Neuropathic pain (NP) can cause substantial suffering and, therefore, it must be diagnosed and treated promptly. Diagnosis of NP can be difficult and if made by an expert pain physician is considered the gold standard, however where expert help may not be easily available, screening tools for NP can be used. The painDETECT questionnaire (PD-Q) is a simple screening tool and has been widely used in several languages. We developed an Arabic version of PD-Q and tested its validity and reliability.The original PD-Q was translated into the Arabic language by a team of experts. The translated version of the PD-Q was administered to the study population, which included patients having moderate to severe pain for at least three months. Reliability of the Arabic version was evaluated by an intra-class-correlation coefficient (ICC) between pre- and post-measures and Cronbach’s α values. Validity was measured by receiver operating characteristic (ROC) curve. Expert pain physician diagnosis was considered as the gold standard for comparing the diagnostic accuracy.A total of 375 patients were included in the study, of which 153 (40.8%) patients were diagnosed with NP and 222 [59.2%] patients had nociceptive pain. The ICC between pre- and post-PD-Q scale total scores for the overall sample, NP group, and NocP group was 0.970 , 0.963 , and 0.962 , respectively. The Cronbach’s α values for the post-assessment measures in the overall sample, NP group, and nociceptive pain group, were 0.764, 0.684, and 0.746, respectively. The area under the ROC curve was 0.775  for the PD-Q total score.The Arabic version of the PD-Q showed good reliability and validity in the detection of NP component in patients with chronic pain. The International Association for the Study of Pain (IASP) defines pain as “an unpleasant sensory and emotional experience associated with actual or potential tissue damage or described regarding such damage” . The twoMoreover, NP does not respond to conventional analgesics such as paracetamol and non-steroidal anti-inflammatory drugs (NSAIDs). Instead, NP is treated with certain other agents such as tricyclic antidepressants , serotonin-norepinephrine reuptake inhibitors , and antiepileptics  . DiagnosReaching the precise diagnosis of NP may be difficult especially in the primary care setting. Based on the distinct features of NP that distinguish it from NocP, several screening tools have been developed to help in the diagnosis . While eThe PD-Q was first used in the German population in the year 2004. Since then, it has become very popular and has been translated into several languages including Spanish, Swedish, Dutch, Turkish, Japanese, and Korean –21. MorePatient access to care is different rural areas as compared to urban areas and different for different socioeconomic statuses. Lack of trained personnel is also a burning issue. To date there are no, regional guidelines for NeuP management, which are largely dependent on the local settings. To raise awareness among healthcare practitioners about NeuP, and to provide them a simple tool for its diagnosis to alleviate the suffering of the patients afflicted with chronic pain due to neuropathological causes, in the Kingdom of Saudi Arabia, especially in the remote areas where expert medical help is lacking. To address this unmet need, we developed an Arabic version of PD-Q and tested its validity and reliability.Adult patients attending any of the three clinics: pain management e diabetes and orthopedics at a single tertiary hospital in Riyadh, King Fahad Medical City (KFMC) were included in the study. Patients with moderate to severe NP or NocP pain  for a minimum of three months were included. Patients with the pain of unknown origin, mixed-type pain, and diffuse pain which includes fibromyalgia syndrome, myofascial pain, complex regional pain syndrome, cancer pain, and headaches wereexcluded. Patients with substance abuse, chronic alcoholism, severe depression, and those incapable of understanding the questionnaire were excluded .The study was approved by the Institutional Review Board at KFMC, Riyadh, Saudi Arabia. Written informed consent was obtained from the participants who met inclusion criteria and agreed to participate in the study.The study participants were randomly approached and invited to participate in this study by the research team over a period of one year from March 2016 to March 2017.The adaptation procedure was monitored by an expert panel including two specialists in pain management, an expert in research methodology, an expert in clinical research, and an expert in linguistics. The international guidelines for cross-cultural adaptation of health questionnaires and diagnostic tests were followed. An internationally-accepted translation methodology was used by a well-established linguistic validation process , 24.Two bilingual translators, with Arabic mother tongue, translated the questionnaire to the Arabic language. While one of the translators was informed about the questionnaire and its applications, the other had no information about the questionnaire. Each translator independently produced a translated version. The two versions were then combined into a single questionnaire. This was followed by backward translation to English by two native English speakers who had no knowledge of the original version. A meeting was then held between the experts and the two translator teams to discuss the original and translated versions. The expert committee consolidated all the versions and developed the pre-final version for pre-testing. The pre-final version was administered to a sample of 30 patients as a part of pre-testing. Each patient was asked to fill-in the form and was then interviewed about his/her understanding of each item of the questionnaire and the corresponding response.The study population was divided into two groups. The first group included patients diagnosed with NP by a pain specialist in pain clinics as per the guidelines established by the IASP . The secThe same investigator administered the Arabic version of the PD-Q to the study population twice (pre-clinic visit and post clinic visit) within a gap of two to four hours .The PD-Q is simple, self-administered, useful screening questionnaire that allows detecting NP component in patients with chronic pain. The PD-Q consists of four main sections. The first section contains three items with 11-point Likert scale format with anchor terms in the scale ends , accompanied by a color grading scale representing pain intensity in analog format. These items assess the intensity of pain at the time of administration of the scale, and maximum pain intensity and average pain during preceding 4 weeks. The first section is used to diagnose the presence of pain and is not included in the questionnaire scoring. In the second section, patients are asked to mark one of the four graphs that best describe their pain course patterns. The possible patterns and their scores are determined as follows: persistent pain with slight fluctuations (0 points), persistent pain with pain attacks (−1 point), pain attacks without pain between them (1 point), and pain attacks with pain between them (1 point). The third section includes a sensory map representing homunculus along with questions designed to mark the pain zone, a dichotomous item about the presence of radiating pain and showing the direction of radiating pain with an arrow. The positive answer about the presence of radiating pain is scored with two points. In the last section, there are seven Likert type items asking about the intensity of the sensation marked over the homunculus. These items are scored with a 6-point Likert format, with corresponding ordinal anchor terms . These Likert-type items enquire about the following sensations: burning, tingling or prickling, allodynia, pain attacks, temperature evoked pain, numbness, and pressure-evoked pain. This last section provides scores between 0 and 35 points. The final score is obtained summing up the scores of the last three sections with a total score of −1 to 38. Two cutoff values are used by the developer of PD-Q for the presence of NP. Scores ≤12 state that an NP component is unlikely and scores ≤19 indicate that neuropathic component is very likely to be present. Scores between 12 and 19 suggest that the result is unclear .The sample size was calculated assuming a significance level of α = 0.05 and prevalence of 41% with power = 0.95 and maximum difference = 0.5. The sample size that is needed to validate the questionnaire for the two groups was 376 patients.Socio-demographic and clinical characteristics of patients were summarized for the whole sample (NP group and NocP group) using frequency (%) for categorical variables and mean  for continuous variables. For continuous variables, normality was assessed using Shapiro Wilk test and histograms. The socio-demographic and clinical characteristics were compared between the two groups using a chi-square test for categorical variables and t-test or Mann Whitney’s U test for continuous variables.The PD-Q responses were collected in two sessions, one pre-clinic visit and the other post- clinic visit. For the test re-test reliability of the items of PD-Q and its total scale score, intra-class-correlation coefficient (ICC) between pre- and post-measures were computed. For the pre- and post- clinic visits, correlation of the items of PD-Q with the total scale scores were assessed using Spearman correlation coefficients. Cronbach’s α for the pre- and post-measures of the items of PD-Q were computed to examine the internal consistency. All the above analyses were performed for the whole sample and also separately for the two study groups.Initially, median (interquartile range) of the total scores for PD-Q were computed and Mann Whitney’s U test was used to compare the scores between NP and NocP groups. Thereafter, to assess the discriminant validity of PD-Q in identifying the NP, we used receiver operating characteristics curve (ROC) analyses. A binary variable with a physician diagnosis of (NP = 1) versus (NocP = 0) was created. Physician diagnosis of neuropathic pain was carried out before the questionnaire was completed by the participants and the physicians were blinded to the responses on the questionnaires. In the logistic regression analysis, we used the physician diagnosis of neuropathic pain (a binary dependent variable) to assess how the total score of PD-Q scale predicts the physician diagnosis. This analysis assessed the validity of PD-Q by establishinga corroboration between physician diagnosis and questionnaire diagnosis for. Next, ith th binary variable, a ROC curve analyss was conducted for the total scale scores of PD-Q. concerning for each cut-off-point of the scores, sensitivity, specificity, correctly classified, positive likelihood ratio, negative likelihood ratio, was also calculated. In this analysis, using Youden’ s index , the besA statistical significance level of p<0.05 was used to reject the null hypothesis. The analyses were conducted using statistical software of SPSS software version 17.0  and Stata version 12 .A total of 375 patients were included in the study. Expert pain physicians diagnosed NP in 153 (40.8%) patients and NocP in 222 (59.2%) patients. The mean age of patients in the NP and NocP groups was similar; however, the range was broader in patients with NocP . Use of medication, diabetes, and heart disease were significantly more common in patients with NP. The duration of pain was more in patients with NP .Paracetamol was the most common painkiller used in both NP  and NocP  groups. Paracetamol was followed by meloxicam , and diclofenac sodium  in both the groups.NP was most commonly seen in patients with low back pain/referred pain (74 [48.4%]), and NocP was most commonly seen in patients with osteoarthritis (111 [50.0%]) .There was a high level of test-re-test reliability for the items of the PD-Q scale in the overall sample and the two study groups. The ICC between pre- and post-PD-Q scale total scores for the overall sample, NP group, and NocP group was 0.970 , 0.963 , and 0.962 , respectively. For the measures of the scale in the post clinic visit, the Cronbach’s α values were 0.764, 0.684, and 0.746, in the overall sample, in NP group, and in the NocP group, respectively. For the pre-clinic visit measures of the items of the PD-Q scale, a similar level of internal consistency was observed .The total scores for both pre-clinic visit and post clinic visit on the fourth section of the PD-Q scale, which includes seven items were evaluated using the Spearman rank correlation co-efficient, a significant (moderate to high) correlation between the total score and each item of the scale was observed .Considering the diagnosis as NP or NocP by a pain physician as the gold standard, when the discriminant validity was examined using the receiving operating characteristics (ROC) curve analyses, the area under the ROC curve was estimated to be 0.775  for the PD-Q total score .P<0.001)  . The finIn this study, the most common agent used for pain, irrespective of the pain type, was paracetamol followed by NSAIDs. This may mean that patients with NP either had no pain relief or had only partial relief, as they were not receiving appropriate treatment. If the diagnosis of NP is made correctly early on, appropriate treatment can be instituted at primary care health facility and patients can benefit from it. A pilot study conducted by Hassan et al. from 10 centers in the Middle East Region in 2004 puts the prevalence of NP as 41% in the chronic low back pain patients and NocP as 59% . In a prTo achieve an accurate diagnosis, assessment by an expert pain physician including meticulous history, clinical examination, and neurological evaluation must be carried out , 29. It The original version of the PD-Q, which was evaluated in a large population of about 8000 subjects, demonstrated high sensitivity, specificity, and positive predictive accuracy. This screening tool can be used in primary care settings worldwide only if it is translated and adapted for local use and tested for psychometric properties. Several translations of the PD-Q have been developed and assessed in the corresponding local populations –21.In this study, the first important step was a translation of the original version to the Arabic language. The translation which was done as per the standard textbook Arabic language understandable by most of the Arabic speakers and was not limited to the Saudi dialect only. A panel of experts was employed to ensure that the translation and adaptation was done as per internationally accepted norms. Additionally, the prefinal version was tested on a sample population before administering it to the study population. Expert pain physicians were involved at several stages.The demographics of both study groups including age, gender, occupation, and level of education were fairly balanced indicating that there was no difference in the level of understanding of the questionnaires between the groups. The Arabic version of the PD-Q showed good psychometric properties. Stability of the PD-Q over time  was excellent. The test-retest reliability was similar to previous studies , 17. TheHence, the Arabic version of PD-Q is reliable and valid scale for detecting NP in the Arabic population. Especially in remote areas where expert help may not be available, the PD-Q can be a real boon. It can serve as a powerful tool for initial screening of patients with pain.Our study also had some limitations. This was a single tertiary hospital study as patients from multiple centers could not be included due to logistic reasons.In summary, we developed the Arabic version of the PD-Q and tested its psychometric properties. The Arabic version of PD-Q demonstrated good reliability and validity. A large-scale study in the Arabic population is required to confirm the results of this study further and further affirm the validity and reliability of the Arabic version of the PD-Q.S1 Arabic version of the painDETECT questionnaire(PUB)Click here for additional data file.


Transient neurologic symptoms (TNSs) can be distressing for patients and providers following uneventful spinal anesthesia. Spinal mepivacaine may be less commonly associated with TNS than lidocaine; however, reported rates of TNS with intrathecal mepivacaine vary considerably.  We conducted a retrospective cohort study reviewing the internal medical records of surgical patients who underwent mepivacaine spinal anesthesia at Toronto Western Hospital over the last decade to determine the rate of TNS. We defined TNS as new onset back pain that radiated to the buttocks or legs bilaterally.  We found one documented occurrence of TNS among a total of 679 mepivacaine spinal anesthetics  that were performed in 654 patients.  Our retrospective data suggest that the rate of TNS associated with mepivacaine spinal anesthesia is lower than that previously reported in the literature. Transient neurologic symptoms (TNSs), characterized by low back pain that radiates to the buttocks or legs after recovering from spinal anesthesia , 2, can Following the University Health Network ethics approval, we conducted a retrospective cohort study of consecutive patients who received spinal anesthesia with 1.5% or 2% isobaric mepivacaine for elective surgery at TWH between January 1, 2006 and March 31, 2017. We adhered to the STROBE recommendations for the design and reporting of observational studies .We identified all mepivacaine spinal anesthetics by searching our regional anesthesia database, which captures data for all neuraxial blocks performed in the block procedure room at our institution. Patients who received multiple mepivacaine spinal anesthetics for more than one operative procedure in separate encounters during the study period and patients who received rescue spinal anesthesia using mepivacaine following failed spinal anesthesia with bupivacaine during the same encounter were both included.All spinal anesthetics were performed by a regional anesthesia resident or fellow under direct supervision of the attending anesthesiologist or by the attending anesthesiologist. The dose and concentration of mepivacaine, along with any additives, were administered at the discretion of the attending anesthesiologist. Since 1.5% mepivacaine is not commercially available in Canada, this concentration was achieved by diluting 2% mepivacaine hydrochloride  with preservative-free normal saline.It is our institutional practice that any patient who develops neurologic symptoms following neuraxial anesthesia, including TNS, postdural puncture headache, and meningeal, neuropathic, or radicular symptoms, is referred to our anesthesiology department for immediate evaluation and management. Severe cases or patients at risk of acute deterioration are referred to the emergency department and/or for urgent neurosurgical consultation.For each mepivacaine spinal anesthetic administered, the corresponding patient's internal medical record was reviewed for any documented indication of TNS postoperatively. We defined TNS as new onset back pain radiating to the buttocks or legs bilaterally , 2. SpecDescriptive statistics were generated for the entire sample using counts and frequencies for categorical variables and means, medians, standard deviations, and interquartile ranges for continuous variables. The distribution of continuous variables was assessed using the Shapiro–Wilk test of normality; none of the continuous variables were normally distributed, and medians with interquartile ranges are therefore reported. We describe the cases developing the outcome, and the absolute risk of developing the outcome with 95% confidence intervals is reported. In a secondary analysis, the characteristics of excluded patients were descriptively compared with those comprising the primary cohort. Statistical analyses were performed using SAS University Edition 9.4 .A total of 679 mepivacaine spinal anesthetics were performed in 654 patients. Twenty-two patients (3.3%) underwent multiple mepivacaine spinal anesthetics, among whom 19 (2.9%) received two mepivacaine spinal anesthetics and 3 (0.4%) received three mepivacaine spinal anesthetics, each in separate encounters. Twenty-four patients (3.6%) received a rescue spinal anesthetic using isobaric mepivacaine following a failed spinal anesthetic with isobaric bupivacaine (5–15 mg) during the same encounter. Our cohort characteristics are presented in n=671, 99%), using a midline approach  with a 25 gauge  Whitacre needle . Paresthesia during intrathecal injection was reported during a single spinal anesthetic, while pain upon intrathecal injection was not reported during any spinal anesthetic. No patients experienced any new postoperative motor deficits.The spinal anesthesia procedures were most commonly performed in the sitting position . One patient was referred to our anesthesia department for consultation regarding postoperative neurological symptoms. None of the remaining 653 patients  received postoperative consultation by any medical or surgical service at our institution, none were referred to our emergency room, and none were readmitted postoperatively.Among the 679 mepivacaine spinal anesthetics performed, a single occurrence of TNS  was found in a 75-year-old female patient (BMI 27) undergoing ankle hardware removal surgery, who received spinal anesthesia with 2.5 mL of 2% mepivacaine at the L4-L5 level. No pain or paresthesia was reported during intrathecal injection. Of note, this patient appeared in our dataset twice, as she received a mepivacaine spinal anesthetic in a separate encounter for knee arthroscopy 43 months prior with no complications, but developed TNS following second exposure.In secondary analyses , the chaOur single-institution retrospective cohort study suggests that the reported rate of TNS following spinal anesthesia with isobaric mepivacaine is very low. Our large cohort of mepivacaine spinal anesthetics allowed for addressing some important shortcomings in previous small randomized trials, wherein a wide range of 0–7.5% rates of TNS have been observed , 5–8. ThUnder the conditions for detection of TNS in the present study, a single case was identified. The single observed case of TNS occurred in an elderly patient who received mepivacaine spinal anesthesia for ankle hardware removal after 43 months of having received mepivacaine spinal anesthesia for knee arthroscopy. While advanced age is independently associated with TNS , the effOur results are subject to several limitations. First, our study comprised all mepivacaine spinal anesthetics conducted in a single-academic tertiary care hospital in a manner reflective of institutional clinical practice. Similar studies at centres with differing case mixes are necessary to generalize our findings to other practice settings. Second, due to the retrospective nature of our study, there is considerable potential for underreporting. It is possible that minor episodes resolved spontaneously and may not have been recalled, reported, and/or documented. Given that TNS is routinely discussed and disclosed during our informed consent process prior to performing spinal anesthesia using mepivacaine, it is also possible that even moderate-or-severe TNS symptoms were appropriately disregarded and unreported by patients. Further, we excluded 339 mepivacaine spinal anesthetics from our analyses due to incomplete or missing postoperative data, most likely because follow-up occurred in a private clinic. However, none of these patients were referred to our anesthesia department for consultation, none received any other medical or surgical consultation, none were readmitted, and none visited our emergency room postoperatively. It is also noteworthy that the characteristics of excluded patients  did not A large cohort comprising a decade of experience with isobaric mepivacaine for short-acting spinal anesthesia at our institution suggests that the rate of TNS is lower than that previously reported in the literature. Practically, TNS following mepivacaine spinal anesthesia has not posed a clinical challenge to our regional anesthesia teaching program over the last decade.


The immunological profiles of human specific IgE (sIgE) and specific IgG4 (sIgG4) vary by genetic predisposition, living conditions in different geographical locations and patient’s age. The aim of our study was to analyze sIgE and sIgG4 patterns and their age-dependent changes in patients from the Moscow region. For identifying sIgE and sIgG4 profiles the blood samples from 513 patients aged 6 months to 17 years who were showing symptoms of allergic diseases were analyzed using microarrays containing 31 allergens. The highest sIgE prevalence was observed for birch pollen (32%) among pollen allergens, cat dander (24%) among indoor allergens, and egg whites (21%) among food allergens. The most common sIgG4 response was developed toward egg whites (80% of patients). Age-related elevation was identified for patients with increased sIgE to pollen allergens and indoor allergens (cat or dog dander and house dust mites). For each allergen, the proportion of cases with significant levels of sIgG4 appeared to increase with patients’ age. The data on allergen-specific sIgE and sIgG4 prevalence show both general trends and some local special aspects that are indicative for the Moscow region. This information should be useful in terms of epidemiology of allergic diseases. Currently, up to 20% of the adult population of developed countries and up to 40–50% of school children are affected by different allergic diseases . It is wAllergen-specific immunoglobulin E (sIgE) is the main marker of type I hypersensitivity reaction to a certain agent. sIgE bound to an allergen is involved in the process of cross-linking of FcƐRI receptors leading to basophil and mast cell activation followed by the inflammatory mediator release . Howeverin vitro analysis interpretation [Allergen-specific IgG and specifically IgG4 acting as blocking antibodies interfere IgE binding with allergens and prevent cross-linking of FcƐRI receptors and further cellular activation . Unlike retation  and is mretation .Currently, there are a number of studies describing the immunological profiles of sIgE ,10 and sTo fill this gap in the research, we performed a screening analysis of blood samples from a representative (513 patients) group of patients from the Moscow region, as well as from a typical urban region of Central Russia that has an adverse ecological situation. The profiles of sIgE and sIgG4 were identified with microarrays containing 31 allergens for patients aged 6 months to 17 years with allergic symptoms.Patients from the Filatov Moscow City Pediatric Clinic No. 13 aged 6 months to 17 completed years with suspected allergic diseases or proven allergic diagnosis  were included in the study. The continuous recruitment was carried out from May, 2016 to December, 2016. Exclusion criteria were acute infectious diseases within 2 weeks before the blood sampling and the treatment by allergen-specific immunotherapy.After the exclusion, totally 513 patients were enrolled. The written informed consent was obtained from one of the parents or guardians of the patients. The study was approved by local ethics committee of the Filatov Moscow City Pediatric Clinic No. 13.All the patients were divided into 6 age groups: 0–1, 2–3, 4–5, 6–8, 9–13, 14–17, each containing from 77 to 94 participants.The surplus of blood serum samples of participants that were left over after routine diagnostic procedures were used for the analysis of allergen-specific IgE and IgG4. The blood samples were measured once for each subject. All of the samples were exposed to a single refrigeration cycle at -45°C.The evaluation of sIgE and sIgG4 levels was performed using multiplex fluorescent microarray assays. Microarrays were developed as the matrix of hemispherical gel pads of 0.1 nl volume ,13, and The protocol for the analysis and verification of this method were described in previous studies ,15,16. BThe registration of fluorescence signals was performed using a two-wavelength microarray analyser (EIMB RAS) based on the concept of digital wide-field fluorescence microscopy . The meaThe patient was considered sensitized (demonstrated increased sIgE level) to the allergen if the level of sIgE to this allergen exceeded the minimal threshold of 0.35 IU/ml. The level of sIgG4 in relation to the allergen was considered to be significant if the concentration of sIgG4 was more than 110 ng/ml.The prevalence of sensitized individuals and the prevalence of subjects with a significant sIgG4 level were calculated for each of the age groups as the proportion of the patients with increased sIgE and significant sIgG4 for all the patients included in the group. 95% Confidence Intervals  for theDiagram plotting was performed using Microsoft Excel 2010.In this study, 513 blood serum samples were analyzed from patients aged 6 months to 17 years who were showing symptoms of allergic diseases, such as pollinosis, atopic dermatitis, and bronchial asthma.The number of patients grouped by age in 6 groups with or without increased levels of sIgE and significant levels of sIgG4 is presented in For comparison, rates of sensitization obtained in the study and the data from the two other investigations that describe sensitization among allergic populations in the Netherlands  and TaiwAs concerns sIgG4 prevalence, sIgG4 antibodies were detected most commonly in response to allergens such as egg whites (80% of patients) and cow milk and its components (46–65%). To a lesser degree, considerable levels of sIgG4 were reached in response to wheat flour (33%), peanuts and hazelnuts , and birch pollen (23%) .For most common allergens, the age-dependent changes in sIgE and sIgG4 prevalence were studied. The percentages of patients with significant levels of sIgE and sIgG4 as well as the 95% confidence intervals corresponding to these proportions are shown in Figs Pairwise evaluation of statistical significance in sIgE and sIgG4 prevalence between the age groups for observed allergens  has showThis study is focused on the analysis of allergen-specific IgE and IgG4 profiles among 513 pediatric patients from the Moscow region who demonstrated allergic symptoms. Most of the previous studies describing allergy occurrence in European countries , Asia 22,23, theThe major survey on allergies and asthma among children in Russia was carried out as part of The International Study of Asthma and Allergies in Childhood (ISAAC) . The stuInhalant allergens that were included in our study are the most common allergens responsible for allergic rhinitis and rhinoconjunctivitis among Europeans. In Central Europe, sensitization to grass pollen, especially to timothy grass, prevails among pollen allergens . AccordiD. pteronyssinus and D. farinae) is less common. sIgE response to D. pteronyssinus was detected only for 7% of the patients included in this study. Also, among the group of older patients aged from 13 to 17 years sensitization to D. pteronyssinus rises to 16% ..D. pteroAccording to the results obtained in this study, the majority of patients who are sensitized to food allergens are children under the age of 10 years . The mosSensitization to peanuts and hazelnuts among atopic children in the Moscow region is rarely observed, compared with allergens such as egg and cow milk. Nevertheless, peanuts and hazelnuts remain very important allergens, as they are responsible for up to 70–90% of all cases of food-induced anaphylactic reactions . The senIncreased sIgE levels for carrots and apples were observed in 7% and 19% of atopic children that were involved in our study . This siAge-related changes in the occurrence of sensitization to allergens in various groups of children in the Moscow region  correspoIn addition to the sIgE profile, sIgG4 profiles were evaluated for all patients. Significant levels of sIgG4 are associated with the formation of tolerance to certain allergens  since sIIn our study, the most significant level of sIgG4 was found for food allergens. Apparently, the increased synthesis of sIgG4 is a normal physiological response to consumption of proteins . At the In our study, we analyzed the levels of IgE and IgG4 which are specific to the main allergens that cause sensitization in children, as well as the age-dependence of the sensitizing activity of allergens in the Moscow region. Information on the allergens that most often cause sensitization in a particular region may be of importance for effective diagnostics and treatment of atopic patients.S1 TableConditional color formatting was used for the lines with percentages .(XLSX)Click here for additional data file.S2 Table(XLSX)Click here for additional data file.


Nontraumatic musculoskeletal disorders are the main reason for presentation to the emergency department (ED), with rachialgia (back pain) being the most common reason to request medical assessment among them. This also generates the highest demand for reassessments due to poor pain control or onset of adverse reactions to the treatment prescribed in the initial assessment.A retrospective observational study based on usual clinical practice was conducted in patients attending the ED due to low back pain during a period of 24 months. The primary objective was to determine the demand for reassessments in the ED by these patients in the following 30 days, according to the type of therapeutic approach used in the initial assessment.P = 0.001, odds ratio [OR] 0.252 with 95% confidence interval [CI] 0.100–0.635) and days 15 to 30 . Patients who received tapentadol also had a better clinical evolution of pain compared to those who did not receive it (P < 10−4) and to those who received tramadol (P < 10−4).A total of 732 patients who requested medical assessment due to back pain in the ED of our hospital were analyzed, 91 of whom were treated with tapentadol whereas 641 received another treatment. In the first month after the initial assessment, reassessments were less common in the tapentadol group; this difference was significant from days 8 to 15 (In this study in patients with back pain, tapentadol shows clear advantages over the other analgesics analyzed, in terms of pain control and less need for reassessments. In English-language medical and scientific literature, the term recidivism refers to the repeated demand for medical assessments by the same individual, due to the same complaint, or with the events derived from the management of that complaint.,2There is no equivalent translation for ,4This demand for new medical assessments represents a major challenge for the healthcare management in hospital and nonhospital units that attend to medical emergencies, and consequently for their efficient indexes. This is especially remarkable in Spain, where emergency departments (EDs) tend to be overused by the population.–8 Those conditions also generates the highest ratio of recidivism. These studies have also shown that, within nontraumatic musculoskeletal disorders, axial lumbar pain (low back pain) stands out as the most common chief complaint in EDs in all the age groups evaluated.–11 Together, this leads to an increase in the care burden and healthcare expenditure, in addition to the social cost derived from the incapacity for work that it generates. In this context, there is a need to study the possible factors that may be causing this demand for reassessments by patients who attend EDs for back pain, to establish actions that can improve clinical outcomes in these patients after the initial assessment, reduce recidivism, and increase the efficiency of the healthcare system as a whole.Recent observational studies have shown that the most frequently assessed complaint in EDs is nontraumatic musculoskeletal pain, reaching a prevalence of around 15%. The Spanish Medicines Agency approved its use in 2010, and since then it has been incorporated into the management of chronic pain in EDs.Tapentadol is a centrally acting opioid analgesic that acts as a mu-receptor agonist and norepinephrine reuptake inhibitor. It is indicated for the management of severe chronic pain that can only be adequately treated with an opioid analgesic.This study was proposed with the aim of analyzing the outcomes obtained by patients with low back pain treated in our ED, in terms of the need for reassessments as well as other secondary characteristics such as pain control, time off work and future need for specialized assessments. The type of drug treatment prescribed in the ED was used as a classification criterion, differentiating between those who received tapentadol and those who did not for the analysis.2A retrospective observational study was conducted in patients who attended our ED for back pain during a period of 24 months. The primary objective or endpoint of the study was to determine revisits to the ED for back pain (REB) in patients seen in our ED, according to the type of therapeutic management used.The study included patients of both sexes from our hospital's influence area who had been diagnosed with any type of back pain in our ED from January 2014 to December 2015 (24 months). Patients in whom signs of inflammatory back pain (spondyloarthritis of any nature) were identified were excluded, as were those in whom radiological data were missing from the patient studies performed or these studies could not be assessed, and those who presented incomplete demographic data in the ED or primary care records. Patients treated with buprenorphine, fentanile, oxycodone, or oxicodone/naloxone were excluded due to its relative insignificant number (<5 cases for each treatment).We retrospectively reviewed the medical records of these patients from the first assessment in the ED and for the following 3 months. The information was obtained from the Hospital Ramón y Cajal ED database of care records , the database of care records of the Primary Care Health Centres in our hospital catchment area (via the Horus interface) and the electronic medical records of the Hospital Ramón y Cajal ED .The variables considered were sex, age, time from onset of painful symptoms to the first visit, presence of comorbidities , concomitant treatment , Kellgren–Lawrence radiological classification grade, intensity of pain at the start of treatment assessed according to a visual analog scale  and Short Form-36 Health Survey , treatment administered in the ED, number of repeat visits to the ED for the same complaint or for complaints derived from the medication administered during the first month since the first visit, number of sick days the patient had, number of appointments scheduled in the 3 months following the initial assessment, patients who were not given an appointment in either specialized outpatients or in primary care in the 3 months following the assessment, and changes in pain intensity (VAS and SF-36) at 7 days.1: There are differences in the REB according to the type of treatment administered” versus “H0: There are no differences in the REB according to the type of treatment administered.” No studies have measured the REB according to the treatments prescribed, although around 16% (5.6) of all patients who consult for a musculoskeletal condition are known to revisit the ED within the first month for the same complaint. Assuming 16% as the worst-case scenario, proposing a reduction in the demand for reassessments of 5 percentage points  and for a 95% confidence level, with a statistical power of 80%, and assuming 15% incomplete records, 679 patients needed to be included.In order to calculate the sample size, we used the hypothesis related to the primary endpoint: “H2 test or Fisher's exact test. The behavior of the quantitative variables  was analyzed by treatment group using the Student's t test or non-parametric tests  in the case of asymmetry. Logistic regression models were fitted, to evaluate the association of those variables with a P-value <0.15 in the crude analysis. Kaplan–Meier curves were constructed to evaluate the rate of events in time and Cox models were fitted. A stratified analysis was also performed to identify possible treatment effect modifications. Accordingly, all these treatment modifications were considered therapeutic failures and ended the Kaplan–Meier curve. In all the hypothesis testing, the null hypothesis was rejected with a type I error or α error <0.05. This study was approved by our institution's independent ethics committee.The software used for data analysis was STATA version 12.0 (StataCorp). A descriptive analysis was performed of all the study variables . The association between qualitative effectiveness variables (REB) and the treatment groups was evaluated using the χ2.1This study was conducted in accordance with Good Clinical Practice Guidelines and the Declaration of Helsinki, and current applicable legislation.2.2No patient was interviewed in person or by telephone, nor did they undergo any diagnostic or therapeutic tests.2.3The authors declare that they have followed their center's protocols regarding the publication of patient data.2.4The authors declare that no patient data appear in this article.3A total of 732 patients diagnosed with back pain in the ED were analyses, of whom 91 received drug treatment with tapentadol (tapentadol group). The remaining 641 patients—who did not receive tapentadol—were considered the control group. Table P = 0.005), with the percentage of men significantly lower in the tapentadol group compared with the control group (P = 0.001). With respect to the associated comorbidity, hypertension, and compression fractures of the back were significantly more common in the tapentadol group . The Kellgren–Lawrence radiological classification grade was significantly higher in the tapentadol group (P < 10−4); no significant association was observed between the grade and baseline VAS or short form health survey 36 (SF-36) scores. Regarding the intensity of pain, in the tapentadol group, the baseline VAS was significantly higher and the SF-36 significantly lower (P < 10−4 in both cases).Patients in the tapentadol group were significantly younger . Table P < 10−4), although it should be taken into account that there are more exceptions due to missing baseline values in the control group. It is important to note that this better clinical evolution of pain in patients who received tapentadol was also observed with respect to patients who received tramadol (P < 10−4), and in patients who received tramadol with respect to those who did not receive either tramadol or tapentadol .Patients who received tapentadol had better clinical evolution of pain (reduction in VAS and increase in SF-36) in the 7 days following the initial assessment, compared with patients in the control group (P = 0.186). It was significant, however, at days 8 to 14  and days 15 to 30 . In this respect, it was observed that patients who received tapentadol returned significantly later than those who did not receive either tramadol or tapentadol (2-sided significance P = 0.004) and especially compared with those to received tramadol (P < 10−4), who in turn also returned significantly earlier than those who received neither tramadol nor tapentadol (P < 10−4). No correlations were detected with the tapentadol or tramadol dose.Regarding the reassessments requested within the first month after the initial assessment according to the treatment administered, reassessments were less common in the tapentadol group compared with the control group Graph , althougρSpearman = −0.076 and τKendall = −0.061, which, given the large number of cases (732) are significant with 2-sided significance P = 0.040 in both cases. No correlation was observed with the tapentadol dose administered.The time to the first reassessment requested tended to decrease very slightly for higher baseline VAS, with nonparametric correlation coefficients Graph P < 10−4 in Fisher's exact test and OR 0.258 with 95% CI 0.147–0.453), whereas the reduction in the (few) reassessments requested for adverse effects is not significant.In total, 50.3% of patients in the control group  and 20.9% of patients in the tapentadol group  visited the ED for reassessment in the first 30 days after the initial assessment in the ED, as shown in Graph 4,14Pharmacological treatment for back pain can include paracetamol, metamizole, NSAIDs , minor opiates (tramadol), tapentadol and major opiates , and depends on a range of factors including the pain intensity, cause of the pain and response to previous analgesic treatments. Tapentadol treatment should be individualized, seeking the optimal dose that provides the patient with pain relief without the onset of toxic effects, and always under proper medical supervision. It is also available in an extended-release formulation (retard), which permits adequate pain control with twice daily administration. In clinical trials conducted on knee osteoarthritis and low back pain, tapentadol was superior to placebo and non-inferior to oxycodone in the reduction of pain intensity.–19 Furthermore, in a systematic review that included 4 clinical trials—3 phase III clinical trials with 12-week follow-up and an open-label safety study with 52-week follow-up, in a total of 4094 patients with osteoarthritis, low back pain or both—extended-release tapentadol was associated with a reduction in pain intensity compared with placebo and oxycodone. However, the clinical significance of these results is limited, due largely to differences in the variables used to evaluate the efficacy in the reduction of pain intensity and the high withdrawal rates in the clinical trials analyzed.,21Tapentadol is a potent analgesic with a dual mechanism of action, as it contains μ-opioid receptor agonist activity and additional noradrenaline reuptake inhibition properties. It is indicated for the control of severe chronic pain in adults that can only be adequately treated with an opioid analgesic.No studies have been published to date that have evaluated the impact of the different drug treatments available for the treatment of back pain on the demand for reassessments by patients due to poor pain control despite the treatment prescribed or due to the onset of treatment-induced adverse reactions.In our study, it was observed that the group of patients initially treated with tapentadol demanded urgent reassessment within 30 days after the initial assessment less often than those who did not receive tapentadol, with this difference being more marked between 8 and 30 days after the initial assessment. This finding is consistent with the fact that patients treated initially with tapentadol had better clinical pain evolution (reduction in VAS score and an increase in SF-36 score) compared with those who did not receive tapentadol. In the set of patients analyzed, almost all the first reassessments requested were for poor pain control, with the proportion of patients significantly lower in those who received tapentadol compared with those who did not receive it.It should be considered that, as this is a retrospective observational study, there were limitations in the availability of some clinical data, particularly in patients in the control group . These limitations were taken into account when analyzing the data and interpreting the results. Other limitations related to the retrospective design of our study are the leak of assessment of the potential psychogenic component for ED visits that had been recently described and may be considered as a focus of interest in further research initiatives.5The prescription of adequate drug treatment for back pain in EDs—always after conducting a complete anamnesis and the pertinent examinations in each case—is essential to achieve optimal pain control, while at the same time avoiding the onset of adverse reactions and ultimately reducing the demand for reassessments in the ED by these patients.In this study in patients with low back pain, tapentadol shows clear advantages as regards pain control and less need for reassessments in the ED over the other analgesics analyzed.


Ambulatory surgery has recently gain popularity, as it is a good method of optimizinghospital resources utilization. To support ambulatory surgery, anaesthetic goals nowrevolve around patients’ early recovery with minimal pain and nausea, expedientdischarge home and prompt resumption of activities of daily living. In this study, weevaluated the effect of a single pre-induction dose of dexmedetomidine on anaestheticrequirements, postoperative pain and clinical recovery after ambulatory ureteroscopy andureteric stenting under general anaesthesia.1  or IV saline . General anaesthesia was maintained with Sevoflurane: oxygen: air, titrated to BIS 40–60. Pain intensity, sedation, rescue analgesics, nausea/vomiting and resumption of daily activities were recorded at 1 h, and postoperative day (POD) 1–5.Sixty patients were randomised to receive IV dexmedetomidine 0.5 μg.kg-p = 0.037; reduced postoperative resting pain at 1 h (VAS 0–10) (mean (SD) 1.00 (1.84) vs. 2.63 (2.78), p = 0.004), POD 1 (mean (SD) 1.50 (1.48) vs. 2.87 (2.72), p = 0.002), POD 2 (0.53 (0.97) vs. 1.73 (1.96), p = 0.001) and POD 3 (0.30 (0.75) vs. 0.89 (1.49), p = 0.001). DEX patients also had less pain on movement POD 1 (3.00 (2.12) vs. 4.30 (3.10), p = 0.043) and POD 2 (2.10 (1.98) vs. 3.10 (2.46), p = 0.040), with higher resumption of daily activities by 48 h compared to placebo, 87% vs. 63%, p = 0.04.Group DEX patients had significant reduction in sevoflurane minimum alveolar concentration (MAC), mean (SD) DEX vs. Placebo 0.6 (0.2) vs. 0.9 (0.1), We conclude that a single dose of dexmedetomidine was a useful adjuvant in reducing MAC and postoperative pain (at 1 h and POD 1–3), facilitating faster return to daily activities by 48 h.ACTRN12617001120369, 31st July 2017, retrospectively registered.The Australian New Zealand Clinical Trials Registry (ANZCTR),  There has been a paradigm shift in modern healthcare and hospital resource utilization, such that most surgical procedures are now performed in an ambulatory outpatient setting versus traditional hospitalization postoperatively . AnaesthDexmedetomidine is a highly selective α-2-agonist exhibiting dose-dependent sedation, analgesia, anti-inflammatory, sympatholytic and anxiolytic effects without relevant respiratory depression , 9. The 1for ureteric stenting after ureteroscopy, and assess its impact on intraoperative anaesthetic agent requirements eg. minimum alveolar concentration (MAC) {primary outcome} and secondarily, postoperative pain scores, perioperative analgesia requirements, PONV, drowsiness and return to normal daily activities.This clinical equipoise left questions unanswered for our institution as to whether it would benefit our patient cohort. We also found no studies evaluating the effect of a single adjuvant dose of dexmedetomidine on patients undergoing ureteroscopy in an ambulatory surgery setting and therefore embarked on this study. We aimed to evaluate the efficacy of a single pre-induction dose of Dexmedetomidine 0.5 μg.kg-−1), advanced liver disease , those with a history of chronic use of sedatives/ narcotics or analgesics, known alcohol or drug abuse, allergy to study medication and those devoid of postoperative telephone access.The study was approved by our institutional review board  and all patients gave their written informed consent. We enrolled patients of American Society Anaesthesiologists physical status I-II, aged between 18 and 65 years who underwent elective ureteroscopy and ureteric stenting in our ambulatory day care centre. We excluded patients with increased serum creatinine  vs. Placebo using a computer generated random number table in blocks of ten. After recruitment, the enrolling investigators opened sealed opaque envelopes that concealed group allocation. Participants, trial investigators, attending anaesthetists providing general anaesthesia for the cases, and outcome assesors were blinded to group allocation.As part of our routine practice, all patients received oral paracetamol 1 g when they entered the preoperative holding bay. Intravenous access was secured with a 20 Gauge catheter, and ECG, noninvasive arterial blood pressure, heart rate (HR), pulse oximetry and a Bispectral Index (BIS) forehead strip was applied. Haemodynamic variables were recorded at five-minute intervals but BIS monitoring  was only commenced later in the operating room.−1 and the placebo group received 20mls of normal saline. The study drug was prepared by one of the investigators by drawing up the required dose and diluting it with normal saline to a total of 20 ml in a 20 ml syringe, which was then infused over 10 min via an infusion pump  by a blinded nurse not involved in the trial. Patients in the placebo group received 20 ml normal saline in a similar syringe by the same method. The investigator that prepared the drug had no further role in assessing the patient’s outcomes.In this holding bay, patients randomised to Group DEX then received intravenous dexmedetomedine  0.5 μg.kg−1 and propofol 2 mg.kg−1 or more until patient’s exhibited loss of verbal response. A classic laryngeal mask airway size 3 or 4 was then inserted when the jaw was sufficiently slack. Effective ventilation was confirmed on capnography and anaesthesia maintained with Sevoflurane in an oxygen: air mixture. Depth of anaesthesia was titrated to maintain BIS values between 40 to 60. Intraoperative IV fentanyl in 50 μg boluses were given if deemed necessary by the attending anaesthetist managing the case. No routine anti-emetics were administered. The blood pressure, HR, mean alveolar concentration (MAC) of Sevoflurane and BIS values were recorded every 5 min. Upon completion of surgery, the LMA was removed in the operating room when patients were awake and obeying commands.Post-administration of the study drug, all patients were transferred to the operating rooms within 10 min, where the patients were positioned on the operating table with their heads resting on a jelly doughnut. Standard monitoring was applied and patients were preoxygenated with 100% oxygen for 3 min, before anaesthesia was induced with IV fentanyl 1 μg.kgPostoperatively, patients were assessed in the Post Anaesthesia Care Unit (PACU) within the hour for sedation levels  1–5. Patients were quizzed on their pain at rest  and movement (rising from supine position), incidence of nausea/ vomiting and their ability to ambulate and resume daily activities.p value of <0.05 was deemed statistically significant.The sample size was based on our primary outcome measure of anaesthetic minimum alveolar concentration (MAC). From an initial pilot study involving 16 patients, we found the mean (SD) MAC of Sevoflurane to be 0.85 (0.19). We deemed a MAC reduction of 0.25 to be a clinically significant difference between the two groups. Prospective power analysis indicated that with α = 0.05 and power of 80%, 25 patients per group were needed. Therefore we recruited 30 patients per group to account for dropouts. Parametric data and non-parametric data was analysed with Student’s t test and Mann-Whitney U-test respectively, and Fisher’s exact test was used to compare side effects using SPSS 15.0 ™  software. A Seventy patients were recruited to the study Fig.  and sixtp = 0.037  which then became insignificant at the 20 min mark  MAC of the anaesthetic agent in patients who had received dexmedetomidine compared to placebo, 0.6 0.2) vs. 0.9 (0.1),  vs. 0.9 p = 0.004. Correspondingly, no patient in the DEX group required rescue fentanyl in the PACU, whereas the placebo group received a median [range] of 0 [0–100] μg, p = 0.040 (Table p = 0.160.Intraoperatively both groups received the same amount of fentanyl but the mean (SD) VAS for postoperative pain 1 h after PACU arrival was significantly reduced in the dexmedetomidine group, 1.00 (1.84) vs. 2.63 (2.78), 9; he remained afebrile with a normal total white count and was discharged the next day.Two patients in the placebo group required unexpected hospital admission. One patient had uncontrolled pain postoperatively (VAS 10/10) despite receiving IV Tramadol 100 mg in addition to fentanyl rescue in the PACU and requested admission for pain control and rest for 3 days. Another patient had persistent haematuria and was admitted for observation overnight on a suspicion of sepsis but his haemoglobin level was 12.8 g/dL, platelets 271 × 10p = 0.037 [Fig. Upon discharge home, patients that had received DEX had significantly reduced pain scores at rest on the first to third postoperative day, and reduced pain on movement on POD 1 and POD 2 compared to placebo Table . Signifi037 Fig. . On POD −1 administered to patients undergoing ureteroscopy and ureteric stenting, not only significantly reduced the MAC of sevoflurane and intraoperative anaesthetic requirements, but also resulted in a significant 60% reduction in immediate postoperative pain that extended to Day 3 post discharge. This was highly clinically significant allowing patients earlier resumption of their daily activities and return to the workforce without being hindered by pain occurring after discharge that would negatively impact patient recovery and place economic burdens on patient and caregiver [In this study, we demonstrated that a single pre-induction dose of dexmedetomidine 0.5 μg.kgaregiver .2-Adrenergic agonists produce clinical effects after binding to α2-adrenergic receptors, of which there are three subtypes . These receptor subtypes are distributed ubiquitously, and early mice studies investigating substitution of a mutant α2A-adrenergic receptor via ‘hit and run’ gene targeting revealed the role of this subtype in sedative, analgesic, and anesthetic-sparing responses in vivo [−1) had a 47% reduction in the MAC of isoflurane, compared to without dexmedetomidine [−1 administered 15 min pre- induction of anaesthesia also decreased the MAC of sevoflurane by 17% in adults aged 55–70 years [−1 with relative haemodynamic stability [−1 resulted in lower mean intraoperative isoflurane concentration in the dexmedetomidine-treated patients than controls undergoing minor orthopaedic and general surgery [α in vivo . In thistomidine . Similar70 years . Other i surgery .−1 reduced immediate postoperative pain by a dramatic 60% in our study. This resonates with Schnabel’s [The analgesic properties of dexmedetomidine are conferred by its central action at the locus coeruleus in the brainstem and spinal cord that inhibits neuronal firing, thereby triggering sedation, anxiolysis and analgesia . We founhnabel’s  meta-anahnabel’s .On the other hand, other investigators have found no difference in immediate postoperative pain but a reduced amount of rescue analgesics after a single dose of dexmedetomidine . A recen−1 resulted in significantly reduced pain at rest for the first three postoperative days, and pain on movement on postoperative day 1 and 2. More importantly, our study significantly showed that this extended analgesic duration actually enhanced the recovery profile and enabled a significantly higher proportion of patients to return to their daily activities after 48 h as compared to the placebo group. In a rat model of incisional pain, pretreatment with dexmedetomidine (administered subcutaneously 30 min before plantar incision) significantly decreased remifentanil-induced postoperative hyperalgesia as indicated by increased paw withdrawal latencies and thresholds to thermal and mechanical stimulation [We demonstrated that a single pre-induction dose of dexmedetomidine 0.5 μg.kgmulation . Westernmulation . These rFurthermore, surgical injury to tissue results in increased stress hormone production , complement system activation, leukocyte migration to the site of injury, and release of cytokines , superoxide radicals, proteases and growth factors . AlthougThere is already evidence that dexmedetomidine decreases production of inflammatory cytokines while lowering intra-abdominal pressure in critically ill patients with sepsis . In animOur study did have the following limitations: preoperative assessment could have ideally screened for motion sickness and smoking as possible confounders of the risk of PONV. We did not measure plasma catecholamine or stress-hormone concentrations; although these measurements could support a direct relationship between sympatholytic and anti-inflammatory properties of dexmedetomidine and postoperative quality of recovery, it was not logistically feasible under the constraints of our study setting. Additionally as MAC decreases 4% to 5% per degree centigrade decrease in core temperature , tempera−1 reduced intraoperative anaesthetic agent requirements, alleviated immediate postoperative pain in hospital and up to the third postoperative day at home, enabling patients earlier resumption of their daily activities. We conclude that dexmedetomidine proved to be a useful adjuvant that benefitted our patients undergoing ureteroscopy and ureteric stenting in our ambulatory clinical practice.In conclusion, we found that the administration of a single pre-induction dose of dexmedetomidine 0.5 μg.kg


Aim. To determine whether preassessment improves bowel preparation quality and prevents renal deterioration for chronic kidney disease (CKD) patients. Methods. Data was collected prospectively starting in January 2011 for 12 months. Patients were divided according to the presence or absence of preassessment and stratified to one of three risk groups based on patient's comorbidities and identified risk factors for poor bowel preparation; group 1 had no risk factors, group 2 had 1 risk factor, and group 3 patients had 2 or more risk factors. The association between preassessment and bowel preparation quality was analyzed using binary logistic regression. Results. 1840 colonoscopies were carried out during the period. Total number analyzed was 1704. 404 patients were preassessed. Preassessment patients had significantly better bowel preparation across all groups . Group 3 patients were 52% more likely to have good bowel preparation (p = 0.04) if they had been preassessed. Eighty-eight patients were identified with an eGFR < 60 mL/min. There was a significant difference in the eGFR percentage change between patients with preassessment and those without (p = 0.006). Conclusions. Face-to-face preassessment appears to improve the quality of bowel preparation and aids in minimizing the risk of renal injury in patients with CKD. Oral bowel cleansing agents (OBCAs) are widely used before endoscopic and radiological assessment of the bowel and preoperatively to aid with diagnostic assessment and minimizing fecal contamination . For theIn 2009, the National Patient Safety Agency (NPSA) issued an alert regarding the use of OBCAs [Consequently, consensus guidelines were developed in late 2009, which were later revised and published in 2012 .NHS trusts in the UK have responded to these guidelines with varying approaches and as such changes have a significant impact on resources. One way of implementing such guidelines has been with the use of a preassessment clinic. Organizing a specific precolonoscopy assessment, either by telephone or in person, has been shown to be beneficial in reducing nonattendance for the procedure itself . As yet,The aim of this study was to assess how preassessment of patients prior to colonoscopy affects a number of factors concerned with the procedure itself, including the quality of bowel preparation, the effect on eGFR, and the use of a risk-stratifying approach, to deploy resources for those patients most at risk of poor bowel preparation.This was an observational, retrospective study of prospectively collected data from our endoscopy database. Data were collected over a period of 12 months starting in January 2011; preassessment began at our NHS trust in September 2011. This study was approved by the local research authority, and the need for informed consent from the patients was waived.Assessment was made in keeping with the 4 recommendations made by the NPSA: (1) clinical assessment prior to prescription of bowel preparation, 2) authorization of bowel preparation by clinical staff, (3) an explanation of the bowel preparation to every patient, and (4) dispensation of the medication by authorized professionals with verbal and written explanation [ authorizn = 1840) into 3 risk groups, based on the presence of “at-risk” conditions. Patients in group 1 had no risk factors, while those in group 2 had at least 1 risk factor, and those in group 3 had 2 or more risk factors.Patients who underwent colonoscopy at our unit were divided into 2 groups, based on whether they received preassessment; patients between January and August were not preassessed, while the majority who were seen between September and December did receive preassessment. All patients from both groups were retrospectively assessed for “at-risk” conditions; “at-risk” patients included those on regular medication considered unsafe by the NPSA and consensus guidelines when administering OBCAs (NSAIDs and diuretics), patients with comorbidities , and those who were “at risk” of poor bowel preparation (constipation and limited mobility). A proforma, based on NPSA recommendations, was used to stratify the study group  test. A p value ≤ 0.05 was considered significant for all statistical tests. The effect of preassessment (with or without) on the quality of bowel preparation was analyzed initially for all 3 groups combined and then for each group separately, using binary logistic regression. All statistical tests were performed using SPSS .Continuous variables are expressed as mean (standard deviation) or median; group comparisons were carried out using then = 1354) of patients had good bowel preparation, while 20.5% had poor bowel preparation (n = 350). Patient demographic characteristics are shown in During the study period, 1840 colonoscopies were performed. Patients were omitted when there was no clear comment on the quality of bowel preparation in the report, giving a final study cohort of 1704 patients. The mean age was 61.7 years (range 16–94). A total of 404 patients received preassessment. With respect to the quality of bowel preparation, 79.5% (p = 0.002). In groups 1 and 2, the likelihood of having a good quality bowel preparation was 80% and 72% higher, respectively, in patients who received preassessment; however, these improvements did not reach statistical significance (p = 0.039) than those who were not preassessed. Age and sex were not shown to affect the quality of bowel preparation in our study.Preassessment significantly increased the quality of bowel preparation across all groups .We examined the reasons for an incomplete colonoscopy . A greatn = 404). 9.7% (n = 39) of the patients within the preassessment group were discussed with the gastroenterologist in view of significant concerns. Out of 28 patients with CKD who underwent preassessment, 12 patients  were hospitalized, such that renal function could be closely monitored. To prevent deterioration in eGFR and to improve quality of bowel preparation, 4.6% of patients (n = 20) had alteration to their medications. Extrabowel preparation was given to 6.7% (n = 27) of patients with history of severe constipation.Additionally, we looked at the interventions  that werp = 0.006, Mann–Whitney)  and those who did not (median = −6.6%) (Whitney) .p = 0.039) in the quality of bowel preparation. This cohort was considered to be at high risk because of the presence of multiple comorbidities, which appear to affect the quality of bowel preparation and may ultimately lead to failed and/or repeat procedures. Preassessment in this group facilitates specific modification of medications and bowel OBCAs to optimize the chance of a successful procedure. While the likelihood of good bowel preparation improved with preassessment in groups 1 and 2, no statistical significance was observed. One could argue that these groups, and especially group 1, have the least to gain from drug alterations or modification of OBCAs, due to a comparatively lower initial risk. As such, this study may not be adequately designed to determine any significant benefits for groups 1 and 2.Our study suggests that face-to-face assessment prior to colonoscopy, as recommended by the consensus guidelines for prescription and administration of OBCAs, improves the quality of bowel preparation across all patients. Most notably, patients stratified into group 3, the “high-risk” group, exhibited a statistically significant improvement . In the context of the fecal occult blood test colorectal cancer-screening program, Rodger and Steele found that when patients were given the option of a telephone interview rather than a face-to-face consultation, nonattendance rates fell from 14.9% to 0.8% . Of thisWith this in mind, the possibility of telephone preassessment looks promising. By streamlining our service, we have the opportunity to save selected low-risk patients time and money, while simultaneously improving the quality of the colonoscopy through telephone preassessment. By freeing up resources at the hospital, more time and attention can be dedicated to patients with greater risk, resulting in better planning, reduced risk of harm to patients, and increased efficiency of the hospital's service. Further work is necessary in this field, specifically related to patient satisfaction, attendance rates, and cost effectiveness.Limitations of our study include the retrospective analysis of clinic letters, which occasionally lacked crucial data, including information about CKD and medication history. This may have led to a bias of Type 2 error and could have underreported the presence of comorbidities. Thus, in some cases, patients may have been wrongly grouped into lower risk groups. There are patients between September and December who have been allocated into the group who had no preassessment. This may have led to a selection bias in this study. There could have been few uncontrolled factors which might have affected the results such as patient education, alteration of medications by gastroenterologists, and timing of procedure (morning versus afternoon) and body mass index. These factors were not described in the current study. The risk-stratification model is limited to some extent in that it was not validated. Additionally, the measurement of eGFR was rarely consistent with respect to timing, which may have altered the observed percentage change in eGFR.Our study has shown that face-to-face preassessment improves the quality of bowel preparation for patients undergoing colonoscopy, with the greatest benefit observed in “at-risk” patients who are at a high risk of poor bowel preparation. We have demonstrated that preassessment significantly reduces the fall in eGFR for those with CKD and helps minimize the risk of renal injury. We conclude that preassessment is a prerequisite for patients who are at risk of poor bowel preparation and with significant comorbidities. Further studies regarding the potential for telephone assessment of low-risk patients may highlight further opportunities to improve patient safety and resource allocation.


The minimum clinically important difference (MCID) is used to interpret the clinical relevance of results reported by trials and meta-analyses as well as to plan sample sizes in new studies. However, there is a lack of consensus about the size of MCID in acute pain, which is a core symptom affecting patients across many clinical conditions.We identified and systematically reviewed empirical studies of MCID in acute pain. We searched PubMed, EMBASE and Cochrane Library, and included prospective studies determining MCID using a patient-reported anchor and a one-dimensional pain scale . We summarised results and explored reasons for heterogeneity applying meta-regression, subgroup analyses and individual patient data meta-analyses.2 = 93%, relative MCID: I2 = 75%) and results were therefore presented qualitatively, while analyses focused on exploring reasons for heterogeneity. The reported absolute MCID values ranged widely from 8 to 40 mm  and the relative MCID values from 13% to 85%. From analyses of individual patient data , we found baseline pain strongly associated with absolute, but not relative, MCID as patients with higher baseline pain needed larger pain reduction to perceive relief. Subgroup analyses showed that the definition of improved patients  and the design of studies (single or multiple measurements) also influenced MCID values.We included 37 studies (8479 patients). Thirty-five studies used a mean change approach, i.e. MCID was assessed as the mean difference in pain score among patients who reported a minimum degree of improvement, while seven studies used a threshold approach, i.e. MCID was assessed as the threshold in pain reduction associated with the best accuracy (sensitivity and specificity) for identifying improved patients. Meta-analyses found considerable heterogeneity between studies (absolute MCID: IThe MCID in acute pain varied greatly between studies and was influenced by baseline pain, definitions of improved patients and study design. MCID is context-specific and potentially misguiding if determined, applied or interpreted inappropriately. Explicit and conscientious reflections on the choice of a reference value are required when using MCID to classify research results as clinically important or trivial. It can be challenging to decide whether a modest effect in a randomised clinical trial, or a meta-analysis of several such trials, is clinically relevant. Statistical tests inform on the probability of a result being a chance finding; however, they convey no information on whether a given effect will be experienced as important by patients. The degree of pain reduction that is considered clinically relevant has an impact on which analgesic interventions are regarded clinically useful. This interpretation problem for clinical relevance has been at the core of debates of the importance of several types of interventions intended for reducing acute pain, for example, paracetamol –3, non-sthe smallest difference in score in the domain of interest which participants perceive as beneficial and which would mandate, in the absence of troublesome side effects and costs, a change in the patient’s management”. The strength of the concept is that it defines a relevant effect size based on clinical considerations and not only statistical significance [Jaeschke et al.  characteificance , 14. It ificance .The minimum clinically important difference (MCID) is sometimes chosen on the basis of expert consensus judgement , statistWe therefore decided to systematically review empirical studies of the MCID in acute pain relief and to examine possible causes for variation between study results, especially their likely dependency on baseline pain levels. We also reviewed studies of the substantial clinically important difference in acute pain relief as well as clinically important differences for worsening of pain.We included prospective studies of patients with acute pain, regardless of age, clinical condition, and intervention, in which pain intensity was assessed on a one-dimensional scale, e.g. a 100 mm visual analogue scale (VAS) or a 0–10 point numerical rating scale (NRS), and in which the MCID was determined using an anchor-based method using patients’ perception of change to determine clinical importance. Pain was considered acute when its duration was less than 1 month or, if duration was not indicated, when it was described as such in a study report.Studies were excluded if they were not clinical  or determined the MCID from objective criteria (e.g. return to work), the distribution of data (e.g. the minimum detectable difference) or expert consensus.A typical eligible study would ask patients to score their pain intensity, e.g. using a VAS, at baseline and follow-up. At follow-up, patients were also asked to categorise their change in pain intensity using response options such as ‘no change’, ‘a little better’/‘somewhat better’, and ‘a lot better’/‘much better’. The MCID was then determined from the change in scores on the pain scale among patients having categorised their change as ‘a little better’ .We included studies with two types of analytical approaches (1) the ‘mean change approach’, i.e. the mean difference in pain scores among patients who reported a minimum degree of pain relief ; or (2) We searched PubMed, EMBASE and Cochrane Library until August 2016 with no language restrictions. The core search string was:  AND (difference OR change OR relief OR reduction) AND  AND (pain) with variations according to the specific database (Appendix 1). The reference lists of all included studies and relevant review papers were read systematically to identify further studies.Screening of titles and abstracts to determine the eligibility of studies was done by the primary author (MFO), while the selected full-text records were examined by two researchers independently . Any disagreement was solved by discussion.Data extraction was conducted by two researchers independently  using pretested data extraction forms generated in EpiData . Any disagreements were solved by discussion.For each study, we extracted descriptive data including publication year, study design, setting, clinical condition, type of intervention, sampling method, sample size, and definition of patients with relevant change (see Appendix 2 for complete list). For studies using a mean change approach, we extracted the following outcome data: the MCID for pain relief  and for pain worsening , as well as the substantial clinically important difference for relief and worsening of pain . We extracted MCIDs as the mean change in pain score among patients who indicated a one-category improvement (e.g. ‘a little better’). If unavailable, we extracted the mean change among patients who were minimally improved by authors’ definition (e.g. some authors defined minimum important change as the mean change in pain score among patients with either a one- or two-category improvement). Similarly, we extracted the substantial clinically important differences as the mean change among patients with a two-category improvement or used the authors’ definition. We extracted the point estimate of outcomes with their corresponding standard error or, if unavailable, other measures of variation such as standard deviation or 95% confidence interval.For studies using a threshold approach, we extracted information about definition of responders (i.e. patients with relevant change) and non-responders and the cut-off point with its corresponding sensitivity (i.e. percentage of responders correctly classified as such) and specificity (i.e. percentage of non-responders correctly classified as such). If studies reported pain scores from several concurrent pain assessments (e.g. back pain and leg pain), we extracted the assessment where more data were available or, if no difference found, we randomly selected which to extract. All scales were standardised to a 0–100 mm scale. When studies reported pain assessments based on both VAS and NRS, we used the assessment based on VAS.If the primary outcome or other key variables were unclear or incompletely reported from a study, we contacted the corresponding author. In cases where authors provided individual patient data, we first checked whether we could replicate a main result of the published paper. We then calculated estimates of absolute and relative MCIDs and their corresponding standard error.For each study, we assessed risk of attrition bias (studies were considered low risk when attrition < 10%) and risk of non-representative study sample (studies were considered low risk if using consecutive or random sampling).For each study, we extracted or calculated the MCID for pain relief (absolute and relative change), and noted results of any study-based exploration of causes for variation, e.g. baseline pain.We then summarised results qualitatively as there was considerable clinical and methodological variation between studies and heterogeneity in their results. To provide an overview, we first reported the range of results for all studies and then the range and median results with inter-quartile ranges (IQR) of studies according to analytical strategy (mean change or threshold approach). To facilitate exploration of reasons for heterogeneity, we then pooled results of studies using mean change approach with inverse-variance meta-analysis using random effects models. We studied the association with baseline pain scores in three different analyses. First, we explored the impact of the average population baseline pain in a mixed-effects meta-regression . SecondWe furthermore used subgroup analyses to explore whether between-study variation was explained by differences in other clinical and methodological factors including clinical condition, type of pain scale (VAS vs. NRS), directionality of global transition scale , definition of minimum clinically important change , number of pain assessments per patients (single vs. multiple assessments adjusted for correlation vs. multiple unadjusted assessments), risk of attrition bias (low vs. high or unclear) and risk of non-representative sample (low vs. high or unclear).Finally, we used the same analysis strategy for the substantial clinically important difference for pain relief (absolute and relative change) and the minimum and substantial clinically important differences for worsening of pain (absolute and relative change).All data analyses were done using Stata/IC version 13.We screened 1553 database records and read 273 full-text publications  or unspecified pain (7 studies). Additional studies included patients with post-operative pain, cancer-related pain, sickle cell crisis, rheumatic pain, abdominal pain, low back pain, or headache , and 12 studies used an 11-point NRS (0–10), while five studies used both scales. In 32 studies, patients compared their current pain intensity with pain at their previous assessment, while they were asked to assess the effect of their treatment in five studies , 49, 58.For studies using the mean change approach, the majority defined the minimum clinically important improvement as a one-category improvement on the transition scale 31 studies). The response categories were similar with wordings such as ‘a little less pain’, ‘a bit better’, ‘slightly improved’, or ‘slight relief’. In four studies, the MCID was defined as the mean change in pain score among patients with a one- or two-category improvement, thereby combing patients answering ‘some relief’ and ‘partial relief’ , or ‘muc studies.P values of analyses were adjusted for correlation between estimates, for example, with Generalised Estimation Equation, while the remaining studies either made no adjustment or did not report this. Access to individual patient data increased the number of studies with adjusted estimates to 11.Pain intensity was assessed at baseline and a single follow-up measurement in 14 studies, while it was assessed at multiple (from 2 to 16) follow-ups with intervals between 10 and 45 minutes in 23 studies. The latter group then derived their outcome as the summarised mean difference in pain score from the patients’ previous pain assessment when they reported minimum relief (i.e. a series of ‘moving baselines’). In eight of these studies, the In 10 studies, the MCID was defined as a numeric change for all patients with minimum change, regardless of whether pain had improved or worsened. After contacting authors, separate estimates for pain relief were available from eight of these.Standardised to a 100-mm scale, the absolute MCID in 30 studies ranged from 8 to 40 mm, and the relative difference in 15 studies ranged from 13% to 85%.The determination of the MCID was based on a mean change approach in 35 studies, of which 30 (6598 patients) were included in our analyses and five were disregarded (see below). Twenty-nine studies (6517 patients) reported absolute values ranging from 8 to 40 mm, with a median of 17 mm (IQR 14–23 mm) Fig. . Only niFor data syntheses, we did not include results from five of the 35 studies (1567 patients) as they did not differentiate between pain relief and pain worsening , 48, bec2 = 93%, P < 0.001 and I2 = 75%, P < 0.001  reporting absolute values and 11 studies (1397 patients) reporting relative values. Meta-analyses of both the absolute and relative values showed considerable heterogeneity: ISeven of the 37 included studies (2602 patients) determined clinically important differences as a threshold to differentiate between patients with or without relevant pain relief. Absolute thresholds ranged from 10 to 35 mm in six studies (2331 patients) with a median of 10 mm, and the relative threshold ranged from 15% to 50% in four studies (534 patients) . Of nine studies assessing absolute change, seven reported an association , 58, 59.P = 0.70) nor relative  estimates of MCIDs.Based on meta-regression, we found no association between baseline pain and either absolute . Thus, for patients with initial pain levels < 40, 40–70, and > 70 mm, the MCIDs were 6 (4–8) mm, 13 (11–14) mm, and 21 (20–23) mm, respectively .However, based on individual patient data meta-analysis from seven studies (918 patients), we found a strong association between baseline pain and absolute MCID, showing that patients with higher baseline pain need larger pain reduction to reach a minimum clinically relevant difference compared to patients with lower baseline pain. For each 10 mm increase in baseline pain, MCID increased by 3.1 mm  mm at 30 minutes to 11 (8 to 14) mm at 120 minutes. However, patients’ pain level declined accordingly during the multiple follow-ups and estimates expressed as a relative change from the previous assessment therefore did not decline.Other causes for variation in the MCID were addressed in 10 studies. These included sex , 57, 59,The MCIDs of subgroups are presented as medians and pooled averages, respectively Table . For mosThe supplementary outcomes for pain relief and worsening were only reported from studies using the mean change approach Table . The resWe included 37 studies (8479 patients) assessing the MCID in acute pain, of which 35 used the mean change approach and seven used the threshold approach. Meta-analyses found considerable heterogeneity between studies and, consequently, no single value of minimum clinical important difference could be meaningfully determined. Study results ranged widely both when they were reported as absolute change (from 8 to 40 mm) and as relative change from baseline (from 13 to 85%). The median of study results based on the mean change approach was 17 (IQR 14 to 23) mm and 23 (IQR 18 to 36) % for absolute and relative values, respectively. Reasons for heterogeneity were explored and baseline pain was identified as a cause of variation in absolute, but not relative, outcomes. In addition, the definition of minimum clinically important change and the use of multiple assessments per patient influenced study results. High heterogeneity was also found for assessments of substantial clinically important difference as well as for worsening of pain.To our knowledge, this is the first systematic review of MCIDs in acute pain. We identified 37 studies involving over 8000 patients and a broad range of clinical conditions, study approaches and pain scales. We gained access to unpublished data from 10 studies, including individual patient data from seven studies (918 patients). This ensured high data quality and uniform analysis, and enabled an adequate assessment of the association with baseline pain avoiding the risk of ecological fallacy  inherentHowever, we were not able to fully explain the large heterogeneity among studies. We found no effects of pain scale, but for the comparison of clinical conditions and directionality of transition scale, subgroups involved too few studies to ensure detection of all relevant associations. In addition, our ability to assess clinical condition was limited by the fact that many studies included a mixed patient group and we did not have access to individual patient diagnoses. Similarly, the studies included a variety of analgesia and other treatments which did not allow an assessment of potential impact of interventions. Regarding the risk of attrition bias and non-representative sampling, the majority of studies were categorised as unclear and potential impact could therefore not be assessed. Most importantly, acknowledging the association with baseline pain, it would have been more accurate to base subgroup analyses on relative outcomes, but the data available only allowed comparison of absolute outcomes. We could not assess impact of various description of pain (e.g. “intensity”) or the follow-up time between measurements since there was not enough variation between studies. Furthermore, the available data did not allow an assessment of the potential influence of pre-existing pain level (e.g. if patients are affected by chronic pain in addition to their current episode of acute pain), pre-existing use of pain relief or the psychological state of patients, since this was not reported by any of the studies. Finally, we cannot dismiss the risk of recall bias in studies where patients simultaneously assess their pain status and perceived change . DiffereOnly few systematic reviews of the minimum clinically relevant change have been published despite a vast number of primary studies. Stauffer  and ErdoThe problem of variability in results of studies of minimum clinically relevant change has previously been addressed primarily when attempting to reconcile dissimilar results from different approaches, e.g. anchor-based and distribution-based studies . Our stuWe included studies with a patient-reported anchor. While some find that using a patient-reported criterion as an anchor for a patient-reported score is circular and basically flawed , we woulThe studies we included varied considerably both in methods and analytical approaches. As would be expected, differences in the definition of patients with minimum important change impacted study results. We also found that the use of multiple measurements per patient resulted in lower outcomes. This corresponded with the finding that outcomes decreased during several follow-ups as patients’ pain declined over time. Furthermore, one in four of the reviewed studies did not differentiate between minimum relief and minimum worsening of pain in their original study reports. The practice of combining groups with minimum change, regardless of its direction, is sometimes based on a seemingly similar distribution of data in the two groups . HoweverThe association between MCID and baseline pain may to some extent be explained by ‘regression towards the mean’ as patients are likely to score closer to the mean if their initial scores were more extreme because of chance . HoweverThis review included studies that determined MCID from an anchor-based method using patients’ perception of change to determine clinical importance. Although this is the most common approach, it is just one among a wide range of alternative methods. Revicki noted that retrospective self-reports of pain relief tend to correlate more strongly with the end-level of pain than the start level, implying that the current status matters to patients more than the degree of improvement . This haIt is likely that the challenges of MCID, apparent for acute pain, may not be isolated to that specific research area. Acute pain stands out because of the many studies that have been conducted, reflecting the status of acute pain as a core symptom in clinical practice. Our study can thus be seen as a model for a more general challenge with empirical assessments of the MCID.The methodological challenges embedded in the empirical assessment of MCID are of such an extent that merits caution for its use and interpretation. It is clearly inappropriate to use and interpret MCID as a kind of clinical scale constant – a characteristic which, once empirically determined, is universally valid. This is, however, often the practice seen . NeverthThe choice of a reference value has large consequences for the number of patients needed in a trial, e.g. four times as many patients will be included, if researchers accept a MCID value of 12 mm as compared to 24 mm. Further, the conclusion about the clinical relevance of a trial result is often based on whether a mean difference exceeds a chosen reference value, but with the large span of MCIDs available in the literature, it is highly problematic to randomly pick one or a few single assessments for guidance. The considerable variation means that it is necessary to conscientiously and explicitly reflect on the span of results in relation to context-specific clinical and methodological factors, as presented in this review, with a particular focus of the baseline pain of patients, whether repeated measurements were used, and how minimum relief was defined. A starting point for such an exercise by individual clinicians or researchers, or by consensus building committees, could very well be our overview of studies and their results.In future studies there is a clear need for uniform guidelines for standardised conduct, analyses and reporting of the MCID, especially for how transition scales and questions are structured and how data are analysed. We strongly encourage using values relative to baseline pain – also for multiple measurements where the patient’s last assessment should be applied as a ‘moving baseline’, standardising the definition of relevant pain relief, and distinguishing clearly between improvement and worsening of pain. In addition, since the influence of clinical and methodological factors is difficult to identify from aggregated data, we encourage improved access to individual patient data to enable further exploration of the causes of heterogeneity.The MCID in acute pain varied greatly between studies. Absolute MCID ranged from 8 to 40 mm in 29 studies, and relative values ranged from 13% to 85% in 14 studies. Baseline pain was strongly associated with absolute, but not relative, values and variation in definitions of minimum relief and study designs influenced study results. Due to the heterogeneity between study results, no meaningful overall value of minimum clinically important change can be concluded. Instead, we recommend that MCIDs are considered context-specific and take account of baseline pain. The MCID in acute pain is central for the interpretation of results of randomised trials and meta-analyses and for determining appropriate sample sizes for new trials, but it is potentially misguiding if determined, applied or interpreted inappropriately. Explicit and conscientious reflections on the choice of a MCID value are required, when using it to classify research results as clinically important or trivial.