In healthcare, a gap exists between what is known from research and what is practiced. Understanding this gap depends upon our ability to robustly measure research utilization.The objectives of this systematic review were: to identify self-report measures of research utilization used in healthcare, and to assess the psychometric properties  of these measures.Standards for Educational and Psychological Testing.We conducted a systematic review of literature reporting use or development of self-report research utilization measures. Our search included: multiple databases, ancestry searches, and a hand search. Acceptability was assessed by examining time to complete the measure and missing data rates. Our approach to reliability and validity assessment followed that outlined in the r coefficients > 0.80. No validity information was reported for 12 of the 60 measures. The remaining 48 measures were classified into a three-level validity hierarchy according to the number of validity sources reported in 50% or more of the studies using the measure. Level one measures (n = 6) reported evidence from any three (out of four possible) Standards validity sources . Level two measures (n = 16) had evidence from any two validity sources, and level three measures (n = 26) from only one validity source.Of 42,770 titles screened, 97 original studies (108 articles) were included in this review. The 97 studies reported on the use or development of 60 unique self-report research utilization measures. Seven of the measures were assessed in more than one study. Study samples consisted of healthcare providers (92 studies) and healthcare decision makers (5 studies). No studies reported data on acceptability of the measures. Reliability was reported in 32 (33%) of the studies, representing 13 of the 60 measures. Internal consistency (Cronbach's Alpha) reliability was reported in 31 studies; values exceeded 0.70 in 29 studies. Test-retest reliability was reported in 3 studies with Pearson's i.e., the Standards) in future research utilization measurement studies.This review reveals significant underdevelopment in the measurement of research utilization. Substantial methodological advances with respect to construct clarity, use of research utilization and related theory, use of measurement theory, and psychometric assessment are required. Also needed are improved reporting practices and the adoption of a more contemporary view of validity ( Clinical and health services research produces vast amounts of new research every year. Despite increased access by healthcare providers and decision-makers to this knowledge, uptake into practice is slow ,2 and haRecognition of, and a desire to narrow, the research-practice gap, has led to the accumulation of a considerable body of knowledge on research utilization and related terms, such as knowledge translation, knowledge utilization, innovation adoption, innovation diffusion, and research implementation. Despite gains in the understanding of research utilization theoretically ,4, a lari.e., focus on specific studies, people, issues, or organizations) when measuring knowledge utilization. Shortly thereafter, Dunn [We located three theoretical papers explicitly addressing the measurement of knowledge utilization (of which research utilization is a component) -11, and er, Dunn , proposeer, Dunn  providedet al. [The only review of research utilization measures to date was conducted in 2003 by Estabrooks et al. . The revThe four papers -12 discui.e., used to measure research utilization by healthcare providers, healthcare decision makers, and in healthcare organizations); and assess the psychometric properties of these measures.The study reported in this paper is a systematic review of the psychometric properties of self-report measures of research utilization used in healthcare. Specific objectives of this study were to: identify self-report measures of research utilization used in healthcare . We also excluded articles reporting on the use of one specific-research-based practice if the overall purpose of the study was not to examine research utilization.Studies were included that met the following inclusion criteria: reported on the development or use of a self-report measure of research utilization; and the study population comprised one or more of the following groups--healthcare providers, healthcare decision makers, or healthcare organizations. We defined research utilization as the use of research-based  information. This information could be reported in a primary research article, review/synthesis report, or a protocol. Where the study involved the use of a protocol, we required the research-basis for the protocol to be apparent in the article. We excluded articles that reported on adherence to clinical practice guidelines, the rationale being that clinical practice guidelines can be based on non-research evidence  and assessed the reference lists of all retrieved articles. The final set of included articles was restricted to those published in the English, Danish, Swedish, and Norwegian languages . There were no restrictions based on when the study was undertaken or publication status.We searched 12 bibliographic databases; details of the search strategy are located in Additional File Two team members (JES and HMO) independently screened all titles and abstracts n = 42,770). Full text copies were retrieved for 501 titles, which represented all titles identified as having potential relevance to our objectives or where there was insufficient information to make a decision as to relevance. A total of 108 articles  comprised the final sample. Disagreements were resolved by consensus. When consensus could not be reached, a third senior member of the review team  acted as an arbitrator and made the final decision (n = 9 articles). Figure ,770. FulStandards for Educational and Psychological Testing (the Standards) [Content evidence refers to the extent to which the items in a self-report measure adequately represent the content domain of the concept or construct of interest. Response processes evidence refers to how respondents interpret, process, and elaborate upon item content and whether this behaviour is in accordance with the concept or construct being measured. Internal structure evidence examines the relationships between the items on a self-report measure to evaluate its dimensionality. Relations to other variables evidence provide the fourth source of validity evidence. External variables may include measures of criteria that the concept or construct of interest is expected to predict, as well as relationships to other scales hypothesized to measure the same concepts or constructs, and variables measuring related or different concepts or constructs [Standards, validity is a unitary construct in which multiple evidence sources contribute to construct validity. A higher number of validity sources indicate stronger construct validity. An overview of the Standards approach to reliability and validity assessment is in Additional File Two reviewers (JES and HMO) performed data extraction: one reviewer extracted the data, which was then checked for accuracy by a second reviewer. We extracted data on: year of publication, study design, setting, sampling, subject characteristics, methods, the measure of research utilization used, substantive theory, measurement theory, responsiveness (the extent to which the measure can assess change over time), reliability , reported statements of traditional validity , and study findings reflective of the four sources of validity evidence  outlined in the andards) . Contentnstructs . In the Standards, considered best practice in the field of psychometrics [Standards' critical reliability information and four validity evidence sources. To assess relations to other variables, we a priori (based on commonly used research utilization theories and systematic reviews) identified established relationships between research utilization and other  variables , contextual characteristics , organizational characteristics , and interventions . All relationships between research use and external variables in the final set of included articles were then interpreted as supporting or refuting validity evidence. The relationship was coded as 'supporting validity evidence' if it was in the same direction and had the significance predicted, and as 'refuting validity evidence' if it was in the opposite direction or did not have the significance predicted.There are no universal criteria to grade the quality of self-report measures. Therefore, in line with other recent measurement reviews ,15, we dometrics . AccordiStandards, no one source of validity evidence is considered always superior to the other sources. Therefore, in our hierarchy, level one, two, and three measures provided evidence from any three, two, and one validity sources respectively. In the case of single-item measures, only three validity sources are applicable; internal structure validity evidence is not applicable as it assesses relationships between items. Therefore, a single-item measure within level one has evidence from all applicable validity sources.The findings from the review are presented in narrative form. To synthesize the large volume of data extracted on validity, we developed a three-level hierarchy of self-report research utilization measures based on the number of validity sources reported in 50% or more of the studies for each measure. In the In total, 60 unique self-report research utilization measures were identified. We grouped them into 10 classes as follows:1. Nurses Practice Questionnaire (n = 1 Measure)2. Research Utilization Survey (n = 1 Measure)3. Edmonton Research Orientation Survey (n = 1 Measure)4. Knott and Wildavsky Standards (n = 1 Measure)5. Other Specific Practices Indices n = 4 Measures)  8. Parahoo's Measure (n = 1 Measure)9. Estabrooks' Kinds of Research Utilization (n = 1 Measure)10. Other Single-Item Measures (n = 39 Measures)e.g., reading research articles) but not research utilization directly. The 13 proxy measures are: Nurses Practice Questionnaire, Research Utilization Questionnaire, Edmonton Research Orientation Survey, and the ten Other General Research Utilization Indices.Table The majority (n = 54) of measures were assessed with healthcare providers. Professional nurses comprised the sample in 56 studies (58%), followed by allied healthcare professionals , physicians , and multiple clinical staff groups . A small proportion of studies  measured research utilization by healthcare decision makers. The decision makers, in each study, were members of senior management with direct responsibility for making decisions for a healthcare organization and included: medical officers and program directors ; managerMost studies were conducted in North America , followed by Europe . Other geographic areas represented included: Australia , Iran , Africa , and Taiwan . With respect to date of publication, the first report included in this review was published in 1976 . The majOur psychometric assessment involved three components: acceptability, reliability, and validity.Acceptability in terms of time required to complete the research utilization measures and missing data (specific to the research utilization items) was not reported.Reliability was reported in 32 (33%) of the studies  of the studies: two studies assessing the Nurses Practice Questionnaire -35, and ls Table . One stue et al.  had traiStandards. For 12 measures [No single research utilization measure had supporting validity evidence from all four evidence sources outlined in the measures -49, eachSix measures were grouped as level one: Specific Practices Indices (n = 1), General Research Utilization Indices n = 3), and Other-Single Items (n = 2) ; Knott and Wildvasky Standards (n = 1); General Research Utilization Indices n = 4); Specific Practices Indices (n = 2); Estabrooks' Kinds of Research Utilization (n = 1); Past, Present, Future Use (n = 1); and Other Single-Items (n = 6)  of research utilization measures identified fell into level three: Champion and Leach's Research Utilization Survey (n = 1); Edmonton Research Orientation Survey (n = 1); General Research Utilization Indices (n = 3); Specific Practices Indices (n = 1); Past, Present, Future Use (n = 1); and, Other Single-Item Measures (n = 19) Table . The majAs part of our validity assessment, we paid special attention to how each measure 'functioned'. That is, 'were the measures behaving as they should'. All six level one measures and the majority of level two measures (n = 12 of 16) displayed 'relations to other  variables' evidence, indicating that the measures are functioning as the literature hypothesizes a research utilization measure should function. Fewer measures in level three (n = 10 of 26) displayed optimal functioning Table . We alsoOur discussion is organized around three areas: the state of the science of research utilization measurement, construct validity, and our proposed hierarchy of measures.et al. [In 2003, Estabrooks et al.  completeThere is ambiguity with respect to the naming of self-report research utilization measures. For example, similar measures have different names. Parahoo's Measure  and Pettet al. [In the earlier research utilization measurement review, Estabrooks et al.  identifiper se.Research utilization has been, and is likely to remain, a complex and contested construct. Issues around clarity of research utilization measurement stems from four areas: a lack of definitional precision of research utilization, confusion around the formal structure of research utilization, lack of substantive theory to develop and evaluate research utilization measures, and confusion between factors associated with research utilization and the use of research Lack of definitional precision with respect to research utilization is well documented. In 1991, knowledge utilization scholar Thomas Backer  declaredi.e., consists of a series of stages/steps) and research utilization as variable or discrete event (a 'variance' approach). Despite debate in the literature with respect to these two conceptualizations, this review revealed that the vast majority of measures that quantify research utilization do so using a 'variable' approach. Only two measures were identified that assess research utilization using a 'process' conceptualization: Nurses Practice Questionnaire [Innovation Decision Process Theory [et al. and based on Knott and Wildavsky's Standards of Research Use model [et al. [A second reason for the lack of clarity in research utilization measurement is confusion around the formal structure of research utilization. The literature is characterized by multiple conceptualizations of research utilization. These conceptualizations influence how we define research utilization and, consequently, how we measure the construct and interpret the scores obtained from such measurement. Two prevailing conceptualizations dominating the field are research utilization as process  and se model ). Some sse model  and Estase model ,74,80,86e.g., [e.g., [e.g., [et al. [e.g., knowledge translation) fields that requires addressing.A third reason for the lack of clarity in research utilization measurement is limited use of substantive theory in the development of research utilization measures. There are numerous theories, frameworks, and models of research utilization and of related constructs, from the fields of nursing e.g., -96), org, orge.g., [e.g., -99), and, [e.g., ). Howeve, [e.g.,  was deve, [e.g., ). The In [et al.  in theirper se. The Nurses Practice Questionnaire [A fourth and final reason that we identified for the lack of clarity in research utilization measurement is confusion between factors associated with research utilization and the use of research ionnaire  and all ionnaire ,101-105)Foundational to the development of any measure is measurement theory. The two most commonly used measurement theories are classical test score theory, and modern measurement (or item response) theory. Classical test score theory proposes that an individual's observed score on a construct is the additive composite of their true score and random error. This theory forms the basis for traditional reliability theory (Cronbach's Alpha) ,107. Iteet al. [Similar to the previous review by Estabrooks et al. , none ofet al. [Standards as a framework for validity. Using this framework, we scrutinized all results , in terms of whether or not they added to overall construct validity.In the previous review, Estabrooks et al.  concludeSeveral additional limitations in research utilization measurement were also noted as a result of this review. They include: limited reporting of data reflective of reliability beyond standard internal consistency (Cronbach's Alpha) coefficients; limited reporting of study findings reflective of validity; limited assessments of the same measure in multiple (> 1) studies; lack of assessment of acceptability and responsiveness; overreliance on the assessment made in the index  study of a measure; and failure to re-establish validity when modifications are made and/or the measure is assessed in a new population or context.Standards) was taken in this review. Using this approach, validity was conceptualized as a unitary concept with multiple sources of evidence, each contributing to overall (construct) validity [Traditionally, validity has been conceptualized according to three distinct types: content, criterion, and construct. While this way of thinking about validity has been useful, it has also caused problems. For example, it has led to compartmentalized thinking about validity, making it 'easier' to overlook the fact that construct validity is really the whole of validity theory. It has also led to the incorrect view of validity as a property of a measure rather than of the scores (and resulting interpretations) obtained with the measure. A more contemporary conceptualization of validity . As a result, we are not able to declare any of the measures we identified in our review as valid or invalid, but only as more or less valid for selected populations, settings, and situations. This deviates substantially from traditional thinking, which suggests that validity either exists or not.All self-report measures require validity assessments. Without such assessments little to no intrinsic value can be placed on findings obtained with the measure. Validity is associated with the interpretations assigned to the scores obtained using a measure, and thus is intended to be hypothesis-based ,111. Hene.g., Duhem [i.e., based on data collected with an specific measure). From this perspective, construct validity exists or not. In the second half of the 20th century, however, movement away from justification to what was described by Feyerabend [i.e., more or less valid) and can change with the sample, setting, and situation being assessed. This is in line with a more contemporary (the Standards) conceptualization of validity.According to Cronbach and Meehl , constru., Duhem  and Laka., Duhem ). Duhem yerabend  and Kuhnyerabend  as 'nonjStandards [Standards) placed equal weight on all four evidential sources. While we were able to categorize 48 of the 60 self-report research utilization measures identified into the hierarchy, several cautions exist with respect to use of the hierarchy. First, the levels in the hierarchy are based on the number of validity sources reported, and not on the actual source or quality of evidence within each source. Second, some measures in our hierarchy may appear to have strong validity only because they have been subjected to limited testing. For example, the six measures in level one have only been tested in a single study. Third, the hierarchy included all 48 measures that displayed any validity evidence. Some of these measures, however, are proxies of research utilization. Overall, the hierarchy is intended to present an overview of validity testing to date on the research utilization measures identified. It is meant to inform researchers regarding what testing has been done and, importantly, where additional testing is needed.The tandards  providedStandards, we may have concluded lower levels of validity for some measures than actually exist. In the latter case, our findings may reflect poor reporting rather than less validity. Third, our decision to exclude articles that reported on healthcare providers' adherence to clinical practice guidelines may be responsible for the limited number of articles sampling physicians included in the review. A systematic review conducted by Grimshaw et al. [Although rigorous and comprehensive methods were used for this review, there are three study limitations. First, while we reviewed dissertation databases, we did not search all grey literature sources. Second, due to limited reporting of findings consistent with the four sources of validity evidence in the w et al.  on cliniper se. Substantial methodological advances in the research utilization field, focusing in the area of measurement  are urgently needed. These advances are foundational to ensuring the availability of defensible self-report measures of research utilization. Also needed are improved reporting practices and the adoption of a more contemporary view of validity (the Standards) in future research utilization measurement studies.In this review, we identified 60 unique self-report research utilization measures used in healthcare. While this appears to be a large and definite set of measures, our assessment paints a rather discouraging picture of research utilization measurement. Several of the measures, while labeled research utilization measures, did not assess research utilization The authors declare that they have no competing interests.JES, CAE, PG, and LW participated in designing the study and securing funding for the project. JES, CAE, HMO, PG, and LW participated in developing the search strategy, study relevance, and data extraction tools. JES and HMO undertook the article selection and data extraction. All authors participated in data synthesis. JES drafted the manuscript. All authors provided critical commentary on the manuscript and approved the final version.Search Strategy. This file contains the details of the search strategy used for the review.Click here for fileExclusion List by Reason (N = 393). This file contains a list of the retrieved articles that were excluded from the review and the reason each article was excluded.Click here for fileThe Standards. This file contains an overview of the Standards for Educational and Psychological Testing Validity Framework and sample predictions used to assess 'relations to other variables' validity evidence according to this framework.Click here for fileDescription of Other Specific Practices Indices and Other General Research Use Indices. This file contains a description of the four measures included in the class 'Other Specific Practices Indices' and the ten measures included in the class 'Other General Research Use Indices'.Click here for fileReported Reliability of Self-Report Research Utilization Measures. This file contains the reliability coefficients reported in the included studies.Click here for fileSupporting Validity Evidence by Self-Report Research Utilization Measure. This file contains the detailed validity evidence on each included self-report research utilization measure.Click here for file


The aim of the study was to verify the association between body composition and physical fitness with bone status in children and adolescents.A cross-sectional study was conducted with 300 healthy students . Weight, height, fat and fat-free mass, and percentage of body fat (%BF) were evaluated, as were physical fitness  and maximum oxygen consumption. Bone parameters  and the Ultrasound Bone Profile Index (UBPI) were evaluated using DBM Sonic BP ultrasonography.2 = 0.274; p < 0.001), and pubertal maturation only for UBPI . For girls, age and %BF were identified as significant independent variables for AD-SoS and UBPI  but height only for AD-SoS .In the study group, girls had higher bone parameter values than boys. A univariate analysis assessed in a stepwise multiple regression model was conducted. It showed that for boys, the %BF and height were significant independent variables for AD-SoS and UBPI, but the horizontal jump test only for AD-SoS  are independent positive predictors for the bone parameters in both boys and girls. %BF is an independent negative predictor. For boys, the horizontal jump test was an independent positive predictor for AD-SoS, indicating that physical fitness related to the neuromotor system can influence the amount of bone present. Physical activity is a key factor for improving physical fitness related to health, and has been indicated as a major determinant of bone mass throughout life . AccordiSeveral methods have been used to measure bone mass, including dual-energy X-ray absorptiometry (DXA) and peripheral quantitative computed tomography. However, in recent years, quantitative ultrasonography (QUS) of the phalanges has been used for indirect evaluation of bone tissue. The rationale is based on the results of a variety of experiments suggesting that US parameters provide information not only about the quantity but also about bone architecture and elasticity . QUS hasThe aim of the present study was to verify the associations between body composition and physical fitness using the bone status in children and adolescents according to their sex.n = 2), use of drugs that might interfere with body composition or bone mass (n = 0), no consent to participate from the parents or students (n = 22), and no execution of any test or absence on any evaluation day (n = 300). The final sample consisted of 300 students: 148 boys (49.4%) and 152 girls (50.6%) aged 11–16 years. Chronological ages were established by calculating decimal ages, with reference to the birth date and survey date, using decimal intervals between 0.50 and 0.49 [This cross-sectional study was conducted on 624 students from the Bradesco Foundation School  who were invited to participate. The sample was selected intentionally. The exclusion criteria were the presence of physical disability (permanent or temporary) that precluded conducting the evaluations (2).Weight (kg) and height (m) were measured using standardized techniques . Body maTriceps and medial calf skinfolds were measured according to the American College of Sports Medicine recommendations  using a Pubertal maturation was performed by self-assessmen, using specific figures for breast stage (B1–5) for girls and male genitalia (G1–5) for boys according to the criteria of Marshall and Tanner ,16. The The third generation of DBM Sonic BP equipment  was used for determining the bone parameters: amplitude-dependent speed of sound (AD-SoS) and Ultrasound Bone Profile Index (UBPI). The equipment is fitted with a probe that attaches two transducers (transmitter and receiver). The probe is positioned at the distal metaphysis of each of the last four proximal phalanges (II–V) of the nondominant hand. The transducer transmitter emits a sound wave of 1.25 MHz, and the transducer receiver picks up the signal and assesses the speed of propagation of sound through the phalange . The quaThe use of this method is interesting because the assessed site  has high metabolic activity during all stages of life and has great similarity to the microstructures of the lumbar spine . The equIn vivo short-term precision was assessed based on the root mean square of the coefficient of variation (RMS-CV) for 80 measurements made in 10 healthy young persons  measured four times each. It was calculated according to Bonnick et al. [One evaluator (E.M.G.) performed all of the US measurements. k et al. . The RMS2peak). A team of experienced previously trained evaluators applied the tests. To avoid random errors, only one evaluator was responsible for the information obtained from each test.To determine physical fitness, a 1-min crunch abdominal exercise , a sit-and-reach exercise (flexibility), and horizontal jump tests (power of the lower limbs) were performed according to the standardizations described by the American Alliance for Health, Physical Education, Recreation and Dance ,19. The 2peak) on the dependent variables . The results were considered statistically significant at p < 0.05.The statistical analyses were performed using SPSS version 16.0 software . The normality of the data was verified using the Kolmogorov Smirnov test. When normal distribution was not observed, the transformation of Blom was completed to achieve homogeneity and normality of variables. Multivariate analysis of covariance was used to verify the differences in dependent variables between sex and pubertal development. Age was used as a covariate. The Bonferroni post-hoc test was used when necessary. Multivariate normality and homogeneity of variances and covariances were evaluated. Pearson’s correlation coefficient was used to verify correlations between the QUS parameter and anthropometric variables, body composition, and physical fitness. Stepwise multivariate linear regression analyses were performed to determine the possible effects of a group of independent variables , and pubertal maturation for UBPI . In the male model, %BF and height were identified as significant independent variables for AD-SOS and UBPI, horizontal jump only for AD-SoS , and pubertal maturation only for UBPI . For girls, age and %BF were identified as significant independent variables for AD-SOS and UBPI , but height only for AD-SOS .Significant variables identified by the univariate analysis were further assessed in a stepwise multiple regression model. Separate models were developed for AD-SoS and UBPI and are shown in Table Our study provides specific QUS parameters that evaluate bone status (AD-SoS and UBPI) by sex and pubertal maturation and their association with physical fitness and body composition in Brazilian children and adolescents. It is the first research using phalange US and physical fitness in Brazilian adolescents.Girls showed significantly higher QUS values than boys. Studies using DXA have shown that girls have higher BMD values than boys at trabecular sites ,21,22. TSeveral studies have investigated the association between the level of physical activity and bone mass in children and adolescents ,28-31. H2peak and QUS parameters. In fact, the results suggested that the bone mass differences between males and females could probably be explained by differences in physical fitness and lean mass [This effect on bone parameters related to physical strength variables is in accord with other studies. For instance, Ginty et al.  showed aean mass ,35. The ean mass ,36, otheean mass ,37 was aean mass .Regarding the variable fat, our study showed a negative correlation of AD-SoS and UBPI for %BF and BMI in boys and %BF for AD-SoS in girls. These results are consistent with those of other studies ,39, whicThe present study demonstrated that in a general regression analysis, sex and measurements related to growth  were positive predictors for both AD-SoS and UBPI. In boys, the standing long jump and height were positive predictors for AD-SOS and pubertal maturation and height for UBPI. Conversely, for girls, age was a positive predictor for AD-SoS and UBPI but height only for Ad-SOS. An interesting finding was that for all groups the %BF was a negative predictor for both AD-SoS and UBPI. These findings are in accord with data from previous studies ,24,40-42In our study, despite the lean mass having shown a relation with AD-SoS in boys and girls and with UBPI in boys, it did not appear as a predictor in the regression analysis. However, the horizontal jump as a positive predictor in boys showed that muscle strength positively influences AD-SoS, and the %BF has a negative influence on QUS parameters in both sexes.The present study has some limitations. They include the large number of subjects lost from the original sample, self-evaluation of sexual maturation, no comparison of US data with DXA data, no implementation survey of fractures, and no evaluation of the ethnicity of the subjects.The results of the present study demonstrated that QUS of phalanges parameters are correlated with growth variables such as age, height, and pubertal maturation. Regarding physical fitness, only the variables related to muscle strength, especially in boys, showed an association with QUS parameters. In addition, fat mass demonstrated a negative association with AD-SoS and UBPI.The authors declare that they have no competing interests.All of the authors have made substantial contributions to the study. AMM: conception and design of the study, acquisition of data, analysis and interpretation of data; drafting the article; final approval of the version to be submitted. EMG: acquisition of data, analysis and interpretation of data; revising the manuscript critically for important intellectual content; final approval of the version to be submitted. VJOB: interpretation of data; revising the manuscript critically for important intellectual content; final approval of the version to be submitted. GG-J: conception and design of the study; acquisition of data analysis and interpretation of data; revising the manuscript critically for important intellectual content; final approval of the version to be submitted. All authors read and approved the final manuscript.The pre-publication history for this paper can be accessed here:http://www.biomedcentral.com/1471-2431/13/117/prepub


China has one of the world's largest health insurance systems, composed of government-run basic health insurance and commercial health insurance. The basic health insurance has undergone system-wide reform in recent years. Meanwhile, there is also significant development in the commercial health insurance sector. A phone call survey was conducted in three major cities in China in July and August, 2011. The goal was to provide an updated description of the effect of health insurance on the population covered. Of special interest were insurance coverage, gross and out-of-pocket medical cost and coping strategies.Records on 5,097 households were collected. Analysis showed that smaller households, higher income, lower expense, presence of at least one inpatient treatment and living in rural areas were significantly associated with a lower overall coverage rate. In the separate analysis of basic and commercial health insurance, similar factors were found to have significant associations. Higher income, presence of chronic disease, presence of inpatient treatment, higher coverage rates and living in urban areas were significantly associated with higher gross medical cost. A similar set of factors were significantly associated with higher out-of-pocket cost. Households with lower income, inpatient treatment, higher commercial insurance coverage, and living in rural areas were significantly more likely to pursue coping strategies other than salary.The surveyed cities and surrounding rural areas had socioeconomic status far above China's average. However, there was still a need to further improve coverage. Even for households with coverage, there was considerable out-of-pocket medical cost, particularly for households with inpatient treatments and/or chronic diseases. A small percentage of households were unable to self-finance out-of-pocket medical cost. Such observations suggest possible targets for further improving the health insurance system. The basic health insurance is run by the central and local governments. It consists of three schemes for different groups of people and takes different forms in rural and urban China. In rural China, the new rural cooperative medical care system (NCMS) was designed as a mutual help and risk-pooling health protection system Commercial health insurance is supplementary to basic health insurance, targeting mainly the upper class In the literature, one of the most relevant studies is The study design strategy is as follows. To provide an up-to-date description, a survey was conducted. To complement existing studies, the survey was focused on three major cities and surrounding areas. To provide a more comprehensive description, both basic and commercial insurances were studied. Our analysis strategy is as follows. As a broad coverage is needed for health insurance to be effective, our first set of analysis focused on coverage. A main objective of health insurance was to reduce cost for the insured. Thus, in the second set of analysis, we studied the cost, both gross and out-of-pocket, as they have different implications. As considerable cost might still exist after insurance reimbursement, the third set of analysis was on how households dealt with such cost. The three sets of analyses are closely linked.The survey was conducted by staff at the Data Mining Center, Xiamen University. A Computer-Assisted Telephone Survey System (CATSS) was used. The collection of phone numbers for survey was obtained from China Telecom Corp. Ltd. and China Unicom Corp. Ltd. The following method was used for RDD  selection of samples. We draw Mitofsky-Waksberg At the beginning of each phone call, the survey staff would gather basic information. A household would be excluded if (1) the interviewee refused to participate, or (2) it was not in the three surveyed cities (defined by “Hukou”), or (3) the interviewee was less than eighteen years old, or (4) the interviewee could not provide reliable information on the household . Verbal consent was obtained from each interviewee and recorded using a voice recording software package. A total of fifteen questions were asked. Some questions, for example household size and insurance coverage, were “snapshots” at the time of survey. Other questions, for example household income, expense, illness conditions and coping strategies, reflected the accumulation over twelve months. On average, one interview lasted five minutes.Data was de-identified prior to analysis. We examined data and found no obviously unreasonable measurements. In this study, we were interested in three aspects of health insurance. The first was coverage. Here we analyzed overall coverage as well as basic and commercial insurance coverage separately. For each household, we dichotomized coverage rates at 50% and created dummy variables. The second aspect was medical cost, including gross medical cost (before insurance reimbursement) and net out-of-pocket medical cost (after insurance reimbursement). In the survey, medical cost was classified into five categories . The third aspect was coping. Strategies for dealing with high and extremely high cost were analyzed separately. We examined the differences between data collected in difference cities using ANOVA and Chi-squared tests and determined that it was appropriate to combine data. As the response variables of interest were categorical, logistic regression was the main analysis tool. Model diagnostics was conducted, and no serious deviation from the model assumption was detected. All analysis was conducted using S-Plus Version 8.2 (TIBCO Software Inc).A survey was conducted by staff at the Data Mining Center, Xiamen University, China, in July and August, 2011. The study was approved by a research ethics review committee at Xiamen University. Three major cities, Beijing, Shanghai and Xiamen, and their surrounding rural areas were included in the survey . BeijingSummary statistics on coverage were first computed. The 5,097 households contained 18,889 members, among which 15,555 (82.35%) and 6,568 (34.77%) were covered by basic and commercial insurance, respectively. 1,147 households had full, 27 had zero, and the rest had partial coverage. With a high overall coverage rate, it was interesting to notice that a very small number of households had no coverage at all. This small group deserves special attention in future studies. In the literature, there were many studies on analyzing percentage data. However, our literature review suggested that none was able to properly accommodate data with “two non-ignorable corners” (nonzero percentages of zeros and ones). We dichotomized the coverage ratio at 50%, and studied the contrast of “majority” versus “minority with coverage”. Out of the 5,097 households, 4,437 (87.05%) had more than 50% of the household members covered. 4,154 (81.50%) and 977 (19.17%) households had more than 50% of the members covered by basic and commercial health insurance, respectively. The calculated basic insurance coverage rate was lower than that provided by the central government but higher than that reported in Household summary statistics were computed for the whole cohort and subgroups generated based on insurance coverage rates . We condSummary statistics suggested that household size, income, expense, health conditions and location of household were potentially associated with coverage and financial impact of health insurance and warranted further analysis. This set of variables was comparable to those in published studies In the coverage analysis , resultsBroad coverage is needed for insurance to be effective. Analysis results presented above provide possible suggestions for future targets to raise coverage. Interesting target populations may include large households and households in a certain income range. It is of interest to note the negative association between coverage and inpatient treatment. Cost associated with inpatient treatment is an important component of catastrophic health expenditure Medical cost had been rising significantly in China In the first set, the contrasting groups were “medical cost >1 K Yuan” versus “≤1 K”. Among the 5,097 households, 3,019 (59.23%) were in the high-cost group. Multivariate analysis suggested that larger households tended to have higher medical expense, which was reasonable as the expense was not normalized by household size. Household income was significantly associated with medical expense. For example, with the lowest income group as reference, the highest income group had an odds ratio of 1.808 . This was partly caused by the lack of normalization. Both presences of chronic disease and inpatient treatment led to higher medical cost . Basic insurance coverage was not significantly associated with medical expense. Commercial insurance coverage was significantly associated with medical expense . This significant positive association reflected its selection nature, with households in worse health conditions more likely to purchase commercial insurance. Urban residents had higher medical cost, as urban health care facilities tended to be more expensive. City was included as a covariate and found to be significant. In particular, both Beijing and Shanghai households tended to have lower expense, compared with Xiamen. The odds ratios for Beijing and Shanghai were similar .In the second set of analysis, we contrasted low and moderate cost group (≤5 K Yuan) with the extremely high cost group (>5 K Yuan). Among the 5,097 households, 473 (9.28%) belonged to the extremely high cost group. Compared with the first set of analysis, the effect of household size was similar. The associations between income levels and expense were mostly insignificant, expect for the “between 50 K and 100 K” group . Presences of chronic disease and inpatient treatment were significantly associated with higher expense . Neither basic nor commercial insurance coverage rate was significant. The difference between urban and rural residents was not significant, and there was no significant difference among cities.The fast rising of medical cost is a challenge encountered by not only China but also developed countries like the United States. Our analysis may assist identifying the factors that contributed to higher medical cost. As a limitation of this study and surveys of a similar kind, we were only able to identify the factors associated with cost, but not necessarily the underlying causal factors. Although there is no one-to-one correspondence between cost and quality of care, they tend to be correlated. An important aspect of basic health insurance is to ensure that service is delivered to all insured in an equal manner. From this perspective, our analysis identified populations with lower medical cost and possibly lower quality of care. From a policy point of view, for example, it is of interest to investigate how to make healthcare more affordable and more accessible to low-income group and people living in the rural area.Gross medical cost can be of significant interest to the government and insurance industry. For households, net out-of-pocket cost – medical cost after insurance reimbursement – is of more importance. It provides a more direct measure of the impact of illness conditions and effect of medical insurance. It has been observed that in several Asian countries, high out-of-pocket cost has been an important contributing factor for poverty In the analysis of low (≤1 K Yuan) versus high (>1 K Yuan) cost, household size was significant, with larger households more likely to have higher cost (odds ratio 1.198). With income <30 K as reference, three higher income levels were significantly associated with higher cost. The only insignificant level was “between 100 K and 150 K”. Presences of chronic disease and inpatient treatment were significantly associated with higher cost . This result suggested that health insurance was not able to fully remove the financial burden caused by illness. Similar observations had been made in recent studies conducted in South Korea and Vietnam Results from the comparison of low and moderate cost (≤5 K Yuan) versus extremely high cost (>5 K Yuan) were comparable to those described above. Household size remained significant. However, most associations between income levels and cost were not significant. Only the “between 30 K and 50 K” group showed a significant higher level of cost, compared with the reference group. Presences of chronic disease and inpatient treatment remained significant, with the odds ratios slightly lower than those for the gross cost. Basic insurance was not significant, whereas a higher commercial insurance coverage rate was positively correlated with higher cost (odds ratio 2.977). Urban residents tended to have a lower probability of extremely high out-of-pocket cost. A possible explanation was that the reimbursement system in urban areas was more developed, leading to a higher amount and percentage of reimbursement and hence a lower probability of extremely high cost. In addition, the urban basic health insurance system had a better coverage for catastrophic expense. Note that this result needs to be interpreted cautiously, as only 137 records fell in the category of “rural residents and extremely high out-of-pocket cost”. Residents of Shanghai had a higher probability of having extremely high cost , compared to Xiamen.With a considerable amount of out-of-pocket cost, the means that households pay for the cost are of interest. In the survey, the interviewees were asked what was the most important financial source to pay for out-of-pocket cost, with answers including (A) salary from last month, (B) saving, (C) help from family and friends, (D) loan, and (E) reducing daily living cost. In the whole cohort, percentages of answering (A)–(E) were 60.53%, 34.06%, 0.57%, 0.65% and 4.20%, respectively. The majority of the households were able to self-finance out-of-pocket cost (answers A and B). Two sets of analyses were conducted . In the In the first set of analysis, we coded the outcome variable “using salary” as 0 and “using other means” as 1. In the second set of analysis, we coded the outcome variable “salary or saving” as 0 and “other means” as 1. The majority of surveyed households were able to self-finance out-of-pocket medical cost using salary and saving, without having to rely on outside financial sources or reducing daily living cost. However, there were still 5% of the households that warranted further attention. Future policy development should focus on this group that may suffer a long financial impact caused by out-of-pocket medical cost.net effect of basic insurance could not be investigated – it is thus not clear how the government should tune the basic insurance policy.Investigating both basic and commercial insurance might provide a more comprehensive description of households’ insurance status. However, a tradeoff is that the policy implications of the study was less lucid as the Another limitation was that because of limitation in data collection, we were not able to fully validate the representativeness of samples. Summary statistics on the household size and income data roughly matched data published by the central government on the three cities. However, as some data (e.g. income) were collected using categorical variables, we were not able to conduct a rigorous test on the representativeness.In the survey, all the samples were drawn from three major cities and surrounding areas. As can be seen from the GDP figures and city locations , the sammost important coping strategy.With phone call surveys, we were able to collect a large number of samples at affordable cost. However, the data collected were inevitably less detailed. Particularly, the data were either snapshots at the time of survey or aggregated data over twelve months. Such data had limitations. For example, the insurance status  and household size might change over time. The aggregated data, including inpatient treatment, presence of chronic disease, out-of-pocket cost and coping strategies, could not describe variations across different disease episodes and their differences in financial consequences. In addition, measuring out-of-pocket cost as a single item might result in a biased estimation  and reached similar conclusions.The healthcare sector in China is undergoing tremendous reform. Its development may provide valuable information for reform in other developing countries. In this study, we reported a phone call survey in three major cities. Our findings suggested that the surveyed population were well covered by basic and commercial health insurance, although there was still room for improvement. Possible target subpopulations to further increase coverage were identified. This study also identified factors that contributed to high gross and net out-of-pocket medical cost. More attention should be paid to these factors in the process of reform. We also identified the group that had to cope with out-of-pocket medical cost by borrowing or reducing daily living cost. Policy interventions should be developed targeting that group. Despite several limitations, this study may provide valuable information on the current status of China's health insurance and serve as basis for future policy development.


Although lower urinary tract symptoms (LUTS) seem to be related to cardiovascular disease (CVD) in men, it is unclear whether this relationship is unbiased. In order to investigate this relationship, we used longitudinal data for establishing the possible predictive value of LUTS for the development of CVD in a primary care population.We performed a registry study using data from the Registration Network Groningen (RNG). All data from men aged 50 years and older during the study period from 1 January 1998 up to 31 December 2008 were collected. Cox proportional hazard regression analysis was used to determine the association between the proportions of CVD (outcome) and LUTS in our population.Data from 6614 men were analysed. The prevalence of LUTS increased from 92/1000 personyears (py) in 1998 up to 183/1000 py in 2008. For cardiovascular diseases the prevalence increased from 176/1000 py in 1998 up to 340/1000 py in 2008. The incidence numbers were resp. 10.2/1000 py (1998) and 5.1/1000 py (2008) for LUTS, and 12.9/1000 py (1998) and 10.4/1000 py (2008) for CVD. Of all men, 23.2% reported CVD . The hazard ratio of LUTS for cardiovascular events, compared to no LUTS, in the adjusted multivariate model, was 0.921.Based on the results, LUTS is not a factor that must be taken into account for the early detection of CVD in primary care. Cardiovascular diseases (CVD) are a major cause of morbidity and mortality. Worldwide, CVD is responsible for an estimated 35 million deaths each year ,2.The recently published guidelines ‘Prevention of cardiometabolic diseases’  urge genResearch in the past decade has revealed an association between CVD and erectile dysfunction (ED) in community based and clinical studies -7. The iLUTS and CVD share some risk factors such as obesity, diabetes, hypertension, smoking, and ageing ,24. The The hypothesis that LUTS could be associated with the development of CVD still needs to be confirmed in primary care. The only evidence to date is one longitudinal, community-based study conducted on a small group of men with severe LUTS . TherefoWe performed a registry study using data from the Registration Network Groningen (RNG), one of several registration networks in the Netherlands. These registration networks carry out research on data derived from the electronic registration of daily patient care in their participating general practices. The Registration Network Groningen was established in 1989, and has three practices in the north of the Netherlands, with an annual population of approximately 30,000 patients .In the Netherlands, the GP is the gatekeeper in the Dutch health-care system controlling access to specialized medical care. Virtually all non-institutionalized Dutch citizens are registered with a GP so the total practice population represents the general population .GPs working in the RNG-practices use a structured medical record, in which all patient contacts are registered. This includes reason for encounter, medical diagnosis  ), applieFrom the RNG registration, we selected all data from men aged 50 years and older (age at any time during the study period: 1January 1998–31 December 2008).Patients with a history of Prostate Cancer, as well as men with a history of CVD were excluded for longitudinal analysis. Access to the patient’s medical history was a prerequisite. We collected the following data from the registration: date of birth, date of entry in the study, date of and reason for leaving the registration, GP code, patient contacts (ICPC codes), prescriptions (ATC codes) and ICPC codes attached to these medications, and hospital referrals. From this information, we calculated age from date of birth and number of person years in study. All of the data were anonymised. We received ethical permission to access the Registration Network Groningen from the Medical Ethical Research Board, University Medical Center Groningen (M13.132482).Cardiovascular event is defined as a documented acute myocardial infarction, chronic ischaemic heart disease, transient ischemic attack or stroke in the medical records of the GPs . Table 1Lower urinary tract symptoms (LUTS) include the sensation of not urinating completely, withholding urinating, and difficulty voiding. This may include having a stop-and-go urinary flow and getting up frequently at night to urinate . There iErectile dysfunction is defined as the persistent inability to achieve or maintain an erection suficient for satisfactory sexual performance , in thisDescriptive statistics were used to compare the baseline characteristics. Continuous variables are presented in means and confidence intervals, and nominal variables reported in modi. Prevalences (and incidence numbers for LUTS and CVD) for LUTS, CVD, hypertension, Erectile dysfunction, and Diabetes mellitus were calculated by means of prevalence/1000 personyears for the first and last studyyear (1998 and 2008).Cox proportional hazard regression analysis was used to determine the association between the proportions of CVD (outcome) and LUTS in our population. We used age as time factor in a Cox proportional hazard regression analysis to determine the association between CVD (outcome) and LUTS in our population. In this open cohort, time to event (CVD) was calculated from the patient’s date of birth to event (i.e. diagnosis of LUTS) and data of subjects were censored in case leaving the cohort, death, or end of study.Initially, an unadjusted analysis was performed. This association regression model was subsequently corrected for confounders. A covariate was considered a confounder in the event the beta coefficient of LUTS changed by 10 or more percent .In this analysis, the following potential confounders  were tesIn all analyses, multiple dummy variables were created for the analyses of categorical data. In addition, model assumptions were tested for compliance. All analyses were performed using SPSS version 16 based on a two-sided 0.05 significance level.From the database, we collected data from 6614 eligible men. Of these, 1165 (17.6%) reported LUTS during the study period. Table From all men, 1539 had cardiovascular diseases during the study period: 23.2% (95% CI 22.2-24.2%) . The incidence numbers for CVD were 12.9/1000 py in 1998 and 10.4/1000 py in 2008.Table There is a significant unadjusted univariate relationship between the proportions cardiovascular events and LUTS . However, the hazard ratio of LUTS for cardiovascular events, compared to no LUTS, in the adjusted multivariate model, was non-significant . Therefore, men with LUTS were not more likely to develop CVD, at any time, than men without LUTS. In our analysis, hypertension, ACE-inhibitors, All antagonists, beta blockers, and calcium antagonists were found to confound this relationship. In this population, cardiovascular disease was not associated with LUTS (Table Nowadays, when a patient consults his general practitioner for LUTS, it is not customary to search for comorbidities such as CVD. However Dynamic cohort and clinical studies suggest a correlation between LUTS and CVD. To the best of our knowledge this is the first study that has been done to establish the relationship between LUTS and CVD in a primary care population. This is an important population because the GP plays an increasingly prominent role with respect to the early detection of patients at risk, for example, for cardiovascular diseases ,4. In thOther studies report the association between LUTS and CVD ,18,36. TRecently, a community-based retrospective, cross-sectional and longitudinal study into this relationship was carried out . WehrberThe strength of our study is the longitudinal analysis, and the large cohort of 6615 men. Using data from electronic medical records from general practices makes it possible to compare the prevalence of CVD between different patient groups in the same study population. We also had information about the sequence of the diseases, LUTS and CVD events, and hence a possible relationship between LUTS and subsequent CVD could be taken into the statistical analyses.According to previous articles about RNG data, this study population is representative for the Dutch population .A limitation of this study is having access to large patient populations in a health registry with data that are not collected in a structural way. Patients are not measured periodically. Important confounders, such as dyslipidemia, smoking and obesity, are not always registered in the electronical medical records of GPs. For example, in this study dyslipidemia is more frequent in men with LUTS. However, the prevalence of dyslipidemia in all subjects is lower than the Dutch prevalence numbers . BecauseBecause of the absence of an ICPC code for LUTS, we assumed the presence of LUTS if one or more of these ICPC codes were identified Table . UnfortuAnother factor that should be considered is the underreporting of LUTS, the so-called iceberg phenomenon: ,40,41 i.To the best of our knowledge, this is the first study that investigated the relationship between LUTS and cardiovascular diseases in primary care. It suggests that a GP does not have to complete a full cardiac risk profile in men who present with LUTS symptoms for the first time.LUTS: Lower urinary tract symptoms; CVD: Cardiovascular disease; RNG: Registration network groningen; GP: General practitioner; ED: Erectile dysfunction; ICPC: International classification of primary care; ATC: Anatomical therapeutic chemical; PY: Personyears; CI: Confidential interval; HR: Hazard ratio; OR: Odds ratio.The authors declare that they have no competing interests.IIB: carried out the analysis and interpretation of data, and drafted the manuscript. BJK caried out the analysis and performed the statistical analysis, and helped to draft the manuscript. WK, KM, and JMN participated in the study design and coordination. WK helped to draft the manuscript and made contributions to conception and design of the study. All authors read and approved the final manuscript.The pre-publication history for this paper can be accessed here:http://www.biomedcentral.com/1471-2296/15/9/prepub


N-methyl-D-aspartate antagonist reproduces symptoms of schizophrenia, we sought to determine whether the rubber-hand illusion is augmented by ketamine.Our sense of body ownership is profound and familiar, yet it may be misleading. In the rubber-hand illusion, synchronous tactile and visual stimulation lead to the experience that a rubber hand is actually one's own. This illusion is stronger in schizophrenia. Given the evidence that ketamine, a noncompetitive We studied 15 healthy volunteers in a within-subjects placebo-controlled study. All volunteers carried out two versions of the rubber-hand task, each under both placebo and ketamine infusions. In one task, they saw a rubber hand being stroked in synchrony with tactile stimulation of their real, hidden hand. In the other, stroking of the real and rubber hands was asynchronous. We recorded subjective changes in sense of ownership, as well as participants' ability to localize their hidden hand.Ketamine was associated with significant increases in subjective measures of the illusion and in hand mislocalization. Although asynchronous visuotactile stimulation attenuates the strength of the illusion during both placebo and ketamine, there remained a significant illusory effect during asynchronous visuotactile stimulation under ketamine compared with placebo. The strength of the illusion during asynchronous visuotactile stimulation correlated with other subjective effects of the drug.Ketamine mimics the perturbed sense of body ownership seen in schizophrenia, suggesting that it produces a comparable alteration in integration of information across sensory domains and in the subjective and behavioral consequences of such integration. Our sense of body ownership, the feeling that our body parts belong to us, although profound, is fragile. As a consequence of brain injury or mental illness, a patient may no longer feel ownership of a body part. Conversely, a person can sometimes feel that an external object is part of his or her body. Such is the case in the rubber-hand illusion (RHI) , in whicThe illusion usually involves a temporally precise combination of visual and tactile stimulation, although a comparable illusion can occur without visual input . SynchroThus, the RHI offers a way of exploring our sense of body ownership. A disturbance in this sense may be an important feature of schizophrenia . Indeed,N-methyl-D-aspartate receptor antagonist, to healthy volunteers produces symptoms like those seen in schizophrenia  right-handed, healthy volunteers, with a mean age of 22.4 years, were recruited through local advertisement. The research was approved by Addenbrooke's National Health Service Trust Research Ethics Committee. All participants spoke fluent English, were nonsmokers at the time of testing, and had no history of clinical drug or alcohol abuse or of psychiatric illness. One participant was later excluded because of a subsequently discovered history of psychiatric illness, and two more were unable to complete the study because of nausea.The task took place at 200 ng/mL blood plasma level of ketamine, after a series of cognitive tasks run at a lower level, 100 ng/mL plasma (the data for which will be reported elsewhere). The participant's right hand rested in an open-side box see , the lifTwo versions of the task were administered during a session, each lasting 5 min; in the Synchronous condition, the visible brush stroking the rubber hand rotated with the same direction, phase, and frequency as the invisible brush stroking their real hand. In the Asynchronous condition, brushes rotated in opposite directions at differing frequencies. The order of the two versions of the task was counterbalanced across sessions and participants.Bilateral intravenous catheters were inserted into the forearms: one for ketamine or placebo infusion and the other to enable serial blood sampling. In total, eight samples were taken. Racemic ketamine (2 mg/mL solutions) or saline solution were administered by a computerized target-controlled infusion (TCI) system, which calculates the infusion rates required to achieve the “target” blood concentration set by the user. Our TCI setup consists of an infusion pump , under the control of a laptop personal computer running the software Stanpump . StanpumFollowing visuotactile stimulation, participants rated their subjective experience of the illusion using a 5-point Likert scale, in response to nine standard questions  1)..1).We explored the impact of the illusion on participants' estimated position of their right index finger see . A rulerSubjective experiences induced by the drug were recorded using a series of clinician-administered questionnaires carried out by a psychiatrist. We used the Clinician-Administered Dissociative States Scales (CADSS) , Britishet al. Delusions Inventory  and task (synchronous vs. asynchronous stroking) on overall experience of the illusion, represented by an average response, of each participant, to all questions (RHI index). Post hoc paired-samples In addition, given previous work suggesting that the first three questions in the scale are most pertinent to the illusion , we carrt tests were used in an attempt to discern further the individual impact of drug and task on this measure.The change in participant's ability to localize their right index finger, represented by “drift,” was assessed using 2 × 2 repeated-measures ANOVA. Post hoc paired-samples Finally, post hoc correlations between measures of the illusion and background personality measures as well as symptomatic effects of ketamine were carried out.The mean plasma ketamine level at the time of task was 258.3 ng/mL (SD = 88.56).The RHI index , as well as the mean response to each question (for completeness), are reported in F = 9.2, p = .01, and of task, F = 15.9, p = .002, was found, with the illusion proving stronger under ketamine and when stroking was synchronous. No interaction was found, F = .12, p = .74.The effects of drug and task on RHI index  were investigated using 2 × 2 repeated-measures ANOVA with order of drug administration modeled as a between-subject effect. The two factors were drug (ketamine vs. placebo) and task (synchronous vs. asynchronous stroking). A significant main effect of drug, t tests revealed that the RHI was greater under ketamine for both synchronous (p = .016) and asynchronous (p = .019) stroking assessed separately. Moreover, synchronous stroking was associated with a greater RHI under both placebo (p < . 001) and ketamine (p = .03).In a more detailed analysis of the overall pattern of findings, paired-samples F = 7, p = .02, and of task, F = 25.3, p < .001, with the illusion proving stronger under ketamine and when stroking was synchronous. No interaction was found, F = 1.2, p = .3. Subsequent paired-samples t tests revealed that the RHI was greater under ketamine for both synchronous  and asynchronous  stroking assessed separately. Moreover, synchronous stroking was associated with a greater RHI under both placebo (p = . 000) and ketamine (p = .005).In the subsequent subanalysis focusing only on mean responses to questions 1 through 3 (as described earlier), again including drug order as a between-subject factor, we observed a significant main effect of drug, In short, using both the overall subjective measure (questions 1–9) and focusing on more specific indexes (questions 1–3), there is a significant impact of ketamine during both synchronous and asynchronous stroking conditions and a significant impact of synchrony of stroking for both placebo and ketamine.p = .771, paired t test. We can therefore be confident that drug effects were due to its impact on the illusion rather than on a general inaccuracy in hand localization. The total drift in position after 5 min, relative to each participant's estimated starting position, was entered into a 2 × 2 (drug × task) ANOVA.For each participant, an initial judgment, before tactile stimulation or viewing of rubber hand, was made to assess baseline judgment of hand position. The difference in this baseline judgment compared with an objective judgment made by the experimenter was assessed across placebo and ketamine sessions. The root mean square error on placebo  and on ketamine  did not differ significantly across the ketamine and placebo conditions, F = 5, p = .042, but not of synchrony, p = .792. Although under ketamine, the amount of subjective drift was numerically greater for the synchronous  than asynchronous  condition, this was not a significant difference (p = .587). Under placebo, drift under asynchronous conditions  was greater than synchronous (mean = 2.3 cm SEM = 1.3 cm), but did not differ significantly (p = .67). Finally, for synchronous conditions, drift was greater under ketamine (p = .039), although the same was not true for asynchronous conditions (p = .236). No interaction was found (p = .489). Results are shown in There was a significant main effect of drug on final (participant-estimated) hand position, r = .59, p = .02, two-tailed, and for the more constrained measure (questions 1–3), Pearson r = .58, p = .023, two-tailed  and CADSS  relatingiled see .Healthy participants were significantly more vulnerable under ketamine to a false sense of limb ownership. The enhanced illusory effects produced by ketamine were manifest both in subjective experiences and in a greater tendency to localize the position of their real hand to that of the visible rubber hand. Note that, before the onset of visuotactile stimulation, participants were equally accurate at localizing their hidden hand under ketamine as they were under placebo. Thus, the effect of ketamine emerges following the onset of visuotactile stimulation. During ketamine administration, the experience of the illusion was enhanced even when visual and tactile inputs were asynchronous.Under normal circumstances, body representation, and hence, presumably, sense of ownership, depend in part on multisensory integration. It has been argued that the sense of ownership inherent to the RHI is contingent on simultaneity of visual, tactile, and proprioceptive inputs , a conclThe impact of ketamine on sense of body ownership was intriguing and in keeping with our predictions based on observations in schizophrenia. We suggest that the boost in sense of ownership is not related to an enhancement of bottom-up integrated sensory processes. If it were, the drug's effects would presumably be unique to, or greater in, the synchronous condition. Rather, it appears that the drug produces an increased tendency to accept ownership of the hand, even in the face of contradictory sensory information. One interpretation of this might simply be a nonspecific dissociative effect. However, this is unlikely given that the enhancing effect of visuotactile synchrony was preserved under ketamine. Thus, it would be wrong to conclude that ketamine attenuates sensitivity to the coherence of visuotactile information. Rather, ketamine produces an overall increase in tendency to the illusion, whereas this sensitivity to the coherence of visual and tactile input is preserved. What might be the neural mechanisms for this observed combination of effects? Two prior sets of electroencephalogram observations are relevant: first, gamma-band oscillations are enhanced in association with acute ketamine administration . Second,A further possibility is that ketamine enhances the salience of the visual input (the sight of a rubber hand in a position compatible with one's own hand) at the expense of information arising from the temporal asynchrony between sensory inputs. This would explain why the presence of the rubber hand is enough to enable the illusion to persist, albeit in attenuated form, when visuotactile inputs are asynchronous. Under normal circumstances, the appearance and orientation of the rubber hand is crucial in eliciting the illusion, and nonhand objects will not produce a sense of ownership even when stroking is synchronous, nor will a convincing facsimile of a hand produce the illusion when its orientation is clearly inconsistent with the real hand . Under kHow do these findings relate to schizophrenia? Peled and colleagues produced evidence that the RHI is acquired more rapidly and profoundly in people with schizophrenia than healthy control subjects , althougIt is also noteworthy that previous work with ketamine has produced a perturbation in right prefrontal sensitivity to mismatch signals , a patteIn summary, we have combined a pharmacologic model of schizophrenia with an illusion to which patients with schizophrenia have been shown to be especially sensitive. With ketamine, just as with schizophrenia, the illusion is enhanced. Although the precise nature of ketamine's impact is speculative, two features of its effects are noteworthy. First, it promotes an overall increase in the subjective and behavioral indexes of the illusion. Second, this effect is found even when a sensory asynchrony is present, a manipulation that would normally obliterate the illusion. Perhaps this pattern may be understood in terms of the drug's impact on the top-down–bottom-up balance that would normally account for the illusion's characteristic features.


We wished to compare the San Francisco Syncope Rule (SFSR), Evaluation of Guidelines in Syncope Study (EGSYS) and the Osservatorio Epidemiologico sulla Sincope nel Lazio (OESIL) risk scores and to assess their efficacy in recognising patients with syncope at high risk for short-term adverse events . We also wanted to test those variables to designate a local risk score, the Anatolian Syncope Rule (ASR).This prospective, cohort study was conducted at the emergency department of a tertiary care centre. Between December 1 2009 and December 31 2010, we prospectively collected data on patients of ages 18 and over who presented to the emergency department with syncope.We enrolled 231 patients to the study. A univariate analysis found 23 variables that predicted syncope with adverse events. Dyspnoea, orthostatic hypotension, precipitating cause of syncope, age over 58 years, congestive heart failure, and electrocardiogram abnormality (termed DO-PACE) were found to predict short-term serious outcomes by logistic regression analysis and these were used to compose the ASR. The sensitivity of ASR, OESIL, EGSYS and SFSR for mortality were 100% (0.66 to 1.00); 90% (0.54 to 0.99), 80% (0.44 to 0.97) and 100% (0.66 to 1.00), respectively. The specificity of ASR, OESIL, EGSYS and SFSR for mortality were 78% (0.72 to 0.83); 76% (0.70 to 0.82); 80% (0.74 to 0.85) and 70% (0.63 to 0.76). The sensitivity of ASR, OESIL, EGSYS and SFSR for any adverse event were 97% (0.85 to 1.00); 70% (0.52 to 0.82); 56% (0.40 to 0.72) and 87% (0.72 to 0.95). The specificity of ASR, OESIL, EGSYS and SFSR for any adverse event were 72% (0.64 to 0.78); 82% (0.76 to 0.87); 84% (0.78 to 0.89); 78% (0.71 to 0.83), respectively.The newly proposed ASR appears to be highly sensitive for identifying patients at risk for short-term serious outcomes, with scores at least as good as those provided by existing diagnostic rules, and it is easier to perform at the bedside within the Turkish population. If prospectively validated, it may offer a tool to aid physicians' decision-making. Syncope is a symptom of cerebral hypoperfusion and is defined as a short, sudden, self-terminating episode of transient loss of consciousness with a failure to maintain postural tone and with a variety of aetiologies. More than two million people are evaluated for syncope each year in the US, with the cost running into the billions of dollars . It accoPatients with syncope pose a difficult diagnostic dilemma. Most patients appear well and are asymptomatic on arrival, and there are often no witnesses to the event. Some patients will require emergency hospitalisation for a workup and for the treatment of life-threatening or potentially life-threatening causes ,5. Even Clinical decision rules are tools designed to assist clinicians in making decisions at the bedside and to risk-stratify the patients. They are derived from original research and incorporate important predictors of outcome from the history, physical examination and basic diagnostic tests. The Osservatorio Epidemiologico sulla Sincope nel Lazio (OESIL) risk score, the San Francisco Syncope Rule (SFSR) and the Evaluation of Guidelines in Syncope Study (EGSYS) risk scores are largely used in emergency settings -9.According to the OESIL study group, syncope score predictors of death after one year are an abnormal electrocardiogram (ECG), a history of cardiovascular disease (including heart failure), age of over 65 years old, and syncope without prodrome. Each risk factor counts as one point. In keeping with the OESIL study, we considered those patients who were characterised by a score of up to 1 to be low risk. Patients with a score of 2 or higher were assumed to be at intermediate or high risk  . The SFSThe aim of the present study was to compare the SFSR, the OESIL risk score and the EGSYS risk score to assess their efficacy in recognising patients with syncope at high risk for short-term (within seven days) adverse events  and to test those variables to designate a local risk score, termed the Anatolian Syncope Rule (ASR). To the best of our knowledge, no comparative study for the Turkish population regarding the use of three clinical decision rules for syncope in a 7-day period and no proposed new decision rule have previously been published.This observational, single-centre, prospective cohort study was conducted in the ED of a research and training hospital. During a 13-month period, between December 1 2009 and December 31 2010, we prospectively collected data on patients who presented to the ED with syncope.Patients were screened at triage on the basis of the following symptoms, findings and complaints for a potential syncopal event: syncope, loss of consciousness, presyncope, fainting, collapse, light-headedness, dizziness, falls, seizures, head injury and bone fractures. A research paramedic, who was blinded to the study protocol, ensured enrolment of all potential syncopal events. A patient evaluation form with the name of the patient written on it was brought to one of the five research physicians. The exclusion criteria used to determine the target population were as follows:1. age below18;2. unable to give either written or verbal informed consent;3. confirmed nonsyncopal syndromes such as vertigo, coma, shock, witnessed seizure, sustained unconsciousness, head injury preceding loss of consciousness, stroke;4. unable or unfeasible to follow-up ;5. presence of a clinical condition that required admission, such as acute myocardial infarction, pulmonary embolism, intracranial haemorrhage, sustained symptomatic bradycardia and tachycardia;6. comorbidities with low survival rate;7. pregnancy;8. history of drug or alcohol abuse.The final decision to enrol the patient was made by the research physicians and they completed the patient evaluation forms, containing approximately 70 variables that were derived from existing syncope rules, patient history, vital signs and physical examination. Where possible, attending physicians independently evaluated patients to measure agreement on subjective variables requiring interpretation. Only the attending physician assessments were used in the analysis and derivation of the model. All patients were treated in the usual manner by the attending physicians, who were blinded to the study protocol.As shown in Figure The primary end-point of our study was the presence of any adverse event within seven days of discharge from the hospital, which included rehospitalisation, a major therapeutic intervention or death. No exceptions were made for these criteria. All discharged and admitted patients or their tutors were surveyed on seventh day of the index event by follow-up examination or interview and the presence of any adverse event was sought.A major therapeutic intervention was defined as cardiopulmonary resuscitation, pacemaker or implantable cardioverter-defibrillator insertion, intensive care unit admittance or acute antiarrhythmic therapy. An ECG was defined as abnormal in the presence of any of the following in keeping with the criteria used in OESIL risk score: atrial fibrillation or tachycardia; a sinus pause of two seconds or more; sinus bradycardia with a heart rate ranging between 35 and 45 beats per minute; conduction disorders; ECG signs of previous myocardial infarction or ventricular hypertrophy; or multiple premature ventricular beats . The ECGThe study was approved by the Ethical Committee on Human Research of the Coordinating Centre, and participants provided written consent.t-test was used for continuous variables. To develop a score predictive of high-risk patients, binary variables were created from continuous variables. Univariate predictors of high-risk syncope were identified by chi-squared analysis. Variables with a high univariate correlation and with P < 0.10 were considered in stepwise logistic regression analyses with a cut-off P-value of < 0.05. Variables with similar meaning were combined to reduce the number of variables to be included in the logistic regression analysis  was used for descriptive data analysis. All continuous variables are presented as means with standard deviation (SD). Categorical variables are presented as percentages with 95% confidence intervals (CIs). A Pearson's chi-squared or Fisher's exact test was used for categorical variables, and a Student's Multivariable logistic regression analysis was then used to model the independent associations, demographic factors and whether any adverse event had happened, while controlling for sex and age as possible confounders. The goodness-of-fit for this regression model at each variable was verified by the Hosmer-Lemeshow test. The proportion of variance explained by the final model was determined using the Nagelkerke R statistics. The results of the multivariable logistic regression analysis were used to develop a risk score to predict the probability of any adverse event in seven days, named the ASR Table . The relWe compared the prognostic accuracy of ASR and other risk scores by generating ROC curves and comparing the AUC using the method described by Hanley and McNeil for comparing ROC curves derived from the same cases . When assessing potential decision thresholds in each respective risk score, our goal was to achieve near 100% sensitivity for need for intervention or death at the highest possible specificity. According to this criterion, we dichotomized ASR as high- and low-risk values from the decided thresholds for any adverse event and for mortality. Then, we calculated sensitivity, specificity, positive predictive value, negative predictive value, positive likelihood ratio and negative likelihood ratio values for any adverse events for each syncope criterion, including the criteria that we had defined (ASR). To determine the concordance of high- and low-risk criteria with the presence of any adverse event, kappa values for each criterion were calculated.P < 0.05 was considered to be statistically significant for all analyses.A two-tailed Six parameters  were found to have significant predictive value according to logistic regression analysis. They were represented by a simple mnemonic, DO-PACE, to compose the newly proposed ASR Table . PrecipiDespite the limitations of our study, to the best of our knowledge this is the first reported cohort study of syncope patients to compare all three clinical rules and the only attempt to propose a clinical decision rule to identify high-risk patients from the Turkish population. The predictor variables used in the evaluation of patients with syncope developed a highly sensitive clinical decision rule that we believe will augment physician judgement and allow physicians to decide rationally which patients with syncope need admission according to their risk for short-term serious outcomes. The rule is not complex and easily remembered by a simple mnemonic: DO-PACE .The demographic variables of age, sex, and race are potential risk factors for cardiovascular disease. Epidemiologic and cohort studies have confirmed the importance of age; however, of course, age alone is a marker for increased mortality. Although increasing age is accompanied by an increased risk of poor outcomes, there is no single cut-off for age but rather a continuum of gradually increasing risk -13. The Orthostatic hypotension may identify some patients with syncope related to volume depletion, autonomic insufficiency or medications. Orthostatic hypotension is common in patients with syncope of unknown aetiology, and this finding is also present in up to 40% of asymptomatic patients older than 70 years old and 23% of those patients younger than 60 years old . A diagnet al. showed that, in patients with syncope, ST-segment and T-wave abnormalities were one of the significant predictors of arrhythmias [Sarasin hythmias . In anothythmias . Arrhythhythmias . In our In a prospective study in Europe of 676 patients with syncope, predictors of short-term unfavourable outcome at 10 days included a mean age of 59 years old, abnormal ECG findings, concomitant trauma, an absence of prodromal symptoms and male sex. The occurrence of long-term unfavourable outcome was correlated with an age older than 65 years, a history of neoplasms, cerebrovascular disease, structural heart diseases and ventricular arrhythmias. These investigators concluded that risk factors for short- and long-term adverse outcomes after syncope were different and that long-term mortality was related to the comorbidities. Significant independent prognostic factors for time to mortality were diabetes, coronary artery bypass graft surgery, a history of malignancy, narcotics use, smoking, volume depletion and atrial fibrillation . In our et al. defined multiple univariate predictors of serious outcomes that included an abnormal ECG, a haematocrit of less than 30%, a complaint of shortness of breath, hypotension and a prior history of congestive heart failure [A recent study by Quinn  failure ,12. In oAn absent or brief prodrome (less than 5 seconds) may be present with dysrhythmias, whereas neurally mediated syncope  may be characterised by longer prodromes and associated with nausea or vomiting .When developing any prediction rule, one can always achieve 100% sensitivity but often at the expense of specificity and by overfitting of the statistical model. For example, in our model, we could have achieved 100% sensitivity by adding an age older than 40 years old to the rule, which would have identified the seven patients not predicted by the rule. We thought that the trade-off was suboptimal because the absolute admission rate for patients deemed to be high risk would be the same as, if not slightly higher than, at baseline. We were also aware that rare, serious outcomes not in our derivation set would make validating any rule with 100% sensitivity and certainty almost impossible. We thus accepted a rule that maximised sensitivity and specificity at the expense of not achieving 100% sensitivity. We purposely did not develop the rule with the physician's decision-to-admit as the primary outcome. Thus, this rule is not a decision rule to predict admission. The reasons to admit often take other factors  into consideration that, although important, we believed were not specific to whether patients with syncope are at an acute risk for serious outcomes that require acute hospitalisation.Our study has some limitations. First of all, power calculations were not made prior to commencement of the study because there were no published articles to guide our three-way comparison. This was a single-centre study, thus, our patient group may not represent the whole Turkish population. Attending physicians reviewed all cases independently and the decisions of attending physicians superseded in case of dispute. However, inter-rater reliability of all subjective findings was reasonable, fair or worse with kappa values of 0.5 or less. So we decided to accept attending senior physicians' decisions for the analysis and derivation of the model.This prospective cohort study has shown that many variables included here, but not described before, are associated with serious outcomes in patients with syncope. SFSR performed better in the prediction process of any short-term adverse events. SFSR also had the best score to rule-out short-term mortality. However, none of these scores either rule-in or rule-out any short-term adverse events or mortality well enough to be suggested. The newly proposed ASR appears to be highly sensitive for identifying patients at risk for short-term serious outcomes, at least as good as other scores, and it is easier to perform at the bedside for the Turkish population. If prospectively validated, it may offer a tool to aid physicians' decision-making. Our proposed decision rule now needs to be validated prospectively in a large cohort of patients before physicians can consider it in clinical decision-making.The authors declare that they have no competing interests.KK made substantial contributions to the conception and design, acquisition of data and interpretation of data. HA made substantial contributions to the conception, design, and analysis and interpretation of data, and was involved in drafting the manuscript. OL made substantial contributions to the design, the analysis and interpretation of data, and was involved in drafting the manuscript. AOE made substantial contributions to the conception and design, and the analysis and interpretation of data. OY contributed to the analysis and interpretation of data, and was involved in drafting the manuscript. SB made substantial contributions to the conception and design, and analysis and interpretation of data. NVB made substantial contribution to the conception and interpretation of the data and was involved in drafting the manuscript. EEU made substantial contributions to the conception and design, and analysis and interpretation of data. All authors read and approved the final version of this manuscript.