma is complicated, and genetic factor may have the role in the malignant transformation of liver cells. IL-10 gene polymorphisms have been investigated for their potential roles in hepatocellular carcinoma This study aimed to investigate the relationship between polymorphisms of IL-10 , and hepatocellular carcinoma by performing a meta-analysis with eligible individual studies.This study followed the PRISMA 2020 Checklist. Relevant studies were searched in health-related databases. The Newcastle-Ottawa Scale criteria were used to evaluate the studies quality. Pooled odds ratio (OR) and its 95% confidence interval (CI) were used to determine the strength of association between each polymorphism and hepatocellular carcinoma using five genetic models. Stratification was done by ethnic groups. Trial sequential analysis (TSA) was performed to determine the required information size.P values > 0.5). TSA plots indicated that the information size for firm evidence of effect was sufficient only for the analysis of IL-10 (-592 A/C), but not for the − 1082 A/G or -819 T/C.Fifteen case-control studies (n = 8182) were identified. Overall, the heterozygous model showed a marginal significant association only between IL-10 (-1082 A/G) and hepatocellular carcinoma risk . On stratification, IL-10 (-1082 A/G) was significantly associated with hepatocellular carcinoma risk in the non-Asian population under dominant , heterozygous  and allelic models . IL-10 (-819 T/C) was significantly associated with hepatocellular carcinoma risk only among non-Asians under the dominant , recessive . For IL-10 (-592 A/C) with 11 studies, there was no significant association with hepatocellular carcinoma in all five genetic models (Findings suggest that IL-10 (-1082 A/G and − 819 T/C) polymorphisms are associated with hepatocellular carcinoma in ethnic-specific manner. However, this evidence is not conclusive because the sample size was insufficient. IL-10 (-592 A/C) polymorphism was not associated with hepatocellular carcinoma albeit with sufficient information size. Future well-designed large case-control studies on IL-10 (-1082 A/G and − 819 T/C) with different ethnicities are recommended.The online version contains supplementary material available at 10.1186/s12885-023-11323-1. Hepatocellular carcinoma is the most common primary liver cancer, accounting 70–80% of total primary liver cancers . WorldwiDue to nonspecific symptoms at the early stages, the mortality of hepatocellular carcinoma is very high , and theAmong the genetic factors, Interleukin-10 (IL-10) is one of the anti-inflammatory cytokines  as well There is a surge of individual studies that investigated the roles of IL-10 in patients with hepatocellular carcinoma. These studies varied in racial decent of participants, sample sizes, and the quality of study design, and these could contribute to heterogenous findings. Meta-analysis is a method that combines results from data collected from all eligible studies. There are published meta-analyses that assessed IL-10 on hepatocellular carcinoma , 21. HowWe conducted the current study in adherence to the PRISMA 2020 checklist for reporting our meta-analysis , and Google scholar. Keywords and MeSH terms were used with Boolean operators: [“Interleukin-10” or “IL-10”] AND [“polymorphism” or “gene polymorphisms”] AND [“hepatocellular carcinoma” or “liver cancer”]. To capture any additional studies, we performed a snowball method of manual cross-referencing of the retrieved studies and relevant systematic reviews. The search was restricted to studies published in English until June 2022. Search strategies are provided in Additional File Human studies that assessed hepatocellular carcinoma were included, if they.i)assessed IL-10 gene polymorphisms, − 1082 A/G , -819 T/C , and/or -592 A/C (rs1800872);ii)conducted case-control or nested case-control studies, irrespective of the method of DNA analysis;iii)compared hepatocellular carcinoma patients with the controls ;iv)provided sufficient information to extract genotype frequencies in both cases and controls; and.v)measured the outcome with odds ratio (OR) along with 95% confidence intervals (CI) .We assessed whether the distribution of genotypes in the control group of the studies included was consistent with Hardy-Weinberg equilibrium (HWE) .Hepatocellular carcinoma in this analysis was as defined in the primary studies. When more than one publication used the same participants, a publication with more comprehensive information was considered.Studies that did not fit to the inclusion criteria  were not considered. Studies that assessed treatment response, disease severity, drug efficacy, and pre-clinical studies were excluded.The two investigators (TYM and HHA) independently extracted information from each study using a piloted data extraction sheet. Information collected were: first author, publication year, study country, number of cases/controls, age group, male%, polymorphism frequencies in cases/controls, the racial descent (Asian or non-Asian), and minor allele frequency (MAF) in the controls, and HWE status . Any discrepancy between the two investigators was resolved by discussion with the third investigator (CN).The two investigators (TYM and HHA) independently evaluated the methodological quality of eligible studies using the Newcastle–Ottawa scale (NOS). The NOS checklist covers three main domains  in eight items, and each item was awarded 1 or 2 stars in maximum for high quality, and a final score obtained was between 0 and 9 stars . We consp > 0.05) . We performed the FPRP test with the use of a pre-set FPRP < 0.2 and assigned prior probabilities of 0.25, 0.1, 0.01 or 0,001 to examine an OR of 1.5  of associated with the hepatocellular carcinoma risk. To evaluate whether an association is “noteworthy”, we used a FPRP cut-off value of 0.2. Statistical power and FPRP were computed by the Excel spreadsheet provided elsewhere [28].To investigate the stability of results, we performed a sensitivity analysis with leave-one-out meta-analysis by removal of one study at a time. We planned to assess the publication bias by visual inspection of funnel plots . This waTrial sequential analysis (TSA), an approach that adjusts for random error risk, was done for estimation of the required information size . For dicMeta-analysis was done with Stata 16 (StataCorp TX), while TSA was with TSA version 0.9 beta .Figure Table Frequency of individual SNP is provided in Additional File Based on the NOS criteria, the studies included achieved the scores between 5 and 9. Less than half of these studies 6/15, 40%) obtained the high scores  % obtaine, 45 (Add(-1082 A/G) showed a tendency toward significant relationship with hepatocellular carcinoma under heterozygous model , however other four genetic models did not  . However, for the Asian population, there was no significant association between the IL-10 (-1082 A/G) and hepatocellular carcinoma in heterozygous model   and allelic model (G vs. A)  with the non-Asian population. However, there was no significant association with hepatocellular carcinoma under the remaining two genetic models (recessive and homozygous models). For a subgroup of Asian population (5 studies), there was no significant association with hepatocellular carcinoma under any of the five genetic models  and hepatocellular carcinoma in protection using the heterozygous model , and heterozygous model  was associated with hepatocellular carcinoma susceptibility under dominant model  and recessive model , and homozygous model  (Table On a subgroup of the non-Asian population (two studies), There was significant association with hepatocellular carcinoma susceptibility under dominant model , the Asian groups as well as the non-Asan group showed no significant association with hepatocellular carcinoma under any five genetic models Table .Table (-819 T/C and − 592 A/C) under all five genetic models (data not shown). These implied the stability of the effect estimates.For robustness of the findings, we performed’ leave-one-out meta-analysis’ by sequentially removing each study. For the SNP − 1082 A/G (nine studies), the omission of any single study did not significantly change the direction of estimates Fig. . This wap = 0.051), homozygous (p = 0.059), heterozygous (p = 0.066), and allelic (p = 0.050) models. Due to less than ten studies included, we could not investigate publication bias for IL-10 (-1082 A/G) and IL-10 ( -819 T/C).For the SNP (-592 A/C) (11 studies), publication bias was detected in the dominant model ((-592 A/C), the cumulative z-curve cut crossed the futility boundary, indicating the information size was adequate to establish a firm conclusion  were not sufficient to provide firm evidence on the relationship with hepatocellular carcinoma. Therefore, further case-control studies investigating with IL-10 -1082 A/G and − 819 T/C are needed to provide a more robust conclusion.The candidate gene approach is increasingly attractive to distinguish susceptibility genes that may trigger the initiation and progression of various types of cancer. In the present meta-analysis, we assessed the role of three important SNPs in the IL-10 gene  in the risk of hepatocellular carcinoma across eight countries.The gene encoding IL-10 is located on chromosome 1q31-1q32, and has three common IL-10 promoter variants  that significantly affect the gene transcription and expression . CytokinOur findings suggest IL-10 pertinent to AA and GG genotypes in -1082 A/G would be a protective role in hepatocellular carcinoma. A published study reported that high levels of IL-10 reduced the risk of hepatocellular carcinoma by reducing hepatic inflammation, which in turn prevents the malignant transformation of liver cells . The sigIt has been established that hepatocellular carcinoma is a multifactorial process, involving genetic factors and other factors such as lifestyle and environment factors, for the carcinogenesis . TherefoFor IL-10 (-592 A/C), no significant association in both Asians and non-Asians might be explored that IL-10 − 592 A/C had neither the causal nor inhibitory effect in hepatocellular carcinoma development in these study population. Preclinical studies reported that − 592 A/C did not play a part in altering the level of production of IL-10 , 55.A review including ten studies , assesseThe difference in the number of included studies along with different total participants between the published reviews , 21, 56 There are several limitations that should be acknowledged. First, the sample sizes in the current study were small. For example, less than ten studies were recruited in the overall meta-analyses of -1082 A/G and − 819 T/C, and only two studies included for subgroup analysis of -819 T/C. Hence, there may be a type II error, regarding the small number of studies.p value, the prior probability that the association between the genetic variant and the disease is real, and the statistical power of the test . In this analysis, the FPRP approach did not support the associations as noteworthy of true associations at the 0.001 level. The FPRP approach is essentially Bayesian in that it formally integrates data from direct observation of study results with other information about the likelihood of a true association [28]. TSA plots also demonstrated the issue of inadequate samples in the two SNPs (-1082 A/G and − 819 T/C). In some studies, the genotypic distribution of controls did not conform to the HWE. Nevertheless, sensitivity analysis showed the stability of the estimates. Moreover, we included only published studies in English. Hence, there might be relevant studies in other languages or non-published studies, which could contribute to selection bias. Lastly, due to the complexity of the etiology of hepatocellular carcinoma, there might be a contextual interaction between polymorphisms of IL-10 gene and other genes together with environmental/lifestyle related factors, which were beyond the scope of our study [It is possible for specific environmental and lifestyle factors to alter those associations between gene polymorphisms and hepatocellular carcinoma risk . However test 28,7. In thiur study . Taken tNevertheless, the current meta-analysis has strengths. We could identify a greater number of case- control studies that could enhance the power of the current meta-analysis, and additional TSA plot. The earlier reviews did not address adequate information size using the TSA method. The present study performed TSA as the required information size is important to classify the effect estimates as ‘firm evidence of effect’ or ‘potentially spurious evidence of effect’ , 31.Based on limited data presented in this review, we still do not know whether IL10 gene polymorphisms (-1082 A/G and − 819 T/C) increases or reduces the risk of hepatocellular carcinoma. Our analysis highlights the need for additional studies on the role of the IL10 gene in the development of hepatocellular carcinoma, and the findings should be used to inform healthcare providers considering screening of genetic risk factors. Such research could aid in identifying patients who have considerably higher risks of disease progression and may guide the development of customized treatment plans for chronic HCV infection. Since IL10 can diminishes the antiviral response, a practical approach could be the indication of earlier HCV treatment for those carrying low expression of IL-10. This would halt the inflammation induced by HCV infection and potentially inhibit the development of hepatocellular carcinoma .The findings suggested that IL-10 -1082 A/G and − 819 T/C have some roles in associated risk of hepatocellular carcinoma in the non-Asians group. The information size for confirmatory evidence was adequate only in the IL-10 (-592 A/C) to reach a conclusion. Future well-designed case-control studies with adequate number of participants in multi-ethnic groups are recommended to substantiate the evidence on the relationship between these two polymorphisms (IL10- 1082 A/G and − 819 T/C) and hepatocellular carcinoma.Below is the link to the electronic supplementary material.Supplementary Material 1: PRISMA 2020 ChecklistSupplementary Material 2: Search strategies in databasesSupplementary Material 3: Generic formulaSupplementary Material 4: Excluded nine studiesSupplementary Material 5: Frequency of genetic polymorphismsSupplementary Material 6: Methodological quality of studies assessed via NOS criteriaSupplementary Material 7: Forest plot of ? 819 C/T (a) dominant model (b) heterozygous model


The aim of this study was to investigate both the presence and severity of collapse in anatomical regions defined by the VOTE score , during drug induced sleep endoscopy (DISE) in patients diagnosed with obstructive sleep apnea, based on the bi-spectral index (BIS) sedation level.In order to conduct a meta-analysis of articles examining the relationship between the VOTE score and BIS sedation level in determining the presence and severity of upper airway collapse during DISE, a literature review was performed.P < .001). In the subgroup analysis of publications with BIS sedation levels in the range of 60 to 65, the distribution in the epiglottis region was heterogeneous, and it was found to be statistically significant according to the random effect model (P < .001). The distribution in the tongue was homogeneous, and it was found to be statistically significant according to the fixed effect model (P < .001). When the publications in which the BIS sedation level is in the range of 65 to 75 are examined according to the areas of obstruction; the distribution in 4 anatomical regions was homogeneous and statistically significant according to the fixed effect model (P < .001).As a result of the search made in the specified databases, a total of 1864 articles were reached. Five articles included in this review that had sufficient statistical data to be included in the meta-analysis were found. A statistically significant correlation was found between the BIS sedation level and the areas of obstruction in the VOTE score. The strongest association is at the epiglottis level, followed by the velum, oropharynx, and tongue, respectively (CC: 0.639, CC: 0.53, CC: 0.49, and CC: 0.346, It was found that BIS sedation levels during DISE application in obstructive sleep apnea patients were associated with obstruction of the anatomical regions of the upper airway. The strongest association was found at the epiglottis level, followed by the velum, oropharynx, and tongue, respectively. It is helpful to monitor the sedation level with BIS in order to better define the collapsed areas during DISE application. However, more studies are needed to better understand the relationship between BIS sedation values and sleep stages. Obstructive sleep apnea (OSA) is a sleep-related respiratory disorder with a high mortality expectation, characterized by hypoxia, hypercarbia, increased sympathetic activity, and medical complications in which the upper airway is partially or completely obstructed during sleep. The most important reason for failure is the inadequate preoperative evaluation required to determine the area or areas of the upper airway obstruction segment. In order to decide on the patient selection and target segment that will benefit from the surgery, the sleep surgeon should know the exact area or areas of the upper airway obstruction segment.The most common surgical treatment method traditionally applied is uvulopalatopharyngoplasty, with an average success rate of 40% in all patient groups. DISE can be performed when considering positive airway pressure alternatives such as upper airway surgery, oral appliance therapy, positional therapy, or a combination of different treatment modalities. Compared to the assessments made in an awake patient, DISE can change surgical treatment techniques in approximately 50% of OSA patients.Drug-induced sleep endoscopy (DISE), which has been developed in the last 3 decades, provides information about the narrowing and obstruction patterns of the upper airway in pharmacologically induced sleep. However, the degree of this effect has not been clarified. It is necessary to measure the depth of sedation during DISE to simulate sleep closest to natural sleep and to understand muscle relaxation levels. Although there is no standard protocol for sedation application and monitoring methods in DISE, it is possible to monitor sedation with a bi-spectral index (BIS). BIS evaluates the electro-encephalography values by converting them into signals between 0 and 100 and gives information about the depth of sedation and the level of consciousness.Anesthetic and sedative agents are effective in the collapse of anatomical regions in OSA patients.,13The VOTE  classification, developed to rank the collapse patterns observed during DISE, is a collaborative comparison of studies based on the anatomical structures that contribute to upper airway obstruction segment, qualitatively evaluating the degree of UAA narrowing and providing a scientific evaluation of DISE in different centers. It is a scoring system that enables language formation.This study aims to systematically review the international literature and make a meta-analysis with available data to evaluate the relationship between the BIS sedation level during the DISE application and the VOTE score, which evaluates the obstructed areas of upper airway in patients diagnosed with OSA by polysomnography (PSG).https://www.prisma-statement.org). Since meta-analysis studies are studies based on data analysis, ethics committee approval is not required in our country. Ethics committee approval was therefore waived.Studies were screened using the Preferred Reporting Items for Systematic Reviews and Meta-analysis statement (PRISMA) guidelines , which is “Do BIS levels have any effect on VOTE score in patients over 18 years of age who were diagnosed with OSA by PSG?”All randomized controlled studies, observational (prospective or retrospective) studies evaluating the effects of BIS levels, indicating the level of sedation, on the VOTE score when applying DISE in patients over 18 years of age with PSG-confirmed OSA were included. While scanning, English, and Turkish were determined as publication languages. No publication year restrictions were imposed.Pediatric age, animal studies, anatomical, genetic, radiological, biochemical, or histopathological studies were excluded. At the same time, studies using a scoring system other than VOTE when evaluating DISE findings, studies in which the explanation of statistical data is insufficient, studies whose publication language is not English or Turkish, and studies that do not have sufficient information to perform the meta-analysis (without the primary endpoint of DISE), and other diseases and Studies focused on surgical procedures , systematic reviews, meta-analyses, and non-original studies  were excluded.www.isiwebofknowledge.com), Pubmed (MEDLINE), Clinical Trials , and gray literature databases. Sources of gray literature included Open Gray and Google Scholar (http://scholar.google.com). In addition to the “cited by” function of Google Scholar, the “related citations” function was used to search for additional studies. Conference papers and abstracts were manually scanned in Web of Science. EndNote software  was used for duplication control. The scanning strategy following the PRISMA guidelines is shown in detail in Figure Searches were done in the Cochrane Library, ISI Web of Knowledge ;-Grade 2: complete obstruction.The spreadsheet specified in Microsoft Excel was used for data extraction. The following data were collected for each included study. These data were the names of the authors, the year of publication, the aims of the study, the country, the number of patients reported, the mean age, gender information, mean apnea-hypopnea index (AHI), BIS values, and the National Institute for Health and Clinical Excellence evaluation score calculated for each study Table . The quality score was calculated for each study included in this meta-analysis and is shown in Table The National Institute for Health and Clinical Excellence quality assessment tool evaluated the quality of the studies included in this meta-analysis.Z transform using the Hedges-Olkin (1985) method to calculate the weighted summary Correlation coefficient under the fixed effects model. CC and 95% confidence interval (CI) values are given. CC value; If it is greater than 0.5, it was evaluated as a high association, 0.3 to 0.5 moderate association, 0.1 to 0.3 low association, if less than 0.1 little if any association. If the correlation coefficient is; 0.10 to 0.39 was accepted as a weak correlation, 0.40 to 0.69 moderate correlation, 0.70 to 0.89 strong correlation, and 0.90 to 1.00 as a very strong correlation.As the effect size, the contingency coefficient (CC) between the rating in the VOTE scoring system (grade 0–1–2) and the obstruction areas  was used. Meta-analysis of these correlation coefficients was performed using the Fisher To determine whether the studies included in the meta-analysis have publication bias, primarily the funnel plot was used, and then the Begg and Mazumdar rank correlation statistics were calculated. The Forest plot is shown with 95% CI for the visualization of the marker size according to the study difficulty.Q statistic, a chi-square heterogeneity test with (k − 1) degrees of freedom, was used to assess heterogeneity. Heterogeneity was evaluated using the I2 statistic. Depending on the degree of heterogeneity, significance was assessed using both the fixed-effect and random-effects models. An I2 value of 25% is considered low heterogeneity, 50% is considered moderate heterogeneity, and 75% is considered high heterogeneity. Regarding publication bias analysis, it was concluded that there was no publication bias as the P values of the articles were greater than .05.In meta-analysis studies, the choice between a fixed-effect model and a random-effects model is based on investigating the heterogeneity of the studies. For this purpose, the Cochrane As a result of the search made in the specified databases, a total of 1864 articles were reached. Among the articles reached as a result of the search, 185 articles were reached after the titles and abstracts of the 704 articles obtained after removing the duplicates were examined in accordance with the exclusion criteria. Then, from the articles whose full texts were examined in detail, 21 articles were found following our PICO question. After detailed analysis, 5 articles included in this review that had sufficient statistical data to be included in the meta-analysis were found. Scanning was performed according to PRISMA guidelines and outlined in accordance with the PRISMA flowchart . In contrast, Park et al the effect size for the velum region of their study was determined as CC: 0.213 and 95% CI low association.The studies of Wang et al, Park et al, and Bilal et al found a high association as CC>0.5 and 95% CI . Won et al the effect size for the oropharynx region of their study was determined as CC: 0.446 and 95% CI as a moderate association. Park et al the effect size for the oropharynx region of their study was determined as CC: 0.184 and 95% CI low association.Bilal et al found a high association in their study because the effect size was CC: 0.561 and 95% CI. Wang et al the effect size of their study was determined as CC: 0.378 and 95% CI moderate association. In contrast, Park et al, Park et al, and Won et al, the effect sizes for the tongue region of their study were determined as a low association .Park et al, Park et al, and Won et al studies found a high association as CC>0.5 and 95% CI . Wang et al and Bilal et al the effect sizes of their studies were found as a moderate association (CC: 0.452 and CC: 0.487 respectively and 95% CI).P values of the articles were found to be greater than .05 in the publication bias analysis, it was determined that there was no publication bias , oropharynx , and epiglottis  regions model is selected. However, the fixed-effect model was chosen because it was homogeneously distributed in the tongue root .When the 5 studies in the current study are analyzed according to their distribution; the random effect as it shows the heterogeneous distribution in the velum .In the presented meta-analysis, a statistically significant correlation was found between the BIS sedation level and the areas of obstruction in the VOTE score. The strongest association is at the epiglottis level, followed by the velum, oropharynx, and tongue root, respectively , oropharynx , and epiglottis  showed heterogeneous distribution and random effect model was applied. However, since there was a homogeneous distribution in the tongue root , the fixed effect model was applied .In the Oropharynx region were found to be 0.403 and 95% CI: −0.0442 to −0.715. Since the distribution in the oropharynx region is heterogeneous, it was not statistically significant according to the random effect model (P = .076).The correlation coefficient values of the articles examined in the study in the Tongue region were found to be 0.32 and 95% CI: 0.176 to 0.450. Since the distribution in the tongue root is homogeneous, P < .0001 was found to be statistically significant according to the fixed effect model.The correlation coefficient values of the articles examined in the study in the Epiglottis region were found to be 0.62 and 95% CI: 0.254 to 0.830. Since the distribution in the epiglottis region is heterogeneous, P < .0002 was found to be statistically significant according to the random effect model.The correlation coefficient values of the articles examined in the study in the P = .78, I2 = 0.00%), the oropharynx , the tongue root , and epiglottis  were found to be homogeneously distributed, and fixed-effect model was applied .In the presented meta-analysis, A statistically significant correlation was found between the BIS sedation level and the areas of obstruction in the VOTE score. The strongest association is at the epiglottis level, followed by the velum, oropharynx, and tongue root, respectively . The distribution in the epiglottis region was heterogeneous, and it was found to be statistically significant according to the random effect model (P < .001). The distribution in the tongue was homogeneous, and it was found to be statistically significant according to the fixed effect model (P < .001). However, while the obstruction at the base of the tongue and the sedation level between BIS: 60 to 65 were found to be a weak correlation (CC: 0.32), a moderate correlation was found with the obstruction at the epiglottis level (CC: 0.62).In the subgroup analysis of publications with BIS sedation levels in the range of 60 to 65, when the velum and oropharynx regions were examined, it was not found to be statistically significant according to the random effect model since the distribution was heterogeneous (P < .001). Although the relationship between the 4 anatomical regions that developed obstruction and BIS: 65 to 75 sedation level was determined as moderate, the highest numerical value was in the oropharynx (CC: 0.628) region. Oropharynx; It was followed by the velum, epiglottis, and root of the tongue .When the publications in which the BIS sedation level is in the range of 65 to 75 are examined according to the areas of obstruction; at this sedation level, the distribution in 4 anatomical regions was homogeneous and statistically significant according to the fixed effect model .Sedation applied during DISE application may change the dynamics of the upper airway during apnea attacks and cause false-positive upper airway collapse due to excessive muscle relaxation. These structures include the soft palate (velum), oropharyngeal lateral walls, and tongue base, each of which can individually contribute to the obstruction. The VOTE score determines both the presence and the degree of collapse in the identified anatomical regions.However, this meta-analysis was based on the VOTE Score and the individual contributions of anatomical regions were investigated. The VOTE classification, developed by Keziran et al based on approximately 8000 experiences, provides a framework for characterizing various structures that contribute to upper airway narrowing and/or obstruction. BIS values in the 53 to 70 range during DISE application were found to be the most affected obstruction area of palatal collapse. Ravesloot and de Vries, on the other hand, found the regions where the obstruction was observed the most, in order of frequency, as palatal collapse (83%), the base of the tongue (56%), epiglottis (38%), and oropharyngeal collapse (7%). Vroegop et al found palatal collapse (81%), the base of the tongue (46.6%), and oropharyngeal collapse (38.7%) rates. In another study on this subject, the rates of obstruction were found to be approximately 85% to 90% for the soft palatal region, 50% to 60% for the tonsil , and 30% for the tongue root and epiglottis, respectively, according to their anatomical areas. In a study, it was determined that the soft palatine region was the most common occlusion area and that half of the patients had collapsed at the base of the tongue. Although the rates of results differ between studies, during DISE application, velopharynx/retropalatal and tongue/retrolingual collapse is seen in most of the OSA patient group (80–90%), and more superficial sedation allows us to catch up with this patient group.We found that obstruction at the base of the tongue was significant at both sedation levels, but the rate of association increased when the sedation level decreased. In addition, we found that obstruction in the velum and oropharynx region was not statistically significant in the BIS: 60 to 65 sedation range but was significant at a superficial sedation level (BIS: 65–75). At the level of superficial sedation, all 3 sites were moderately associated. Therefore, this level of sedation may identify more obstruction sites. Similar to our research, Viana Ada et al, When examining the patient population in this study, it was found that they had similar AHI values and were categorized as having moderate to severe OSA (AHI > 15) and severe OSA (AHI > 30). Ravesloot and de Vries found a correlation between high AHI values and high collapsibility. However, given the limited number of studies on this topic, our research focused on the patient population diagnosed with OUA as defined by AASM. Additionally, there is a relationship between BMI and collapsibility. The patient population included in the presented meta-analysis had similar BMI values ranging from 25 to 30, classified as overweight according to the World Health Organization definition. Although it is known that the upper airway muscles relax mainly during the rapid eye movement period of sleep in OSA patients, the views are not clear as to what range the BIS sedation level is in the REM period. In the literature, studies show the stages of sleep with different results. Nieuwenhuijs et al classified sleep periods according to BIS levels as evoked conditions (98–94), stage 2 sleep (65–44), stage 3 sleep (49–35), and REM sleep (51–44). they detected. Sleigh et al determined light sleep (75–90) for slow-wave sleep (20–70) and REM period (75–92). Dahaba et al suggested that for stage 2 sleep (70–80), stage 3 (50–60), stage 4 (40–50), and REM sleep (60–70). Considering that the half-maximum effect of the collapsibility range for all obstruction patterns and anatomical levels varies between 65 and 80, the BIS: 65 to 75 range can catch the collapse of the anatomical regions included in the VOTE score, as we found in our study.In this meta-analysis study, we conducted a literature review based on the PICO question. The aim was to examine the relationship between the severity of collapse in anatomical regions, as determined by the VOTE score, and BIS sedation levels. The meta-analysis included patients diagnosed with OSA (AHI > 5) according to the American Academy of Sleep Medicine definition.,35 Propofol is a short-acting hypnotic/amnestic agent. Bolus use during DISE can potentially result in excessive muscle relaxation and respiratory depression.,36 In the presented analysis Wang et al (BIS: 60–65) and Bilal et al (65–75) in their study, none of the anatomical regions in the VOTE score were found to be associated with a low level of collapse. The common feature of both studies is that they are performed with propofol infusion without using a TCI device in sedation. This makes us think that the properties of drugs applied other than sedation levels may also be adequate for the degree of obstruction.In addition, another condition affecting the collapse patterns may be different doses of different anesthetics/sedatives, even if the BIS is in the same sedation range. Three studies included in the meta-analysis were conducted with midazolam infusion, and 2 studies were conducted with propofol infusion and without using a target-controlled infusion device. Midazolam is a short-acting benzodiazepine. Since midazolam has a muscle relaxation effect, it has been determined that it relaxes more pharyngeal muscles than propofol. found that the obstruction in the base of the tongue and larynx improved when turned from the supine position to the lateral position. They found that the obstructions seen in the lateral position are usually the obstruction originating from the oropharyngeal lateral wall. All obstruction patterns shown in our analysis are valid for patients in the supine position.Again, the common feature of DISE applications in the publications included in this study is that they are applied while the patients are in the supine position. Collapse patterns may vary in the lateral position. In their study, Lee et alThe present study has several limitations. First, although there are many studies in the literature for DISE application, there are not many studies examining the degree or location of upper airway narrowing according to the depth of sedation during DISE. Most studies do not meet the inclusion criteria because of the lack of a standard protocol for DISE, depth of anesthesia, patient position, differences in classification systems (such as two-level versus four-level), the definition of obstruction, and fundamental differences in the classification of outcomes. Therefore, the number of studies included in the presented meta-analysis was limited. Second, the relatively small sample size in the studies and the highly selected group make it difficult to generalize the results to all OSA patients. The third limitation of our study is that although the patient populations in the 5 studies had similar AHI values, they were not exactly the same, such as the moderate and severe groups. However, based on the authors’ knowledge of the literature, no studies were found that specifically examined patient groups with identical AHI values. Therefore, we believe that further research is needed in this regard. Finally, some of the results of the included studies were heterogeneous. In order to reduce this heterogeneity as much as possible, it was tried to compensate by choosing the proposed random-effects model.In our study, it was found that BIS sedation levels during DISE application in OSA patients were associated with obstruction of the anatomical regions of the upper airway. The strongest association was found at the epiglottis level, followed by the velum, oropharynx, and tongue root, respectively. According to the findings of the meta-analysis, during superficial sedation range (BIS: 65–75), the collapse of anatomical regions represented by the VOTE score can be identified. However, although an association was observed between collapsibility and BIS sedation levels during DISE application, we believe it is not possible to recommend a specific BIS level for DISE. This is because the relationship between BIS levels and sedation achieved through natural sleep or anesthetic drugs has not been fully elucidated. Therefore, further research is needed to better understand the correlation between BIS sedation values and sleep stages.The statistical consultancy of the presented meta-analysis was provided by MedicReS Good Biostatistical Consultancy Services. We would also like to thank Professor Arzu Kanik for her valuable contributions to the statistics department.Conceptualization: Özlem Öner, Mustafa Cenk Ecevit, Ali Necati Gökmen.Data curation: Özlem Öner, Mustafa Cenk Ecevit, Ali Necati Gökmen.Formal analysis: Özlem Öner, Mustafa Cenk Ecevit, Ali Necati Gökmen.Investigation: Özlem Öner.Methodology: Özlem Öner, Mustafa Cenk Ecevit, Ali Necati Gökmen.Project administration: Mustafa Cenk Ecevit.Supervision: Ali Necati Gökmen.Validation: Özlem Öner, Mustafa Cenk Ecevit.Writing – original draft: Özlem Öner, Mustafa Cenk Ecevit, Ali Necati Gökmen.Writing – review & editing: Özlem Öner, Mustafa Cenk Ecevit, Ali Necati Gökmen.


Emerging evidence suggests that there may be an association between a history of fractures and dementia risk, but the epidemiological findings are inconsistent. We, therefore, conducted a meta-analysis to systematically assess the risk of dementia among people with a history of fractures.We comprehensively searched four electronic databases  for relevant literature published from inception to 10 January 2023. Longitudinal observational studies that investigated the association between any type of fracture occurrence and the subsequent risk of dementia were included for qualitative and quantitative analysis. Risk estimates were pooled using fixed-effects or random-effects models according to the level of heterogeneity. The Newcastle-Ottawa scale was used to evaluate the risk of bias in the included studies.A total of seven population-based studies involving 3,658,108 participants  were eventually included. Pooled results showed a significant association between fracture and subsequent risk of dementia  in cohort studies. Patients with fractures at different sites showed a similar trend toward increased risk of subsequent dementia. No gender, age, region, duration of follow-up, study quality, or study design specificity were observed. Sensitivity analysis indicates that the current results are robust. No publication bias existed. The results were similar in the cohort study with the standardized incidence ratio (SIR) as the statistical measure  and in the case-control study . Of note, the causal relationship between fracture and dementia was not demonstrated in this meta-analysis.People with a history of fractures are at increased risk of developing dementia. Enhanced screening and preventive management of dementia in people with a history of fractures may be beneficial. Dementia is a syndrome characterized by progressive impairment in cognitive function, memory loss, and a decline in the ability to perform daily activities . ApproxiRecent studies suggest that dementia and fractures may be closely linked , 7. AlthIdentifying populations at high risk of dementia for early intervention is a critical strategy to address this global challenge. As the incidence of dementia is rising at an alarming rate and there is a large population base of fractures, any association between the two could have important implications for medical practice. Several observational studies have investigated the association between a history of fractures and dementia risk but yielded inconsistent evidence , 16. ForThe present study was reported following the guidelines of the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA 2020) . The proWe systematically searched the databases of PubMed, Web of Science, Embase, and the Cochrane Library for literature exploring the association between a history of fractures and the risk of dementia. The recent search date was 10 January 2023, and no filters were used. The search strategy included terms related to fracture and dementia, such as “fracture,” “dementia,” “Alzheimer disease.” The Medical Subject Headings terms and free-text phrases provided in each database were combined through Boolean operators. The specific search strategy for each database is provided in The records generated from the database search were imported into EndNote X9 for review. After removing duplicates, the titles and abstracts of all the records were first screened to exclude reports that were clearly irrelevant to the study topic. Potentially eligible reports were then read in full to identify them for final inclusion. This process was done independently by two investigators (LS and XX), with disagreements resolved by consensus among all authors.Inclusion criteria were as follows: (a) the study design was a longitudinal observational study exploring the association between any type of fracture and subsequent risk of dementia and (b) the study outcomes provided relative risk estimates for the sample with a history of fractures compared to controls, such as hazard ratio (HR), odd ratios (OR), risk ratio (RR), or provided sufficient data to calculate effect sizes. Exclusion criteria were as follows: (a) single-arm studies without a control group; (b) cross-sectional studies with a non-longitudinal design; (c) case reports; (d) study outcomes other than risk of dementia; (e) pre-clinical studies; and (f) articles that did not generate primary data, such as reviews, commentaries, and opinions.Information from eligible studies was extracted independently by two authors and cross-checked. The following data were extracted: first author, year of publication, study region, sample source, study design, sample size, mean age, method of identification of fracture and dementia, confounding factors considered, duration of follow-up, and effect estimates with corresponding 95% confidence intervals (CIs) for the risk of dementia in people with a history of fractures.The quality of the included studies was assessed using the Newcastle-Ottawa Scale (NOS) in three main areas: participant selection, comparability between study groups, and outcome assessment in cohort study/exposure ascertainment in case-control studies . A NOS sI2 statistic and Cochran's Q test  or random effects (DerSimonian-Laird) models according to the level of heterogeneity . We perfs Q test . If bothThe database search yielded a total of 10,845 records. After screening according to pre-defined eligibility criteria, seven studies were eventually identified for qualitative and/or quantitative analysis , 21–25. n = 3), Korea (n = 1), the USA (n = 1), the UK (n = 1), and Denmark (n = 1). The mean age of the participants was over 50 years, and the follow-up period ranged from 9 to 15 years. Six studies were cohort studies, five of which used HR as the statistical measure , and therefore the random-effects model was used. The pooled results showed a significant association between fracture and subsequent risk of dementia   .Moreover, both SIR and OR endpoints were reported in only one study , 25, witTo examine differences in the association between history of fractures and dementia risk in samples with different characteristics, we conducted a stratified analysis according to fracture site, age, gender, region, follow-up time, study design, and dementia subtype .For the HR endpoint, results showed that almost all sites of fracture were significantly associated with higher subsequent dementia risk, including skull , vertebrae , upper limb , and thigh/leg/ankle  fractures. For hip fractures, the direction was consistent, although not significant . The risk was elevated for both Alzheimer's disease  and vascular dementia . Stratified analysis based on follow-up time showed that after excluding cases of dementia that occurred during the first 5-year follow-up period, patients with a history of fractures also had a significantly higher risk of long-term dementia beyond 5 years . Tsai et al. found that even 9 years after the fracture, the risk of dementia remained higher compared to the non-fracture sample  . No gendTwo studies reported the SIR and OR of dementia risk in patients with a history of fractures, respectively , 25. A cSince there was only one study for the SIR and OR endpoints and no pooled analysis was performed, sensitivity analysis was not applicable. For the HR endpoint, sensitivity analysis was conducted by comparing the results of the random-effects and fixed-effects models. As shown in P-values of 0.812 and 1.000, respectively.No risk of publication bias was detected by Egger's and Begg's tests with Through quantitative and qualitative analysis of the current evidence, we found a 28% increased risk of dementia in patients with a history of fractures, with a similar increase observed at almost all sites of fracture. No gender, age, region, duration of follow-up, study quality, and study design specificity were observed. The results from case-control studies were similar.Current research suggests that people who experience fractures are at a higher risk of developing dementia in the future than the general population. Whether as a predictor or a risk factor, this highlights the need to focus on the preventive management of dementia after a fracture. Multidisciplinary, including social, psychological, and medical, measures such as post-operative rehabilitation, mental health support, regular exercise, and nutritional supplementation, may be worth considering . In addiThere are no mechanistic studies that suggest that fractures directly cause dementia. Some of the physical changes that occur after a fracture, such as inflammatory responses, complications from fracture surgery, and reduced quality of life after a fracture, may contribute to cognitive decline and dementia risk. There is a prolonged inflammatory response during fracture healing, and a range of inflammatory cytokines, including interleukin (IL)-2, IL-6, IL-8, tumor necrosis factor-alpha, and C-reactive protein, are elevated , 28; preTo the best of our knowledge, this is the first study to systematically analyze the risk of dementia in people with a history of fractures using a meta-analysis approach in both cohort and case-control studies separately. Although a causal relationship between fracture and dementia risk cannot be established based on the current study, this meta-analysis provides relatively robust evidence for an increased risk of dementia in people with a history of fractures. The level of quality of the included studies was generally high, and a meta-analysis of over three million participants can be considered a rigorous estimate.P < 0.001); Tsai et al. found that patients with fractures still had a higher risk of long-term dementia 9 years later than non-fractured individuals , suggesting that studies using the general population as controls may have overestimated the results due to frailty-related factors (Unadjusted confounders are another important bias that affects the reliability of observational studies. Both dementia and fractures occur most commonly in older patients, which may be subject to confounding by shared exposures. Frailty-related factors are evident as potential confounders; for example, obesity, micronutrient deficiency, low educational level, lack of social support, chronic disease, and lifestyle factors are all strongly associated with the development of both fractures and dementia , 41. Des factors .There are several limitations to the current meta-analysis that need to be noted. First, almost all studies identified cases by disease diagnosis codes, which inevitably leads to underdiagnosis, misdiagnosis, and delayed diagnosis. Second, data from most of the current studies were collected retrospectively and from health insurance administrative claims databases, which are less complete than prospective studies and have residual confounding factors, particularly lifestyle factors such as smoking, alcohol consumption, and physical activity. It was therefore impossible to establish a causal relationship. Third, the majority of participants were older than 50 years, and therefore the association between a history of fractures and dementia risk in the young population is unclear. Fourth, although we conducted a stratified analysis, the level of heterogeneity in most results of the meta-analysis remained high, which may reflect differences in medical practice and environmental factors across regions. Therefore, we used the random-effects model for effect estimation to provide more conservative results, which accounts for the disappearance of statistical significance in some subgroups. Fifth, in some studies exploring the association between specific fractures and dementia, the control group did not exclude patients with other fracture types. If the finding of a positive association between history of fractures and dementia risk suggested by the current meta-analysis holds, then these studies would underestimate the effect size.This meta-analysis based on population studies suggests that people with a history of fractures are at increased risk of developing dementia in the future. Fractures may be one of the risk factors or predictors of dementia, and combining genetics, lifestyle, and other environmental risk factors may improve dementia prevention and screening work.The original contributions presented in the study are included in the article/YLi and LS: conception and design. LS, YLia, and XX: acquisition of data. LS and XL: statistical analysis. LS, YLia, and XX: drafting of the manuscript. YLi: critical revision of the manuscript. All authors contributed to the article and approved the submitted version.


Dear Editor,HOTAIR rs1899663 G>T Polymorphism with Colorectal Cancer in the Turkish Population: A Case–Control Study” by Yalınbaş et al.1 published online in The Turkish Journal of Gastroenterology. The article aimed to evaluate the association between the HOTAIR gene rs1899663 G>T polymorphism and colorectal cancer (CRC) risk in a Turkish population. The authors stated the G allele had a protective role against CRC in this population but that it would be appropriate to conduct this research with a larger sample to confirm the finding. Undoubtedly, this study is relevant because it is the first to demonstrate an association between CRC and the HOTAIR rs1899663 polymorphism in a Caucasian population. However, we would like to make a few observations. We read with great interest the paper entitled “Association of 2 which could bias the results.First, the authors genotyped the patients using DNA extracted from tumor cells. However, genetic association studies should be performed on genomic DNA extracted from non-cancerous cells because cancer cells tend to develop genetic changes related to tumor evolution,χ2 = 0.2066, P = .65), but the CRC sample showed an HWE deviation . This can be explained by the overrepresentation of the T allele, and thus the TT genotype, in the patients.Second, the Hardy–Weinberg equilibrium (HWE) test was omitted. We performed this analysis based on the reported genotype frequencies. The control group was in HWE , P = .0005], but the GT genotype did not . Furthermore, this increased risk was confirmed under the dominant and recessive models .Third, while the conclusion is appropriate, the statistical analysis is not. The authors concluded that there was an association between the T allele and CRC susceptibility, but they overlooked the corresponding odds ratios. Instead, they reported a protective effect for the G allele in Table 2. We respectfully suggest an analysis that we believe is consistent with the conclusion and helps clarify the study by specifying the risk for T allele carriers under different inheritance models. Thus, according to the logistic regression test, the T allele showed a significant association with CRC .On the other hand, in this order of ideas, we propose another statistical approach to analyze the association between the polymorphism and the clinicopathological characteristics of the patients, which yields similar results but shows the odds ratios. Thus, in line with the authors’ findings, the only feature that showed a significant risk for the TT genotype was cancer recurrence under a recessive model [GG + GT vs. TT; OR = 9.3 (1.0-88.0), We believe that these observations do not discredit this study. Rather, they complement it.


C-peptide is considered a peptide with active function in the body, which can affect people’s health. However, the results of previous studies on the possible association of C-peptide with the risk of cardiometabolic disorders have not been fully understood. This systematic review and meta-analysis aimed to investigate the association between serum C-peptide level and the risk of cardiovascular disease (CVD) events.The various important databases, including PubMed, Scopus, and Web of Science, were searched comprehensively to November 2022 to identify the relevant studies. The HR(95% CI) or OR(95% CI) for observational studies were extracted and converted into log HR or log OR and their standard deviation(SD) was computed. A random-effects model with an inverse variance weighting method was conducted, to calculate the pooled effect size.2 = 34.7%; P-heterogeneity = 0.140). For cross-sectional studies, the pooled results indicated a positive association between serum C-peptide and the odds of CVD outcomes .Sixteen observational studies, including one case-control study, eight cohort studies, and seven cross-sectional studies were included in the current meta-analysis. The sample size ranged from 90 to 7030, with an age range from 12 to 85 years. During the follow-up time (ranging from 5 to 17 years), 4852 CVD events occurred. Based on cohort and case-control studies, the pooled results showed no significant association between serum C-peptide with CVD events risk (RR = 1.02;95%CI:0.91–1.15, IThe pooled results of the current study suggested that C-peptide level was not related to the risk of CVD events in cohort studies, however, the meta-analysis of cross-sectional studies showed a significant association between C-peptide and an increased risk of CVD events.The online version contains supplementary material available at 10.1186/s13098-023-01142-6. Cardiovascular diseases (CVDs) is a general term covering the conditions affecting the heart or vessels . CVDs arPrevious studies have shown inconsistent findings on the possible association of C-peptide level with the risk of chronic diseases such as cardiometabolic disorders. A recent meta-analysis reported that low serum C-peptide level were significantly associated with increased incidence of coronary heart disease (CHD) and cerebral infarction. Also, this study showed that serum C-peptide is inversely associated with blood lipid level and promotes lipid deposition . Chung eOn the other hand, a recent retrospective cohort study found that the relationship between serum C-peptide level and cardiovascular risk is different depending on the history of T2DM. Unlike the participants without previous T2DM, the serum C-peptide level had a “V” shaped relationship with cardiovascular biomarkers, the associations of C‑peptide with these biomarkers and events were not significant in the patients with previous T2DM .Considering the inconsistent results in the past studies on the association of C-peptide level and risk of cardiovascular disorders, the present meta-analysis was intended to be an effective step in the correct and clear understanding of the relationship between the serum C-peptide level and the risk of cardiovascular diseases by combining all the data and presenting a general result. The findings of this study can also be used as a basis for future research regarding the therapeutic use of C-peptide in cardiometabolic disorders.To detect the published papers until November 2022, a comprehensive and systematic search was conducted through PubMed, Scopus, and Web of Science databases without any language, time, and study design limitations. The systematic search was carried out using the following keywords and MeSH terms: “C-Peptide” or “proinsulin” AND “cardiovascular” or “heart disease” or “vascular” or “stroke” or “cardiac” or “myocardial” or “coronary artery”. To prevent missing any related publications, we manually searched the reference lists of all obtained studies. The present systematic and meta-analysis was performed according to the Preferred Items for Systematic Review and Meta-Analyses (PRISMA) guidelines.Two reviewers independently scanned the title and abstract of all publications. The studies that met the following criteria were considered eligible for inclusion: (1) the original article, (2) adolescent and adult populations (3) cohort, cross-sectional or case-control design, and (4) reported Hazard Ratio (HR), Odds Ratio (OR), and Relative Risk (RR) with 95% Confidence Intervals (CIs). Studies with the following features were excluded: (1) Review, meta-analysis, and randomized clinical trial studies (2) unpublished data, (3) studies without sufficient data, and (4) congress abstracts, patents, dissertations, letters, comments, and conference papers.The following information was collected from eligible studies: author’s name, publication year, location of the study, design, number of participants and CVD events, age and gender of subjects, tools used for serum C-peptide measurement, outcomes, reported HR or OR with 95% CI for CVD events, adjusted covariates, and the duration of follow-up. Two investigators independently extracted the information to increase the accuracy of data collection.The quantitative content evaluation of each article was conducted according to the STROBE statement checklist. Two rev2 quantity[This meta-analysis includes fourteen eligible papers which consist of sixteen studies . The HR (95% CI) for cohorts and the OR (95% CI) for cross-sectional and case-control studies were extracted and converted into log HR, then their standard deviation (SD) was computed. A random-effects model with an inverse variance weighting method was conducted, to calculate the pooled effect size. The I quantity and Coch quantity were useThe flow diagram of included studies is presented in Fig. The detailed characteristic of 16 studies is shown in Table 2 = 34.7%; P-heterogeneity = 0.140).Figure 2 = 83.6%; P-heterogeneity < 0.001).The association of serum C-peptide levels with the odds of CVDs in cross-sectional studies is presented in Fig. Table However, in cross-sectional studies that subgroup analyses conducted on the region  and participants’ diabetes status, a positive significant relationship in Asia , America  and Europe  were observed, also there is significant heterogeneity between region subgroups .In diabetes status subgroups, there is a positive significant association among the diabetic population  and the general population , also a significant heterogeneity between diabetes status subgroups  was detected. Regarding Table According to visual inspection of the funnel plot Fig. sup, A, BTo explore each study’s impact on the pooled effect size, we used a random-effects model. After removing each study, we find out any significant change in the pooled effect size of the association between serum C-peptide levels with CVD in the meta-analysis on the cohort (0.98–1.03) and cross-sectional (1.23–1.46) studies. Findings showed that cohort studies were not affected by each included study, however, pooled OR of the cross-sectional meta-analysis becomes non-significant by omitting each included study except for studies conducted by Winocour et al. and Chung et al.We have two studies that have not adjusted the effect of the confounder variable. To compute the pooled RR for adjusted studies, we conducted subgroup analysis and omitted Faglia et al. study and Winocour et al. study in the cohort and cross-sectional study, respectively. There is no significant change in RR obtained from subgroup analysis in the cohort  and cross-sectional  versus their overall pooled RR.The findings from this meta-analysis systematically reviewed fourteen studies, which examined the association between C-peptide level and CVD events. Our results revealed no significant association between serum C-peptide levels with the risk of CVD events in a case-control and eight cohort studies. However, a significant positive relation between serum C-peptide and the odds of CVDs was seen in cross-sectional studies. Also, the results of the subgroup analysis based on the region and diabetes status indicated no noticeable association between serum C-peptide and CVDs in studies with case-control and cohort design. Nevertheless, cross-sectional studies conducted in Asia, America, and Europe and those studies in the diabetic and general populations had a significant association between C-peptide level and CVD events.+/K+-ATPase enzymes by interacting with membrane receptor on a molecular level which might have a favorable impact on both diabetic nephropathy and neuropathy. Also, C-peptide enhances skeletal muscle and skin blood flow, augments nerve function, diminishes urinary albumin excretion, and decreases glomerular hyperfiltration, only in individuals who suffer from type 1 diabetes and lack of C-peptide, and not in those with healthy status [In line with our findings from pooled HR of cohort studies, a meta-analysis by Autier et al. revealed no evidence of an association between breast cancer risk and serum C-peptide values . Most ofy status . In conty status . It has y status . Since oy status , type 2 y status –26, and y status , reliancOur findings on the association between serum C-peptide and CVDs based on the cross-sectional studies are similar to some of the previous meta-analyses claiming that greater circulating C-peptide level could be a predictive marker for chronic diseases risk susceptibility, such as various cancer , 29. AnoIt should be noted that since some evidence believed that dietary patterns with higher amounts of whole grains, vegetables, fruits, fiber, and coffee intake might be associated with lower plasma C-peptide level, while higher levels of refined grains and animal products might be associated with higher C-peptide level , evaluatOne of the reasons that can explain the non-significant relationship between C-peptide and CVD events based on the pooled results of cohort studies is that, in most of these studies, serum C-peptide has been measured as the main exposure of the study only at one point (at the beginning of the study), while it is possible that the baseline level of C-peptide of the individuals at the beginning of cohort studies did not reflect its level in the participants during the long term follow-up, because the C-peptide level may have changed in the study population during follow-up. So, if serum C-peptide was measured at several points in time in these cohort studies, it more could help determine the serum C-peptide accurately in participants and determine its real relationship with the risk of CVD events in them. Also, considering that cross-sectional studies cannot show the real causality, the pooled results reported in these studies on the significant relationship between high serum C-peptide level and increased risk of CVD events may be somewhat unrealistic; considering that, in cross-sectional studies, the levels of exposure (serum C-peptide) and outcome (CVD events) for individuals have been determined at a specific time point, it is possible that, the serum C-peptide level of patients with CVD events has subsequently increased due to their inappropriate and unstable metabolic condition, such as insulin resistance, metabolic syndrome, obesity, etc., and therefore, the high serum C-peptide level maybe not a definite risk factor for CVD events in individuals.The current study has several noteworthy strengths. To our knowledge, our study is the first meta-analysis investigating this topic with large sample size. However, a few limitations in this study should be kept in mind. Although the pooled results of cross-sectional studies were found to be remarkably significant, it had weak sensitivity and by discarding any of the 5 articles that reported significant findings from the analysis, the OR value became non-significant. In addition, based on the publication bias assessment, if the two missing studies were included in the analyses, the cross-sectional results would be insignificant. Also, different methods were used to assess circulating C-peptide levels. Moreover, included studies evaluated various micro and macrovascular complications which results in high heterogeneity. It seems that more studies are needed to evaluate our hypothesis.The evidence presented in our study indicates no significant association between C-peptide and an increased risk of CVD events in cohort studies, while a meta-analysis of studies with the cross-sectional design suggested that the higher C-Peptide level is associated with the increased risk of CVD events. Additional investigations are needed to shed light on this issue.Below is the link to the electronic supplementary material.Supplementary Material 1


Dietary factors may affect the incidence of colorectal serrated polyps (SP). However, its effects on SP are unclear as epidemiological studies on this topic have showed inconsistent results. The present systematic review and meta-analysis sought to evaluate the effects of dietary factors on SPs.I2 statistics. The meta-analysis was conducted by using a random-effects model, and the pooled effects were expressed with odds ratios (OR) and 95% confidence intervals (95% CI). Probable sources of heterogeneity were identified through meta-regression. Subgroup analysis were based on lesion types, study designs, countries, and so on.Studies regarding the association between dietary factors and SPs were identified by searching PubMed, Cochrane library, Embase and Chinese Biomedical Literature database from inception until 27 February 2023. Search terms include serrated, hyperplastic, adenoma, polyps, colorectal, rectal, rectum and risk. Heterogeneity was assessed using 28 studies were ultimately eligible after scanning, and five dietary factors including vitamin D, calcium, folate, fiber and red or processed meat were excerpted. Higher intakes of vitamin D , calcium  and folate  were not significantly associated with SP. Fiber intake  was a protective factor against SPs. Red meat intake increased the risk of SPs by 30% for the highest versus lowest intakes . For different lesion types, higher folate intake was associated with a decreased risk of HPs , and higher vitamin D intake decreased the risk of SPs including SSA/P .Higher dietary fiber intake plays an effective role in preventing SP, while red meat intake is associated with an increased risk of SP. This evidence provides guidance for us to prevent SP from a dietary perspective.https://www.crd.york.ac.uk/prospero/display_record.php?, RecordID=340750. Colorectal cancer (CRC) is considered to be caused by the accumulation of various aberrant mutations in the epithelial cells lining the colorectal mucosa . In 2020in situ and invasive carcinoma   and comp22340750 The search terms were: (Serrated OR Hyperplastic) AND (Adenoma OR Polyps) AND  AND Risk . BesidesThe following inclusion characteristics based on “PICO(S)” criteria were agreed for screening papers: (1) Adults aged 18 years and over, undergoing endoscopic investigation of the colorectum. (2) Assessment of a dietary factor, for example, vitamin D, calcium, folate, fiber and red meat. (3) Comparisons of the risk of SPs between dietary factors exposure group and non-exposure (or lower exposure) group. (4) Risk of serrated colorectal polyps, encompassing HP, SSA/P and/or TSA. (5) Original studies in English or Chinese.The exclusion criteria were as follows: (1) Study populations of other co-morbidities, for example, Crohn's Disease, ulcerative colitis, Barrett's esophagus, acromegaly; (2) Studies did not provide sufficient data for a meta-analysis; (3) The control group comprised a population with serrated polyps. (4) Reviews, comments, letters, and animal studies.Two reviewers (ZX Zhu and XF Guan) used Endnote X9 software to screen titles and abstracts independently. All data were extracted by two researchers (NW Liu and XX Zhu) independently using a standardized form. If there was any disagreement between two reviewers, the report would be sent to a third researcher (XY Li) and fully discussed. Data extraction included: the name of the first author, published year, country, study design, age, cases, sample size, type of SPs, dietary factors and adjusted confounders. The Newcastle-Ottawa scale was used to assess the quality and bias risk of observational studies, and a study with 7 scores (or more) was defined as a high-quality study. The Cochrane collaboration tool were used to assess the quality of RCTs .Q-test and I2 statistic, and heterogeneity was considered to be significant if I2 > 50% and P < 0.05. The summary OR was calculated through the random-effects model, which takes into account the presence of heterogeneity in their calculations   . Subgrou= 0.018) . BesidesI2 = 52.7%; P = 0.025). In all subgroups, calcium intake was not associated with risk of SPs, and countries, study designs and adjustments did not influence the heterogeneity of included studies . Protective effects were detected in studies not adjusting smoking  or alcohol intake   .I2= 75.8%; P < 0.001). Meta-regression showed statistically significant differences in the heterogeneity of subgroups with different countries . This association was found in almost all subgroups   . Followi= 0.010) .This systematic review and meta-analysis is the latest and most comprehensive study investigating dietary factors and their influence on risk of serrated colorectal polyps. Meta-analyses revealed a statistically significant increased risk of SP associated with red or processed meat intake, and fiber intake was protective against SP. Higher intakes of vitamin D, calcium and folate were not significantly associated with SP risk.Together, vitamin D and calcium are involved in regulating inflammation, differentiation, apoptosis and carcinogenesis in various cells, and have been widely studied as protective factors against colorectal cancer . Lopez-CFolate is abundant in leafy green vegetables, legumes, and cereals. It has been a focus of CRC chemo-prevention research for several decades because of its role in DNA methylation, repair, and nucleotide synthesis . In a raThe current meta-analysis suggested fiber intake decreased the risk of SPs by 10% for the highest vs. lowest intakes. Dietary fiber affects colorectal cancer risk by increasing fecal volume, diluting fecal carcinogens in the colonic lumen, and shortening fecal transit time through the gut . Recent The WHO has classified red meat as a class 2A carcinogen. A recent expert panel report also supported the role of red and processed meats in CRC risk . ProcessThe study provided a comprehensive analysis of associations between the above dietary factors and SP risk, which remain controversial today. Although most previous studies reported that calcium and folate intake could protect people from colorectal adenomas and cancer, the current meta-analyses found they had no such protective effect on SP.This work was subjected to some shortcomings. First, as most studies were conducted based on U.S. populations, it is not appropriate to generalize our findings to African and Asian populations. Second, the results of folate and fiber intake may not be very stable according to sensitivity analyses. Besides, the included studies varied in many ways such as study design, adjustment for confounders, and different ascertainment methods for factor exposure, which might have an impact on the results. To minimize the effect of this limitation, we adopted random-effect models to pool the estimates. Subgroup analyses and meta-regression were also conducted to detect probable sources of heterogeneity, which enabled us to account for these differences.The results of this meta-analysis indicate that higher dietary calcium, folate and fiber intake can reduce the risk of SP, and vitamin D intake may have the effect of preventing SPs, which needs to be determined by more evidence. Red meat intake is associated with an increased risk of SP. This evidence provides guidance for us to prevent SPs from a dietary perspective, such as moderately increasing fiber intake and reducing red meat intake. Further high-quality research is needed to clarify the role of vitamin D, folate, and calcium intake in SPs.The original contributions presented in the study are included in the article/XL, DX, and ZZ conceived and designed the meta-analysis. ZZ, XZ, XG, and NL searched the literature. ZZ, XZ, XG, NL, and SD extracted the data. ZZ analyzed the data, contributed analysis tools, and wrote the manuscript. XL revised the manuscript. All authors contributed to the article and approved the submitted version.


Associations between reproductive factors and breast cancer (BC) risk vary by molecular subtype . In this systematic review and meta-analysis, we summarized the associations between reproductive factors and BC subtypes.Studies from 2000 to 2021 were included if BC subtype was examined in relation to one of 11 reproductive risk factors: age at menarche, age at menopause, age at first birth, menopausal status, parity, breastfeeding, oral contraceptive (OC) use, hormone replacement therapy (HRT), pregnancy, years since last birth and abortion. For each reproductive risk factor, BC subtype, and study design (case–control/cohort or case-case), random-effects models were used to estimate pooled relative risks and 95% confidence intervals.A total of 75 studies met the inclusion criteria for systematic review. Among the case–control/cohort studies, later age at menarche and breastfeeding were consistently associated with decreased risk of BC across all subtypes, while later age at menopause, later age of first childbirth, and nulliparity/low parity were associated with increased risk of luminal A, luminal B, and HER2 subtypes. In the case-only analysis, compared to luminal A, postmenopausal status increased the risk of HER2 and TNBC. Associations were less consistent across subtypes for OC and HRT use.Identifying common risk factors across BC subtypes can enhance the tailoring of prevention strategies, and risk stratification models can benefit from subtype specificity. Adding breastfeeding status to current BC risk prediction models can enhance predictive ability, given the consistency of the associations across subtypes.The online version contains supplementary material available at 10.1186/s12885-023-11049-0. Breast cancer (BC) is the most common cancer type and a leading cause of cancer death among women globally , 2. EstaBC is recognized as a heterogeneous cancer due to differences in tumor and genomic features that indicate different etiology and prognosis . Based oThis systematic review and meta-analysis were conducted according to the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) recommendations Fig.  11, 12], 1211, 1Included studies focused on BC subtypes and reproductive factors of interest. Studies were excluded if they were published in languages other than English; full texts were not available; examined BC subtypes other than luminal A, luminal B, HER-2, TNBC, or outcomes other than BC subtypes; only analyzed non-reproductive risk factors or none of our a priori reproductive risk factors of interest; if effect estimates  were unavailable; or were study designs other than case–control, case-case, or cohort.Two authors  reviewed the titles, abstracts, and full text of all studies retrieved from electronic databases. Discrepancies in selection were resolved by consensus, and disagreements were resolved in consultation with a third author (XM). A total of 12,243 studies were retrieved from database searches. We excluded 7,702 duplicate articles, and 4,396 articles after the title and abstract review, resulting in 145 articles for full-text review. Of these, 40 studies were excluded due to non-relevant outcomes ; 11 studies due to other subtype classifications; 12 studies due to study design; 4 full texts were not available in English; and 3 studies lacked reproductive variables, resulting in 75 eligible studies. Of these, 56 studies reported estimates for the association between reproductive factors and BC subtypes and were included in meta-analysis  and independently reviewed and verified based on original full texts by the two other authors (SK and AJ). Information on author, year, study design, study characteristics , BC subtypes , reproductive factors, covariates, and corresponding risk estimates and their 95% confidence intervals (CI) were recorded.I2-statistic was used to examine the proportion of variation between studies due to heterogeneity ; [Odds ratios (OR), relative risks (RR), and hazard ratios (HR) for reproductive factors comparing the most extreme  categories were extracted. To ensure consistency in the meta-analysis, if needed, ratio measures were inverted to ensure reference categories matched across studies. Odds ratios and hazard ratios were converted to approximate RRs and 95% CIs. Studies were included in the meta-analysis when at least three studies of the same exposure-outcome combination were available. The meta-analysis was conducted separately for each combination of exposure and BC subtype. All 11 reproductive factors were included in the systematic review, but only eight that were examined in at least three studies were included in the meta-analysis: 1) age at menarche, 2) age at menopause, 3) age at first birth, 4) menopausal status, 5) parity, 6) breastfeeding, 7) OC use and 8) HRT use. Pregnancy, years since last birth, and abortion were qualitatively summarized. Study-level results are presented in supplemental materials ;  and the -value); . Analysen = 59), and about 40% of the studies were conducted in the United States. Of the seventy-five studies, thirteen were cohort studies, thirty-six studies were case–control, twenty-two were case-case studies, and four studies were both case–control and case-case study designs. Nineteen of the seventy-five studies reported data that could not be meta-analyzed, resulting in fifty-six studies included in meta-analysis. Four cohort studies [Seventy-five studies met the inclusion criteria and were deemed suitable for the systematic review Fig. . The cha studies –19 that Forty-six studies evaluated the association between age at menarche and BC subtypes  and 14% lower risk of TNBC . Associations with luminal B  and HER2-overexpressing  subtypes were lower but not significant. There was evidence of heterogeneity and publication bias among the studies . In analysis of thirteen case-case studies .In meta-analysis of two cohort and twenty-seven case–control studies Fig. , later vTwenty-one studies evaluated the association between age at menopause and BC subtypes . In the cohort studies and eleven case–control studies , 85–88, I2 = 56.1%, p = 0.025), 6% higher risk of luminal B  and 13% higher risk of HER2-overexpressing , while the association with TNBC subtype . In the case-case analyses .In meta-analysis of two cohort and eleven case–control studies Fig. , later vForty-three studies evaluated the association between age at first birth and BC subtypes . In the cohort and case–control analysis , 85, 88,I2 = 87.6%, p = 0.000), 19% for luminal B  and 15% for HER2 , while there was no difference in risk for TNBC . There was heterogeneity among the studies and evidence of publication bias . In meta-analysis of case-case studies  and 16% lower risk of TNBC . Lower risk was also observed in studies evaluating luminal B but the association was not significant . There was heterogeneity among the studies but no evidence of publication bias .In the meta-analysis of two cohort and twenty-six case–control studies Fig. , later aEighteen studies evaluated the association between menopausal status and BC subtypes . In the cohort and case–control study analysis , 85, 87,I2 = 79.0%, p = 0.001), 20% lower risk of luminal B ; however the association was statistically non-significantly higher for HER2-overexpressing  and lower for TNBC . There was heterogeneity among luminal A studies, as well as evidence of publication bias . In the case-case meta-analysis , though the association was not significant with luminal B  or TNBC . There was significant heterogeneity but no evidence of publication bias .In the meta-analysis of one cohort and six case–control studies Fig.  post/perFifty-three studies evaluated the association between parity and BC subtypes . In the cohort and case–control studies , 85, 86,I2 = 92.4%, p < 0.001) and 39% lower risk of luminal B . There was no significant association for HER2-overexpressing BC  or TNBC . There was no heterogeneity between the studies and no evidence of publication bias . For the sixteen case-case studies . No significant association was found in luminal B  and HER2-overexpressing BC . There was heterogeneity between the groups and evidence of publication bias .In the meta-analysis of three cohort and twenty-six case–control studies Fig. , higher ies Fig. , compareForty-seven studies evaluated the association between breastfeeding and BC subtypes . In the cohort and case–control studies , 83, 88,I2 = 93.6%, p < 0.001), 29% lower risk of luminal B , 23% lower risk of HER2-overexpressing , and 42% lower risk of TNBC  with significant heterogeneity between the studies, but no evidence of publication bias . In case-case studies , while luminal B breast cancer  and HER2-overexpressing BC  results were not significant. There was heterogeneity between the studies reporting HER2 subtype but no evidence of publication bias .In the meta-analysis of thirty studies Fig. , ever/loies Fig.  comparedTwenty-three studies evaluated the association between OC use and BC subtypes . In the case–control analysis , 66, 80,I2 = 26.0%, p = 0.174). The higher association was not statistically significant for luminal A , luminal B  or HER2 . There was no heterogeneity between these studies, but there was publication bias . In the meta-analysis of nine case-case studies . There was a higher risk of HER2-overexpression  and TNBC subtypes , but these were not significant. There was no evidence of heterogeneity or publication bias .In the meta-analysis of fifteen case–control studies Fig. , ever/loTwenty-one studies evaluated the association between HRT use and BC molecular subtypes . In the cohort and case–control analysis , 73, 80,I2 = 90.3%, p < 0.001) and 23% higher risk of luminal B , however no significant association with HER2-overexpressing  and TNBC  with significant heterogeneity between the studies, but no publication bias . For the five case-case studies .In the meta-analysis of eleven case–control studies and two cohort studies Fig. , ever usies Fig. , any useFour studies , 72, 89 Six studies , 59, 68 Five studies , 85, 87 Out of eight reproductive factors included in this meta-analysis, three were consistently associated with BC risk across subtypes: age at menopause, age at first birth, and breastfeeding, with the strongest associations observed in the luminal A subtype among case–control and cohort studies Table . The assThis systematic review and meta-analysis summarizes the current published literature examining eleven reproductive factors and the risk of breast cancer molecular subtypes. To ensure consistency in contemporary classifications of BC based on molecular subtypes, the scope of this review was limited to luminal A, luminal B, HER2-overexpressing, and TNBC/basal-like breast cancer. In the 75 included studies, we observed that the strength and consistency of associations across subtypes differed by risk factors.Late age at menarche was protective for all four BC subtypes. However, the other factors were most consistently associated with the luminal A subtype. Except for OC use, most of the associations between reproductive risk factors were observed for the luminal subtypes, similar to a prior systematic review showing stronger associations among hormone receptor-positive BC .The prior meta-analysis on this topic evaluated three risk factors: parity, age at first birth, and breastfeeding . Our metHigher parity was protective against BC risk regardless of study design, except for TNBC, where no significant association was observed across 27 studies. Lambertini et al. and Islami et al. observed that breastfeeding was associated with a lower risk of luminal and TNBC , 8, consThis review identified several gaps in the literature on luminal A, luminal B, HER2-overexpressing, and TNBC/basal-like subtypes and reproductive factors. History of abortion, pregnancy, and years since last birth could not be included in the meta-analysis due to an insufficient number of studies examining risk associations by molecular subtype. However, the existing studies indicate that women with a history of abortion have an increased risk of BC regardless of subtype, even though studies on abortion with BC subtypes are lacking . PregnanAccording to the American Cancer Society preventative guidelines, not having children, not breastfeeding, use of OC, and use of HRT after menopause increase the risk of breast cancer . HoweverPotential risk of biases of included studies should be considered. Meta-analytic associations between age at menarche, age at first birth, menopausal status, and parity with BC subtypes were often in opposing directions for case–control/cohort studies compared to case-case studies. Selection bias in the case–control/cohort studies is possible; additionally, interpretation of case-case studies has its limitations, as true risks relative to a cancer-free population cannot be directly estimated. Further, some of the results were unexpected in regards to the hypothesized associations in relation to biological mechanisms. For instance, hormone-dependent subtypes are more common in post-menopausal women, though we observed the opposite in case–control/cohort studies. Misclassification of menopausal status may be a contributing factor to these findings. Additionally, exposure to OC is expected to increase hormone-dependent BC, though we found a protective effect for luminal B in the case-case studies. Comprehensive data on OC formulation was not available for our analysis, though warrants further investigation to assess heterogeneity of effect by estrogen and progestin dosages. This review has several limitations: first, eligible studies were restricted to papers published in English, which may have excluded relevant studies from other populations. We did not conduct a formal risk of bias assessment, though we have discussed potential biases in the observed associations. Additionally, there were not enough studies examining pregnancy, abortion, and time since last birth associated with luminal breast cancer and TNBC subtypes, and we only included studies from the year 2000 to 2021. However, this study serves as the first comprehensive review and meta-analysis of the association between luminal A, luminal B, HER2-overexpressing, and TNBC subtypes and a wide range of reproductive factors.This study offers a comprehensive and up-to-date evaluation of the scientific literature on the associations between reproductive factors and luminal A, luminal B, HER2-overexpressing, and TNBC molecular subtypes. We identified a need for additional studies examining abortion as a risk factor and studies examining reproductive risk factors for BC subtypes stratified by patient menopausal status. Across all reproductive factors examined, age at menarche, age at first birth, parity, pregnancy, and breastfeeding showed relatively consistent risk associations across breast cancer subtypes. Considering this finding, common risk prediction models may be improved upon with the inclusion of breastfeeding status as a predictor.Additional file 1:Supplemental Table 1. Summary statistics for studies included after full-text review (N=75). Supplemental Figure 1. Age at menarche by molecular subtypes of Breast cancer. Supplemental Figure 2. Age at menarche by molecular subtypes of Breast cancer . Supplemental Figure 3. Age at menopause by molecular subtypes of Breast cancer. Supplemental Figure 4. Age at menopause by molecular subtypes of Breast cancer . Supplemental Figure 5. Menopausal status by molecular subtypes of Breast cancer. Supplemental Figure 6. Menopausal status by molecular subtypes of Breast cancer . Supplemental Figure 7. Age at first birth by molecular subtypes of Breast cancer. Supplemental Figure 8. Age at first birth by molecular subtypes of Breast cancer . Supplemental Figure 9. Parity by molecular subtypes of Breast cancer. Supplemental Figure 10. Parity by molecular subtypes of Breast cancer . Supplemental Figure 11. Breastfeeding by molecular subtypes of Breast cancer. Supplemental Figure 12. Breastfeeding by molecular subtypes of Breast cancer . Supplemental Figure 13. OC use by molecular subtypes of Breast cancer. Supplemental Figure 14. OC use by molecular subtypes of Breast cancer . Supplemental Figure 15. HRT use by molecular subtypes of Breast cancer. Supplemental Figure 16. HRT use by molecular subtypes of Breast cancer .Additional file 2. Supplemental Appendix.