This comparative effectiveness study evaluates whether the effectiveness of microdecompression alone is noninferior to decompression with instrumented fusion in a real-world setting among patients with lumbar degenerative spondylolisthesis. How do outcomes compare between microdecompression and decompression with instrumented fusion for patients undergoing surgery for degenerative spondylolisthesis?In this comparative effectiveness study including 285 pairs of propensity score–matched patients, 150 of 219 patients (68%) in the microdecompression group and 155 of 215 patients (72%) in the instrumented fusion group achieved an improvement in Oswestry Disability Index of at least 30%, a clinically meaningful noninferior difference. Microdecompression was associated with shorter operation time and shorter length of hospital stay.The findings suggest that microdecompression alone should be considered as an option for most patients undergoing surgery for spinal stenosis with degenerative spondylolisthesis. Conflicting evidence and large practice variation are present in the surgical treatment of degenerative spondylolisthesis. More than 90% of surgical procedures in the United States include instrumented fusion compared with 50% or less in other countries.To evaluate whether the effectiveness of microdecompression alone is noninferior to decompression with instrumented fusion in a real-world setting.This multicenter comparative effectiveness study with a noninferiority design assessed prospective data from the Norwegian Registry for Spine Surgery. From September 19, 2007, to December 21, 2015, 1376 patients at 35 Norwegian orthopedic and neurosurgical departments underwent surgery for lumbar spinal stenosis with degenerative spondylolisthesis without scoliosis. After excluding patients undergoing laminectomy alone, fusion without instrumentation, or surgery in more than 2 levels and those with a former operation at the index level, 794 patients were included in the analyses, regardless of missing or incomplete follow-up data, before propensity score matching. Data were analyzed from March 20 to October 30, 2018.Microdecompression alone or decompression with instrumented fusion.A reduction from baseline of 30% or greater in the Oswestry Disability Index at 12-month follow-up.P < .001) and shorter hospital stay .After propensity score matching, 570 patients  were included for comparison, with 285 undergoing microdecompression  and 285 undergoing decompression with instrumented fusion . The proportion of each type of procedure varied between departments. However, changes in outcome scores varied within patients but not between departments. The proportion of patients with improvement in the Oswestry Disability Index of at least 30% was 150 of 219 (68%) in the microdecompression group and 155 of 215 (72%) in the instrumentation group. The 95% CI (–12% to 5%) for the difference of −4% was above the predefined margin of noninferiority (–15%). Microdecompression alone was associated with shorter operation time (mean [SD], 89 [44] vs 180 [65] minutes; Among patients with degenerative spondylolisthesis, the clinical effectiveness of microdecompression alone was noninferior to that of decompression with instrumented fusion. Microdecompression alone was also associated with shorter durations of surgery and hospital stay, supporting the suggestion that the less invasive procedure should be considered for most patients. The standard surgical procedure has been to decompress the stenosis.2 In the early 1990s, 2 landmark studies4 recommended additional fusion surgery. In the following decades, the rate and complexity of fusion procedures increased dramatically.6 The fusion rate in the United States more than doubled from 2005 to 2014, and degenerative spondylolisthesis accounted for most fusions.7 In 2015, the hospital costs for elective lumbar degenerative fusions exceeded $10 billion, the highest aggregate costs of any surgical procedure in the United States.8 Adding instrumented fusion to decompression has been supported by 1 randomized clinical trial (RCT)9 and clinical guidelines and reviews.13 Another RCT,14 registry studies,16 and systematic reviews18 have recommended decompression alone.Degenerative spondylolisthesis is a forward slip of one vertebra relative to the vertebra below, occurring in a spondylotic and narrowed spinal segment .21 and a large surgical practice variation between hospitals is reported. In 2011 to 2013, approximately 50% of patients with degenerative spondylolisthesis in Norway and Sweden underwent fusion procedures22 compared with 90% to 95% in other countries.23 Industrial boosting with differences in industrial encouragement and lucrative financial reimbursement might explain some of the differences in practice.25The role of fusion surgery is controversial,29 have evaluated the performance of less invasive methods of decompression alone, preserving potentially stabilizing structures of the spine. In this study from the Norwegian Registry for Spine Surgery (NORSpine), we hypothesized that in real-world daily clinical practice, microdecompression alone works as well as decompression with instrumented fusion.Only a few small-sample studiesSTROBE)30 recommendations and the International Society for Pharmacoeconomics and Outcomes Research (ISPOR) reporting guideline for cohort studies.31 Relative effectiveness was studied using prospectively collected data from NORSpine, a national comprehensive registry for quality control and research. According to NORSpine’s annual report for 2015, the coverage rate for lumbar spine surgery was 93% at the hospital level and 63% at the individual level. The registry receives no funding from industry. At hospital admission (baseline), the patients completed questionnaires, which included patient-reported outcome measures and questions about demographics and lifestyle. The surgeons recorded surgical parameters such as diagnosis, treatment, and occurrence of complications. At the 3- and 12-month follow-up, NORSpine’s central unit sent questionnaires by mail to the patients without the involvement of the surgical units. Written informed consent was obtained from the participants preoperatively, and the Norwegian Committee for Medical and Health Research Ethics Central approved the study.The reporting and interpretation of this comparative effectiveness study followed the Strengthening the Reporting of Observational Studies in Epidemiology , a procedure in more than 2 levels, or a procedure with an interspinous device or with an anterior approach. Patients were included regardless of missing or incomplete follow-up data.32 Data were analyzed from March 20 to October 30, 2018.The primary and secondary outcomes, the criterion for noninferiority, and the statistical methods were defined before statistical analysis.Patients who underwent microdecompression alone had preservation of the midline , and one of the following techniques was used: (1) unilateral laminotomy, (2) bilateral laminotomy, or (3) unilateral laminotomy and crossover decompression. Magnifying devices (microscopes or loupes) were used. Patients who underwent instrumented fusion had a decompression with or without preservation of the midline structures and with or without visual enhancement and additional posterior pedicle screw instrumentation with or without an intervertebral cage.34 is a pain-related disability score of 10 items ranging from 0 (no impairment) to 100 (maximum impairment). The primary outcome was a reduction from baseline of 30% or greater at the 12-month follow-up .36 A patient achieving this amount of improvement was defined as a responder. Secondary outcome measures included the following.The Oswestry Disability Index (ODI), version 2.0The mean change scores and the mean 12-month follow-up scores for the ODI and the Numeric Rating Scale (NRS), which ranges from 0 (no pain) to 10 (worst pain imaginable) for leg pain and for back pain experienced in the last week;37 with 7 response alternatives, including completely recovered, much improved, slightly improved, unchanged, slightly worse, much worse, and worse than ever, that were trichotomized into substantially improved (completely recovered and much improved), little or no change , and substantially deteriorated (much worse and worse than ever);The Global Perceived Effect instrumentDuration of surgery and hospital stay;The rate of perioperative complications and adverse events registered on the surgeon form; andThe rate of complications and adverse events reported by the patients at the 3-month follow-up.38 A propensity score, derived from a logistic regression model, is defined as a patient’s baseline probability for receiving decompression plus instrumented fusion, conditional on prespecified plausible confounders . We used the technique of 1:1 matching without replacement, forming paired cases of microdecompression alone and decompression plus fusion, which had a difference in propensity scores of less than 0.2 in logit of the standard deviation.38To make the distribution of observed baseline patient characteristics in the 2 treatment groups as similar as possible, propensity score matching was performed.40 This margin is consistent with analysis in other relevant studies.41 The margin corresponds to a number needed to treat of 7 patients (100/15 [6.67]),42 that is, if 7 patients or more need instrumented fusion to achieve 1 additional responder, the cheaper, safer, and less comprehensive method of microdecompression alone should be considered good enough .The null hypothesis was that the proportion of patients with a clinically important improvement (the responder rate) was at least 15 percentage points lower in the microdecompression group than in the fusion group. The null hypothesis was tested by forming a 2-sided 95% CI for the between-group difference in responder rate and would be rejected if the lower bound of the CI was greater than −15%. A noninferiority margin of −15% was assumed to reflect the advantage of performing microdecompression alone instead of the more extensive and expensive instrumentation.43 under the assumption of missing at random. Owing to nonlinearity, the models were specified as a latent difference score model, including changes from baseline to 3 months, from 3 to 12 months, and the 12-month follow-up (intercept level). The proportion of each type of procedure varied between departments. Patient data were nested within hospital departments and could show clustering effects within units. However, multilevel analyses showed low interclass correlation values for the ODI of 0.023 (baseline) to 0.042 (12 months), for NRS leg pain of 0.026 (baseline) to 0.067 (3 months), and for NRS back pain of 0.013 (3 months) to 0.066 (12 months). The estimated design effect, taking cluster size and interclass correlation into account, showed the highest value to be 2.00 (leg pain at 3 months), which is in the borderline for nonignorable clustering.44 However, multilevel models including random slope variance at the hospital department level showed no department differences in change scores in the 2 intervals.Level and change in the ODI and NRS leg and back pain scores were estimated by multisample latent growth curve (LGC) models, with full information maximum likelihood2 = 0.49; P = .03) but not for the ODI  or NRS back pain . Due to this level of clustering and the focus on observations within patients, single-level LGC models were estimated with robust standard errors corrected for clustering with the maximum likelihood robust.44The intercept variance was found to be statistically significant for NRS leg pain  was used for descriptive statistics, analyses of continuous variables with 2-sided t tests or Mann-Whitney tests depending on the distribution of data, analyses of binary variables with Fisher mid-P value and Newcombe hybrid score CIs,45 and propensity score matching. The LGC analysis was performed with Mplus 8 (Muthén & Muthén).46P < .05 indicated statistical significance.For secondary outcomes, comparisons of treatment groups and corresponding estimates of 47 A previous study from NORSpine showed similar clinical outcome measures for compliers and noncompliers,48 so longitudinal outcome variables were analyzed under the assumption of missingness at random. An additional missing data LGC analysis under the missing data at random assumption was performed for the matched cohort. Multiple imputation49 was used with baseline patient characteristics; several clinical, surgical, and radiological parameters; and outcome variables at baseline and follow-up to generate 70 data sets with complete follow-up scores for the ODI and NRS leg pain and back pain. This procedure is recommended if missing data at random may only be partly assumed.50 Including such covariates may increase the probability for missing data at random and reduce the probability of missingness not at random.50A loss to follow-up of about 20% was anticipated.15 rate at 12 months required 263 patients in each group.51For the primary outcome, choosing a 15% noninferiority margin, a type 1 error of 0.05, and power of 0.90 gave a total sample size of 394. An expected 75% follow-upOf 794 patients who met study eligible criteria, 476 (60%) underwent microdecompression alone , and 318 (40%) underwent decompression plus instrumented fusion . After propensity score matching, 570 patients  remained, of whom 285 patients underwent microdecompression alone , and 285 underwent decompression and instrumentation . Follow-up scores of the ODI and NRS leg pain and back pain are shown in eTable 2 in the The proportion of patients with a clinically important improvement in the ODI at the 12-month follow-up was 150 of 219 (68%) in the microdecompression group and 155 of 215 (72%) in the instrumented fusion group. The lower bound of the 95% CI (–12% to 5%) for the between-group difference of −4% did not cross the −15% limit of noninferiority . An absoP = .42). At 12 months, the microdecompression group had statistically significantly higher scores for NRS leg pain  and NRS back pain  compared with the instrumented fusion group.P = .32) and the rate of substantial deterioration  did not differ between groups.According to the Global Perceived Effect at the 12-month follow-up (eTable 3 in the P < .001) and the length of hospital stay  was statistically significantly shorter for microdecompression alone than for instrumented fusion (P = .003). In both groups, the most common complication was a dural tear . Patients undergoing microdecompression alone reported a statistically significantly higher incidence of superficial wound infection than the fusion group during the first 3 months postoperatively . Other registered complications are listed in The duration of surgery .eTable 4 in the Microdecompression alone was noninferior to decompression plus instrumented fusion. The result of the primary outcome was supported by the patients’ global perceived effects and by analyses of the mean ODI scores both before and after imputation of missing data. Furthermore, microdecompression alone was associated with considerably shorter duration of surgery and hospital stay and somewhat fewer surgeon-reported perioperative complications. Patients treated with instrumented fusion had slightly less leg and back pain and fewer patient-reported superficial wound infections.29 have found nondifferent outcomes between microdecompression alone and decompression plus instrumented fusion. Unlike our study, these studies did not reveal any between-group differences in outcome scores for leg or back pain.Other unmatched observational studies14 the role of fusion has been debated.53 In the RCT by Ghogawala et al,9 51 of 71 patients underwent procedures performed by 1 surgeon. Decompression alone resulted in less improvement in the physical component of the generic 36-Item Short Form Health Survey than instrumented fusion 2 years after surgery. The improvement in ODI scores was not statistically significantly different between the groups. The more pragmatic multicenter RCT by Försth et al14 (n = 135) used the ODI as the primary outcome measurement and revealed baseline scores as well as follow-up scores in accordance with our results.Following 2 simultaneously published RCTs,54 This study was designed to reflect real-world relative effectiveness between carefully matched groups. The aim was to study patients recruited in daily clinical practice at several different hospitals and the treatments chosen according to surgeon and patient preferences.55 Our study provides knowledge about how treatments work in the more pragmatic delivery of health care.57Randomized clinical trials are the criterion standard for studying treatment efficacy, but their generalizability has been questioned owing to strictly recruited patients and clinicians and enforced treatment allocation.16 and considering the large clinical practice variation,23 the high rate of instrumented fusion seems unreasonable. In accordance with a priori expectations and former studies,58 our findings of shorter operation times and hospital stays indicate that a microdecompression alone is associated with acceptable clinical results at lower costs. Although instrumented fusion was associated with somewhat more pain reduction, the high number needed to achieve 1 additional responder and the somewhat higher perioperative complication rate showed disadvantages of instrumentation. Altogether, we consider the noninferior clinical effectiveness and the potential health-economic benefits of microdecompression alone to surpass the procedure’s potential inferiority.Based on the present study as well as previous pragmatic studies39However, this study does not provide evidence that microdecompression alone should be the preferred method for all patients. Adding fusion to decompression may still be a good treatment option for subgroups. Owing to a lack of evidence for defining such subgroups, future research should endeavor to identify variables associated with the best treatment for each individual.59 were fulfilled. Incorrect diagnoses may therefore have been undetected. Furthermore, data on reoperations and data beyond the 12-month follow-up are lacking. Some studies have found lower reoperation rates when a decompression has been supported by fusion,60 whereas other studies have demonstrated similar14 and even higher27 reoperation rates after an additional fusion. For a mixed population undergoing spinal surgical procedures, the clinical outcomes at the 12-month follow-up seem to be the same as the 2-year outcomes62 and stable even at the 5-year follow-up.64The diagnoses of spinal stenosis and degenerative spondylolisthesis are based on the surgeons’ evaluation of radiographs, the radiological descriptions, and clinical signs and symptoms. We have not retrospectively checked whether all established diagnostic criteriaFinally, it is important to recognize that, unlike an RCT, this study was not able to detect treatment-related differences in efficacy. Although the propensity score matching equalized the baseline scores regarding the observed parameters, the distribution of unobserved parameters  might have been unbalanced. A risk of residual bias does therefore still exist.This study found that microdecompression alone seems to be not appreciably worse than decompression and instrumented fusion for treatment of degenerative spondylolisthesis. We would carefully suggest the less extensive and less expensive treatment as the primary surgical choice for most patients with lumbar degenerative spondylolisthesis.


As a component of multimodal analgesia, the administration of systemic lidocaine is a well-known technique. We aimed to evaluate the efficacy of lidocaine infusion on postoperative pain-related outcomes in patients undergoing totally extraperitoneal (TEP) laparoscopies inguinal hernioplasty.− 1 followed by an infusion of 2 mg. kg− 1. h− 1), or an equal volume of normal saline. The infusion was initiated just before the induction of anesthesia and discontinued after tracheal extubation. The primary outcome of the study was postoperative morphine equivalent consumption up to 24 h after surgery. Secondary outcomes included postoperative pain scores, nausea/vomiting (PONV), sedation, quality of recovery (scores based on QoR-40 questionnaire), patient satisfaction, and the incidence of chronic pain.In this randomized controlled double-blind study, we recruited 64 patients to receive either lidocaine 2% (intravenous bolus 1.5 mg. kg p < 0.001). Postoperative pain intensity at rest and during movement at various time points in the first 24 h were significantly lower in the lidocaine group compared with the saline group (p < 0.05). Fewer patients reported PONV in the lidocaine group than in the saline group (p < 0.05). Median QoR scores at 24 h after surgery were significantly better in the lidocaine group (194 (194–196) than saline group 184 (183–186) (p < 0.001). Patients receiving lidocaine were more satisfied with postoperative analgesia than those receiving saline (p = 0.02). No difference was detected in terms of postoperative sedation and chronic pain after surgery.The median (IQR) cumulative postoperative morphine equivalent consumption in the first 24 h was 0 (0–1) mg in the lidocaine group and 4 [1–8] mg in the saline group (Intraoperative lidocaine infusion for laparoscopic TEP inguinal hernioplasty reduces opioid consumption, pain intensity, PONV and improves the quality of recovery and patient satisfaction.NCT02601651. Date of registration: November 10, 2015.ClinicalTrials.gov-  On the other hand, excessive use of opioids produces several adverse effects and might delay recovery , 2. Ther effects . Besides effects –7. Addit effects .Several studies have shown that perioperative lidocaine infusion reduces postoperative pain intensity and opioid consumption, while others have found lidocaine to be ineffective . These iAlthough lidocaine infusion was effective for postoperative analgesia in open inguinal hernia surgery , its useclinicaltrials.gov (Ref No. NCT02601651). The trial was conducted according to Good Clinical Practice and the Consolidated Standards of Reporting Trials (CONSORT) guidelines.This prospective randomized double-blind clinical trial was conducted at the BP Koirala Institute of Health Sciences (BPKIHS) from December 2015 to March 2017. Ethical approval for this study (Ref No. IRC/520/015) was provided by the Institutional review committee of BPKIHS, Dharan, Nepal (Member secretary Dr. Ashish Shrestha) on 24 June 2015. Before enrollment of patients, the trial was registered by the principal investigator (AG) at Patients were screened for eligibility (AG) during the pre-anesthetic visit at the in-patient-unit, the night before surgery. Male patients aged between 18 and 65 years, of ASA physical status I–II, planned for laparoscopic TEP repair of the inguinal hernia were eligible. Patients were excluded if they were obese, unable to comprehend the pain assessment scale, allergic to local anesthetics, on pain medication or anti-arrhythmic drugs, or had, psychiatric disorders, cardiac arrhythmia, hepatorenal disease or epilepsy.After obtaining written informed consent, all eligible participants were randomly assigned, in a 1:1 ratio, to receive either lidocaine (intervention) or normal saline (placebo comparator) infusion. The anesthesia supporting staff created the trial-group assignment from the computer-based randomization list, which remained secured in sequentially numbered sealed opaque envelopes and concealed until after enrollment.− 1 lidocaine  followed by a continuous infusion of 2 mg. kg− 1. h− 1 until the tracheal extubation; The saline group received an equal volume of IV 0.9% normal saline (NS) bolus followed by a continuous infusion. Patients, attending anesthesiologists, and the investigator who collected the data and assessed the outcomes were unaware of the trial-group assignment.On the day of surgery, an anesthesia assistant not involved in the study prepared the drug solution after breaking the codes. Patients received one of the two assigned study medications just before the induction of anesthesia: Lidocaine group received an IV bolus of 1.5 mg. kg− 1 and propofol 2–2.5 mg. kg− 1 till the cessation of verbal response and the tracheal intubation was facilitated with vecuronium 0.1 mg. kg− 1 IV. The lungs were mechanically ventilated in volume control mode, maintaining the end-tidal carbon dioxide (ETCO2) between 35 and 45 mmHg.Patients received no premedication. During the pre-anesthetic visit, they were educated on the numeric pain rating scale  for postoperative pain, where 0 is no pain and 10 is the worst imaginable excruciating pain. In the operating room, standard monitoring was applied. Just before the induction of anesthesia, patients received the study drug, according to the group allocation. Anesthesia was induced with IV fentanyl 1.5 μg. kg− 1 was supplemented intraoperatively if MAP and heart rate increased by 20% from the baseline after ensuring adequate end-tidal concentration of isoflurane, neuromuscular blockade and targeted range of ETCO2. The adequate neuromuscular blockade was achieved with supplemental doses of vecuronium IV bolus after observing curare notch in capnograph. Any episode of intraoperative hypotension (MAP < 65 mmHg) and bradycardia (heart rate < 50 beats. min− 1) was treated with ephedrine 5 mg and atropine 0.4 mg IV respectively.Intravenous paracetamol 1 g was administered for 15 min after tracheal intubation. Pre-incisional infiltration in the three trocar sites was done with 2 ml of 0.25% bupivacaine. Anesthesia was maintained with an air / oxygen mixture (inspired oxygen fraction 0.40) and isoflurane, adjusting the end-tidal concentration of isoflurane to maintain mean arterial pressure (MAP) within 20% of the baseline. IV fentanyl 0.5 μg. kg− 1 and glycopyrrolate 0.01 mg. kg− 1. Following successful tracheal extubation, the study drug was discontinued and the patient was transferred to the postanesthesia care unit (PACU).An experienced surgeon performed the TEP laparoscopic surgery for inguinal hernia repair as described elsewhere . KetorolThe blinded investigator assessed the postoperative outcomes. The primary outcome was total IV morphine equivalent consumed in the first 24 h. Secondary outcomes were postoperative pain scores (NRS) at rest and on movement, sedation scores recorded using a 5-point scale  , the incwww.clincalc.com/Opioids). Ondansetron 4 mg IV was administered for persistent nausea (lasting > 5 min) or vomiting. CPSP was defined as pain that developed after a surgical procedure and persisted at least 3 months after surgery [Pain and sedation scores were assessed at PACU  and surgical unit . If the NRS score for pain was > 3 at rest, morphine 1 mg IV bolus was administered in the PACU, and repeated at 5 min interval until NRS was ≤3. After 2 h of the stay in the PACU, the patients were transferred to the ward. In the surgical unit, tramadol 50 mg IV was administered for NRS score > 3 and 50 mg was repeated at 10 min interval, up to a maximum dose of 300 mg in the first 24 h for maintaining VAS score for pain ≤3. The amount of tramadol consumed was converted to an equivalent dose of morphine from an online dose equivalent calculator . Histograms and the Shapiro-Wilk test was used to check the normality of the data. Normally distributed data were compared using a 2-tailed t-test for independent samples. Non-normally distributed data were analyzed using the Mann-Whitney U test. For ordinal data, the Kruskal-Wallis test was applied. Chi-square test or Fischer’s exact test was used for analyzing the categorical variables as appropriate. The finding with an associated p < 0.001).Of the 82 screened patients, 18 patients were excluded . In the surgical unit, patients consumed a lesser median (IQR) tramadol in the lidocaine group, 0 (0–0) mg compared with the saline group 0 (0–50) mg (p < 0.001). The median NRS scores at rest and during movement were significantly lower in the lidocaine group than in the saline group at all time points after surgery .The cumulative median IV morphine equivalent consumption at 24 h postoperatively was significantly reduced in the lidocaine group than in the saline group .A significant number of patients in the saline group had PONV and needed antiemetic compared to the lidocaine group Table . PostopeOur study showed that intraoperative infusion of low dose lidocaine decreased postoperative opioid requirement and pain intensity in comparison with normal saline in patients undergoing laparoscopic TEP inguinal hernia surgery. Patients receiving lidocaine had fewer occurrences of PONV, a better quality of recovery and were more satisfied with postoperative pain relief than those receiving saline. Patients complained of pain later in the lidocaine group than the saline group. No significant difference was observed for postoperative sedation and the incidence of chronic pain in 3 months.It is well-established that lidocaine acts on voltage-gated sodium channels when administered locally for peripheral nerve block. However, at lower concentration systemic lidocaine is insufficient to produce direct analgesia solely by blocking the neuronal sodium channels . Althougp < 0.001; I2 = 73%; 40 studies, 2201 participants). The results of our study also indicated a similar reduction in total postoperative opioid consumption in the first 24 h after surgery in the lidocaine group compared to the saline group , despite using multimodal analgesia in both the groups.Because of its influence in several pain pathways, systemic lidocaine is widely investigated adjuvant in the regimen of multimodal analgesia to reduce postoperative opioid consumption and pain. Although the majority of studies have demonstrated the analgesic effect of lidocaine, several other trials failed to confirm it. A recently updated Cochrane review in 2018 has provided a much-needed insight on the analgesic property of systemic lidocaine . Random-P < 0.0001). This was equivalent to an average pain reduction between 0.37 cm and 2.48 cm on a VAS 0 to 10 cm scale in the lidocaine group. Likewise, at intermediate time points (24 h postoperatively) the standardized mean pain score at rest in the lidocaine group was 0.14 lower . This was equivalent to an average pain reduction in the lidocaine group between 0.48 cm and 0.10 cm on a VAS 0 to 10 cm scale. These results showed that lidocaine exerted a clinical difference of at least 1 cm on a 0–10 VAS scores for pain at rest during early time points (1 to 4 h); however, this difference was not observed at intermediate (24 h) time points. We too observed statistically significant difference in pain scores up to 24 h postoperatively, while the clinical difference of approximately 1 cm in NRS scores at rest was observed only up to 1 h.Further, the aforementioned meta-analysis  demonstrDue to substantial heterogeneity between studies, the authors of the same meta-analysis performed a sub-group analysis based on type of surgery, duration and dose of lidocaine infusions . In the − 1. h− 1) and high (≥ 2 mg.kg− 1 h− 1) lidocaine doses in combination with either short (until the end of surgery or until PACU) or long (≥ 24 h postoperatively) duration of infusion [The optimal dose and time to terminate lidocaine infusion are still an unsolved issue. We had limited the duration of lidocaine infusion until the patients trachea was extubated due to a lack of dedicated infusion pumps and monitoring at the surgical unit. One might hypothesize that longer infusions would lead to more lasting analgesia but studies are yet to confirm this. The current meta-analysis (2018) had categorized the studies according to the usage of low  reported a significantly lower frequency of nausea in the lidocaine group than in the control group, but the vomiting rates did not differ . Althoug2 = 0%; 6 studies, 306 participants) [Recently, there is a growing interest in patient-reported outcomes such as postoperative QoR and patient satisfaction. We observed better recovery profiles at 24 h of surgery in the lidocaine group as evident from the QoR scores. Similar to our study, De Oliveira and his colleagues reported greater QoR-40 scores at 24 h with perioperative lidocaine infusion for laparoscopic abdominal surgery . Likewiscipants) . Furthercipants) . As our In summary, intraoperative lidocaine infusion decreases overall opioid requirement and postoperative pain intensity in patients undergoing laparoscopic TEP inguinal hernioplasty. It also lowers the incidence of PONV, improves the quality of recovery and patients satisfaction without any sedative effect.


Focus on the frequency of migraine pain may undervalue the total burden of migraine as pain duration and severity may present unique, additive burden. A composite measure of total pain burden  may provide a more comprehensive characterization of pain burden and treatment response in patients with episodic migraine (EM) or chronic migraine (CM). The impact of galcanezumab versus placebo on TPB among patients with EM or CM was analyzed.N = 444 galcanezumab, N = 894 placebo) and Months 1–3  were compared using a mixed-model repeated measures model. Correlation of the Migraine Specific Quality of Life Questionnaire (MSQ) and Migraine Disability Assessment Scale (MIDAS) to TPB at baseline was assessed.Patients from randomized, double-blind, placebo-controlled episodic (two 6-month studies pooled) and chronic migraine (3-month) studies received once-monthly subcutaneous injection of galcanezumab 120 mg or placebo. A post hoc analysis of TPB for a given month was calculated as severity-weighted duration by multiplying duration (hours) and maximum pain severity  of migraine for each day and summing these over the days in a month. Least square mean change from baseline in monthly TPB across Months 1–6  and MIDAS (r = 0.34 and r = 0.32).At baseline, the duration of migraine on a given migraine headache day accounted for the greatest unique proportion of variability  to TPB after adjusting for frequency of migraine headache days and maximum pain severity. The decrease from baseline in monthly TPB was greater with galcanezumab than placebo for patients with EM (68.6 versus 36.2) and CM (102.6 versus 44.4). The average percent reduction of TPB from baseline was significantly greater with galcanezumab compared with placebo in patients with EM (50.8% versus 17.2%) and CM (29.7% versus 11.0%). In patients with EM and CM, TPB correlated with MSQ total score , #NCT02614196 , and #NCT02614261  – all 3 trials were registered on 23 November 2015. It is well established that migraine can interfere with the occupational, educational, household, family, and social facets of daily life –10. ConsGalcanezumab is a humanized IgG4 monoclonal antibody that binds calcitonin gene-related peptide (CGRP) and prevents its biological activity without blocking the CGRP receptor and is indicated for the preventive treatment of migraine . The effThe efficacy of galcanezumab for reduction of migraine frequency is established. However, it is unclear whether galcanezumab impacts total pain burden of migraine over time. Indication of this would provide greater clarity to providers regarding the overall potential clinical value of galcanezumab. In this post hoc study, total pain burden was compared between those taking galcanezumab 120 mg once-monthly injection  and placebo among patients with episodic or chronic migraine.This post hoc analysis of Phase 3, randomized, double-blind, placebo-controlled studies in adult patients utilized data from two, 6-month episodic migraine studies pooled (EVOLVE-1 and EVOLVE-2) and one 3-month chronic migraine study (REGAIN) to analyze the impact of galcanezumab 120 mg once-monthly injection relative to placebo on total pain burden among patients with episodic migraine or chronic migraine . PatientClinicalTrials.gov .In all studies, patients entered a prospective baseline period of 30 to 40 days in which they completed an electronic patient-reported outcome diary to record the occurrence of headaches, headache duration, headache features, severity of headache, and use of headache medication. Eligible patients were randomized (1:1:2) to subcutaneous injections of galcanezumab 120 mg/month , galcanezumab 240 mg/month, or placebo. For the episodic migraine studies, patients were to have discontinued the use of medication or other treatments for the prevention of migraine for at least 30 days, and use of botulinum toxin A and B for at least 4 months, prior to the prospective baseline period. For the chronic migraine trial, patients were allowed to continue using topiramate or propranolol if they were on a stable dose in the 2 months prior to the prospective baseline period and remained on that dose throughout the baseline and double-blind periods. During all three studies, patients were permitted to continue acute migraine medications including triptans, ergots, nonsteroidal anti-inflammatory drugs, aspirin, and acetaminophen. Patients were permitted to continue acute migraine medications including triptans, ergots, nonsteroidal anti-inflammatory drugs, aspirin, and acetaminophen. Opioid- and barbiturate-containing medications were limited to 3 days/month, and only one corticosteroid injection was allowed during any period. Exclusion criteria of note for all three studies included, but was not limited to, prior exposure to any CGRP antibody or any therapeutic antibody 12 months prior to screening, using opioids or barbiturates more than twice per month, persistent daily headache, cluster headache, head or neck trauma within the past 6 months, possible posttraumatic headache, primary headache, or a medical or psychiatric illness that would preclude study participation . The stufrequency of migraine headache days in a month, duration of migraine headache on a given day, and maximum severity of migraine headache on a given day. The total pain burden for a given month (severity-weighted duration) was calculated by multiplying duration (hours) of migraine headache and maximum pain severity  for each migraine headache day and summing these over the days in a month. As an example, consider a patient who has 2 days of migraine headache in a month. The patient reports 2 h of migraine headache on Day 1, which is of mild severity (score = 1) and 3 h of migraine headache on Day 2, which is of moderate severity (score = 2). The total pain burden score for that month would be calculated as the sum of (2 h × 1) and (3 h × 2) which equals 8 severity-weighted hours of total pain burden. The change from baseline in monthly total pain burden measure over the double-blind period was analyzed for both episodic (6 months) and chronic migraine (3 months) studies using a mixed-model repeated measures (MMRM) model. The analysis of the episodic migraine studies included the following fixed effect variables: baseline, treatment, month, study indicator (EVOLVE-1 or EVOLVE-2), pooled region/country (nested within study), and the treatment by month, and baseline by month interaction effects. For the chronic migraine study, the fixed effect variables included: baseline, treatment, month, pooled country, baseline medication overuse (yes/no), concurrent prophylaxis use (yes/no), the interaction effects of treatment by month, and baseline by month. A marginal unstructured covariance structure was assumed to account for the correlation induced due to repeated measures on patients. The mean percentage change in total pain burden was analyzed using an MMRM model, as described earlier, for the change from baseline analyses. A similar MMRM modeling approach was also used to analyze change from baseline in each of the individual components of frequency of migraine headache days, duration per migraine headache day, and severity.The primary outcome of the EVOLVE-1, EVOLVE-2, and REGAIN studies was the overall mean change from baseline in the number of monthly migraine headache days during the double-blind treatment phase. This post hoc analysis evaluated the galcanezumab 120 mg and placebo treatment arms of the 3 studies. The objective of these analyses was to derive and compare between treatment groups, a “total pain burden” measure that incorporates: The Type II squared semi-partial correlation of each individual component within the total pain burden measure at baseline was obtained from the following regression model on the log transformed variables: log  = log (monthly migraine headache days), log (mean hours per migraine headache day), and log (severity of remaining migraine headache days). These provide the additional or unique proportion of variability explained by each component in its ability to predict total pain burden after adjusting for the other two components.To explore the construct validity of the total pain burden measure, the Spearman’s rank correlation was used to determine the degree of correlation of the Migraine Specific Quality of Life Questionnaire (MSQ) and Migraine Disability Assessment Scale (MIDAS) to total pain burden at baseline. The MSQ is a self-administered instrument that evaluates the physical and emotional limitations of specific concern to patients with migraine . The MIDThe pain severity scores  can be thought of as arbitrary. Therefore, a sensitivity analysis was performed by calculating (and comparing between groups) the total pain burden score using the square root of severity score in the calculation instead of severity  and using the square of the severity score in the calculation instead of severity . To further understand whether the changes observed in total pain burden can be fully explained by the change observed in frequency of migraine headache days, the original MMRM model analyzing change from baseline in total pain burden was repeated after including an additional time-varying covariate of the change from baseline in the number of monthly migraine headache days.® Enterprise Guide version 7.1. All analyses done were post hoc and results should be considered exploratory.Only patients in the intent-to-treat (ITT) population were considered for inclusion in this post-hoc analysis. Baseline demographics and descriptive summaries were provided for continuous variables using means and standard deviation and/or median and quartiles; for categorical variables, frequency and percentages were used. Treatment effects from MMRM models were expressed as least squares (LS) means and 95% confidence intervals (CI). No specific methods, such as multiple imputation, were employed to handle missing data. Furthermore, all repeated measures models included patients that had data at baseline and at least one time point during the double-blind phase. All statistical tests done were two-sided assuming a significance level of 5%. All statistical analyses were done using SASN = 894, galcanezumab 120 mg N = 444) and the chronic migraine REGAIN study  were included in this post hoc analysis. The general baseline demographics and characteristics were balanced between treatment groups in the two episodic and one chronic migraine studies and have been previously well-described [Only patients in the ITT population from the placebo and the galcanezumab 120 mg arms of the pooled episodic EVOLVE-1 and EVOLVE-2 studies  (data not shown). The overall monthly LS mean severity-weighted hours reduction from baseline across Months 1 to 6 for episodic migraine and across Months 1 to 3 for chronic migraine were greater for the galcanezumab groups compared with the placebo groups . Likewise, for patients with chronic migraine, the average percent reduction across 3 months was greater with galcanezumab (29.7%) compared with placebo (11.0%); mean difference was 18.7 . As illustrated in Supplemental Material p < 0.001) and the significantly greater percentage reduction was maintained at each subsequent month (p < 0.001).The average percent reduction from baseline in total pain burden in patients with episodic migraine across 6 months was greater with galcanezumab treatment (50.8%) compared with placebo 17.2%); mean difference was 33.6%  of migraine pain on the days when a patient had a migraine headache in a month contributed the most variability to total pain burden rather than the number of days with migraine headache in a month.The duration of migraine on a given migraine headache day accounted for the greatest unique proportion of variability  to the baseline total pain burden after adjusting for the frequency of migraine headache days and maximum pain severity Table . In otheResults from the sensitivity analyses related to pain severity scaling revealed that total pain burden remained significantly lower for the galcanezumab group compared with placebo even when using the square root of severity scores or the square of severity scores in the calculation of total pain burden. Thus, altering the severity rating scores on the calculation of total pain burden did not considerably alter the findings.r = − 0.35 and r = 0.34, respectively; p < 0.0001) and chronic migraine  populations.The MSQ and MIDAS scores at baseline were associated to total pain burden in the episodic  and supports findings from the analysis of the individual components of total pain burden Table . Shown iTreatment with galcanezumab 120 mg once-monthly injection  resulted in greater reduction of total pain burden relative to placebo. This study, combined with previous studies demonstrating the benefit of galcanezumab for reducing monthly migraine headache days, provides a more holistic view of the effect galcanezumab can have in patients with episodic or chronic migraine.A composite measure of total pain burden produces a measure that may be more aligned to the personal pain experience in migraine. The almost exclusive focus on pain frequency as the metric for evaluating treatment “benefit” emanated from a desire articulated earlier this century for a standard, unified method for assessing treatment benefit . In doinThe evidence presented here suggests that galcanezumab could benefit a patient beyond reduction in migraine headache days per month. This observation may be especially relevant for those with chronic migraine as these patients will likely to continue to experience migraine headache more frequently than those with episodic migraine. If those remaining migraine headache days feature pain that lasts fewer hours and/or is less severe, then patients may be experiencing benefit that goes beyond those identified through unidimensional frequency counts.The results support the hypothesis that total pain burden is a composite of various components that contribute unique information. Further, total pain burden is not simply a proxy to the number of migraine headache days since the reduction of total pain burden was greater with galcanezumab relative to placebo even after accounting for the reduction in monthly migraine headache days. This implies that the patient is experiencing an added benefit beyond experiencing fewer migraine headache days per month. Moreover, the positive correlation of total pain burden with MSQ and MIDAS further supports the construct validity of this composite measure and that disability is significantly correlated with total pain burden.This study provides data regarding total pain burden in which published research pertaining to migraine is sparse. The outcomes from this study are limited by the post hoc nature of the evaluation and the clinical trial setting that restricts generalizability to the real-world population of patients with migraine. Another limitation of note may be the 4-point scale used to assess severity and its use in calculating the total pain burden given that a 4-point scale may have less sensitivity in detecting changes in severity than an 11-point scale. Finally, the scope of the analysis did not fully account for the influence of acute medication use on total pain burden.The study outcomes are strengthened by the large sample size for both the populations of patients with episodic and chronic migraine and by the controlled nature of the study that allows for comparisons between groups providing greater confidence in considering the veracity of observed outcomes. Additionally, in the current analysis of total pain burden, we calculated severity-weighted hours such that 1 h of moderate pain was considered equally burdensome as 2 h of mild pain . Others may consider different weighting algorithms to be more representative of total pain burden. However, our sensitivity analysis showed that changes in severity weights did not significantly alter the results when assessing the impact of galcanezumab on total pain burden.Greater reduction in total pain burden was seen in patients with episodic or chronic migraine treated with galcanezumab 120 mg once-monthly injection relative to placebo. This allows for a more robust evaluation of the potential benefit of preventive treatments  in reducing the overall pain experience. Moreover, the discussion of total pain burden is well situated to support patient-centric discussions regarding treatment expectations when clinicians are evaluating options for migraine prevention.Additional file 1.The video animations for the episodic migraine studies (Animation 1) and for the chronic migraine study (Animation 2) tracks the change in monthly total pain burden as a percent of baseline (y-axis) for each patient in the placebo  and the galcanezumab 120 mg  group. The x-axis shows individual patients rank ordered by their final percent of baseline value at Month 6 for episodic migraine studies (Month 3 for chronic migraine study), ordered from greatest improvement on the left to greatest worsening on the right within each plot). Patients with missing data during the last month are placed towards the upper end of the x-axis. All patients start at a baseline of 100%. The solid purple line represents the mean percent of baseline in total pain burden within each treatment group at any point in time. In each scatterplot, black dotted reference lines are shown for baseline (100%) and 50% reduction from baseline in total pain burden. Any patient with a percent of baseline value below 100% has shown an improvement in total pain burden from baseline while patients with values below the 50% reference line have shown at least a 50% reduction in total pain burden compared to their baseline score. The bar plot in the middle shows the observed mean change in total pain burden from baseline to the end of each month for each treatment group . The timeline bar across the top tracks the month of treatment being shown, from baseline (Month 0) to Month 6 for episodic (Month 3 for chronic) in the double-blind treatment phase.


IntroductionThe association of pain and discomfort of moderate to high severity and a high incidence with the intravenous (IV) administration of propofol is well known. Various physical and pharmacological methods are used to minimize propofol-induced pain, but the best intervention is still unknown. Therefore, our aim was to determine the analgesic efficacy of IV paracetamol when used in combination with lidocaine pretreatment in reducing propofol injection pain.Materials and methodsThis double-blind, randomized controlled trial was conducted after receiving the approval of our institutional research ethics board. A total of 74 patients were included after providing informed consent, and participants were placed into two equal groups: group A received IV paracetamol (1 g) in combination with lidocaine pretreatment prior to the injection of propofol, and group B received lidocaine pretreatment alone prior to propofol injection. After propofol injection, all participants were asked to evaluate pain on the visual analog scale.ResultsPatients who received the lidocaine-paracetamol combination reported significantly more pain-free responses (51.35%) than those from patients who received lidocaine pretreatment alone . The analgesic efficacy of group A was positive in 36 patients (97.3%), and for group B, the analgesic efficacy was positive in 24 patients (64.9%).ConclusionThe administration of IV paracetamol with lidocaine pretreatment was more effective than lidocaine pretreatment alone in reducing the pain caused by the injection of propofol. Physicians should consider using IV paracetamol in combination with lidocaine pretreatment when patients require IV propofol to ease patient suffering and reduce pain, which may help provide optimal patient care. Propofol is a popular intravenous (IV) anesthetic induction agent that causes pain during administration. The incidence of propofol-induced pain is 28% to 90% in patients receiving propofol alone with no other intervention or medication . The exaParacetamol is a synthetic, non-opiate, centrally acting analgesic derived from p-aminophenol. It is widely used for pain management preoperatively as co-analgesia ,7. The aThis double-blind, randomized controlled trial was conducted in the operating rooms of a tertiary care center after approval from the institutional ethical review committee. The study included patients aged 18 to 60 years and in the American Society of Anesthesiologists (ASA) status I and ASA II, who were admitted for elective surgery under general anesthesia after providing informed consent to participate. Patients who provided consent were selected via a nonprobability consecutive sampling technique. Any patients taking other analgesics preoperatively, those with known allergies to any drug used in the study, those with deranged hepatic functions, those with ASA III or higher status, those with a language barrier, those unable to understand the visual analog scale (VAS), and any pregnant women were excluded from the study. A unique randomization number was allotted to each participant, who was randomly assigned into two equal groups of 34 patients. Group A patients received IV paracetamol (1 g) with lidocaine pretreatment, and group B served as the control group and received lidocaine pretreatment alone. The study used 74 similar looking bottles labeled as “study drug” with the randomization number for masking the contents from the primary investigators and the patients. The enrolled patients were taught how to use the 100-point VAS to rate their pain levels. All patients were premedicated with midazolam one hour before surgery. In the preoperative area, after standard monitoring, the study drug  was infused over 15 to 20 minutes under the observation of the primary anesthetist. Afterward, the patients were moved to the operating room. After a pneumatic tourniquet was applied, the vein which would be used for propofol injection was pretreated with 2 mL of 2% lidocaine. After one minute of pretreatment, the tourniquet was released and one-fourth of the calculated dose of propofol was injected intravenously over five seconds by an electronic pump. After injection, the patient was asked whether he or she felt pain, and the pain score was marked according to the 100-point VAS. Pain severity levels were defined as VAS 0 for no pain, VAS 1 to 30 for mild pain, VAS 31 to 70 for moderate pain, and VAS >70 for severe pain. Following that measurement, the complete anesthetic dose of propofol was administered and anesthesia continued as per the plan.Data analysisAll statistical procedures were performed using IBM SPSS Statistics for Windows, Version 19.0. . We computed the mean and standard deviation for all quantitative variables including age, weight, and height, while qualitative data like sex, pain, and efficacy of paracetamol were calculated in frequency and percentage. Chi-square and Fisher’s exact tests were applied to compare the proportion of pain between the two groups. P-value <0.05 was considered to be significant. Stratification was done to control effect modifiers like age, sex, height, and weight to observe an outcome (pain).A total of 74 patients were enrolled in the study (36 men [48.6%] and 38 women 51.35%]). The study population age, weight, height, and hemodynamic characteristics are presented in Table .35%). ThThe incidence of pain in the control group (group B) was higher than group A. In group B, 34 patients (91.89%) reported pain on injection, while only 18 patients (48.64%) in group A reported pain on injection. For pain severity, four patients in group B (10.81%) and no patients in group A reported severe pain on injection. One patient in group A 2.7%) reported moderate pain, while nine patients in group B (24.32%) reported moderate pain. A total of 17 patients in group A (45.95%) reported mild pain and 21 patients in group B (56.76%) reported mild pain. Nineteen patients in group A reported no pain on injection (51.35%), while only three patients in group B (8.11%) reported no pain on injection. The assessment of analgesic efficacy, defined as positive (“YES”) for VAS score of 0 to 30 and negative (“NO”) for VAS scores of 31 to 100 for each group is shown in Table .7% reporNo untoward effects related to paracetamol were found in patients in group A. There was no significant difference in analgesic efficacy in groups A and B with respect to sex and height (P<0.05). Analgesic efficacy results in each group with respect to age and weight are described in Table Propofol has a fast onset, smooth anesthesia induction, favorable effect on upper airway reflexes, a smooth and rapid recovery, and minimal association with postoperative nausea and vomiting, making it a popular IV anesthetic induction agent ,11. HoweCanbay et al. studied the effect of 50 mg paracetamol pretreatment on non-premedicated  patients and concluded that paracetamol pretreatment was an effective intervention to reduce propofol injection pain . Ozkan eDedic et al. studied the effect of oral paracetamol (1 g) on prevention of propofol pain using the VAS scoring system; they found that no patients reported severe pain, 75 (72.81%) reported no pain, and 28 (27.19%) reported mild, bearable pain on propofol injection . HoweverThe strength of this study is that it is a randomized, double-blind controlled trial, and all the cases were performed by one anesthetist to eliminate observer bias. Also, it was a therapeutic study in which patients benefited. To our knowledge and according to the available literature, this study is the first of its kind in our population. One potential weakness of the study is the sample size; a larger size may have added further support to our findings.The administration of IV paracetamol in combination with lidocaine for pretreatment before propofol injection is a superior method to lidocaine pretreatment alone in reducing pain caused by propofol injection. For patients who require propofol during their care, physicians should consider using IV paracetamol in combination with lidocaine pretreatment, when appropriate, to ease patient suffering and reduce pain, which may help provide optimal patient care.


To investigate if intra-arterial lidocaine administrated immediately after the embolisation endpoint reduces the pain.Forty patients were randomised and 36 completed the study for purposes of analysis. In one group, the patients got 1% 10 ml lidocaine (100 mg) administered into each uterine artery immediately after embolisation with microspheres. The other group was embolised without supplementary lidocaine. The patients scored their pain on a visual analogue scale (VAS) 2 h, 4 h, 7 h, 10 h and 24 h after embolisation, and the total amount of used morphine was noted. Three-month follow-up MRI control was scheduled for all the patients to investigate the infarction rate.p < 0.02). There was no significant difference in pain score 4 h, 7 h, 10 h and 24 h after UFE. In the lidocaine group, the mean amount of used morphine was significantly less with 11.2 mg compared with 20.2 mg in the control group (p < 0.03). Three months of MR follow-up control showed no significant difference in the grade of fibroid infarction.Embolisation was performed without any complications and with embolisation of both uterine arteries in all cases. Intra-arterial lidocaine was administered in all 20 patients without complications, and 20 patients in a control group did not receive lidocaine intra-arterial. VAS schemes showed a significant reduction in pain experience 2 h after UFE where mean pain score in the lidocaine group was 42.7 ± 21.4 compared with the control group in which the mean pain score was 61.1 ± 20.4 (Intra-arterial Lidocaine administration after embolisation is safe and effective in reducing post-procedural pain in the early hours and opioid usage in the first 24 h following UAE. Uterine fibroid embolisation (UFE) is a well-described and well-established interventional radiology procedure to treat symptomatic uterine fibroids. The efficacy and long term outcome is documented in several randomised studies and long-term follow-up results are confirmed . Between August 2017 and July 2019, all patients with symptomatic uterine fibroids who accepted participation were block randomised were smaller blocks of fewer patients randomised to the two treatments at a time intending to have 20 patients in each group. The patients in one group had 1% 10 ml lidocaine (100 mg) administrated intra-arterially immediately after embolisation in each uterine artery, so the total doses given to each patient was 200 mg of lidocaine. In the other (control) group of patients, the procedure was performed according to standard principles without supplementary injection of lidocaine. The patient’s age and symptoms, fibroid number, localisation and total uterine and dominant fibroid volume changes before and after an embolization were analysed.The standard embolisation technique was performed in local anaesthesia via either transfemoral or transradial access using diagnostic 5F catheter advanced into the internal iliac artery. Further, in all cases, micro-catheter  was used and advanced into the horizontal part of the uterine arteries. Embolisation with microspheres tris-acryl gelatin (500–900 μm)  to near stasis defined as slow forward flow trough the main uterine artery with at least five heartbeats for clearance of contrast from uterine artery with the pruning of peripheral vessels. The total amount of the used microspheres was recorded and compared between the groups. All patients had patient-controlled analgesia pump (PCA) loaded with 2 mg morphine sulphate and an intermittent dose of 1 mg morphine with lockout intervals of 10 min. Before the beginning of the embolisation, a loading dose of 2 mg morphine was given to all patients. The patients were instructed to push the button of the PCA for the first time before the pain developed. No antibiotics were given and no urinary catheter was deployed. There was no standardised regimen concerning usage of adjunctive analgesia and sedation, and if needed, some patients take additional drugs such as non-steroid anti-inflammatory medications on the ward individually.Inclusion criteria were symptomatic fibroids in a premenopausal woman with bleeding and/or bulky symptoms. Size and number of fibroids were not exclusion criteria. Patients who did not want to participate in the study were excluded. Allergy to morphine or lidocaine, heart block and active pelvic infection were exclusion criteria. After embolisation, the patients were transferred to the ward and a visual analogue pain scale (VAS) with oral and written instructions was given to the patients  program was used for analysis.Descriptive statistics were used for baseline patients and fibroid characteristics and were presented as number and percentage and as mean and standard deviation (SD). The student’s t-test was used for comparison between the groups regarding the used amount of morphine and pain experience. p < 0.78), fibroid numbers (p < 0.89), fibroid location (p < 0.22), dominant symptoms (p < 0.54), or amount of used microspheres p < (0.19). Intra-arterial lidocaine was administrated in all 20 patients without complications and was well tolerated. VAS schemes showed a significant difference in pain experienced between the groups 2 h after UFE. The median pain experienced in the lidocaine group was 45 ± 21.1 compared with the control group 60.1 ± 15.2 (p < 0.02) . The total amount of used morphine was recorded 24 h after UFE. In the lidocaine group, the mean amount of used morphine was 11.2 mg compared with 20.2 mg in the control group (p < 0.03). The detailed scores of the pain experience and comparison between the groups and the use of morphine are presented in Table Fifty-four consecutive patients were treated with embolisation in the period August 2017 –July 2019 at the Department of Radiology. Fourteen patients did not want to participate in the study and were excluded. There were no patients with allergy to lidocaine or morphine or with heart problems. Thus, in all 40 patients were included in the study and all gave written consent. In all patients, UFE was performed without complications and with embolisation of both uterine arteries as intended. The demography of the patients, baseline fibroid description and patient symptoms are presented in Table The present randomised study confirmed that intra-arterial lidocaine administration after UFE reduce the pain in the first hours after embolisation. Lidocaine has a half-life about 90–120 min, and the effect was confirmed in this study during the first 2 hours after UFE. Noel-Lamy et al. Further, Katsumori et al. The pain experienced in our study was lower compared with the control group after 4 h,7 h,10 h, and 24 h but without statistical significance, similar to Katsumori et al. The intra-arterial lidocaine administration was safe in all patients without any complaints and adverse events and was easy and safe to deliver after the embolisation endpoint was achieved. In previously published studies, there were no reported cases of adverse reactions during the intraarterial lidocaine administration as well (Noel-Lamy et al. The limitation of this study are small number of the patients, not double-blinded design and drop-out of four patients that withdrew or did not fill in the VAS schema properly. Further, different access via transfemoral or transradial access could be a possible confounding factor due to early mobilisation after transradial access and eventualy use of additional non-steroid anti-inflammatory drugs in the immediately post-embolisation period. Finally, the specific size and volume of used embosphere was not standardised or analysed systematically.Intra-arterial Lidocaine administration after embolisation is safe and effective in reducing post-procedural pain in the early hours and opioid usage in the first 24 h following UAE. Further studies of how further to reduce pain are warranted.


Spinal pain is the leading cause of patient-years lived with chronic pain and disability worldwide. Although opioids are well documented as an effective short-term pain-relieving medication, more than a few weeks of treatment may result in a diminishing clinical effect as well as the development of addictive behavior. Despite recognition of opioid addiction in pain patients as a major problem commonly experienced in the clinic, no reference material exists on the scope of long-term problems in novel opioid users and the link to clinical outcomes.The main aims of this study are to describe baseline and follow-up characteristics of the Spinal Pain Opioid Cohort (SPOC), to evaluate the general use of opioids in spinal pain when an acute pain episode occurs, and to demonstrate the prevalence of long-term opioid therapy (LTOT).Prospective clinical registry data were collected from an outpatient spine center setting during 2012-2013 including patients with a new spinal pain episode lasting for more than 2 months, aged between 18 and 65 years who had their first outpatient visit in the center. Variables include demographics, clinical data collected in SpineData, the Danish National Patient Register, and The Danish National Prescription Registry. The primary outcome parameter is long-term prescription opioid use registered from 4 years before the first spine center visit to 5 years after.This is an ongoing survey. It is estimated that more than 8000 patients fulfill the SPOC inclusion criteria. In 2019, we began the intellectual process of identifying the most relevant supplementary data available from the wide range of existing national registries available in Denmark. We have now begun merging SpineData with relevant opioid data from Danish national registers and will continue to extract data up to 2021-2022. We will also be looking at data regarding somatic or psychiatric hospitalization patterns, patient usage of health care resources, as well as their working status and disability pensions.To our knowledge, this survey will be the first to document the scope of long-term problems regarding LTOT and opioid addiction following new spinal pain episodes and comparing descriptive follow-up data between substance users and nonusers.ISRCTN Registry ISRCTN69685117; http://www.isrctn.com/ISRCTN69685117DERR1-10.2196/21380 Spinal pain is the leading worldwide cause of patient-years lived with chronic pain and disability ,2. The iDespite the increasing long-term dependence risk of change in addictive behavior associated with opioid treatment , most puThe aim of this prospective Spinal Pain Opioid Cohort (SPOC) research program is to collate relevant individual patient data over a decade to illuminate both the overall group data developments as well as developments relating to individual usage of different opioids during this period, and to correlate these findings with the individual patient’s physical, psychological, and social data over 10 years. In Denmark, a large number of different national registers are available, and the present data will provide the opportunity to assess different long-term effects of opioid use.This is a longitudinal cohort study based on prospectively collected data as part of routine daily clinical practice in The Spine Center of Southern Denmark during 2012-2013. This outpatient secondary care department is a unit in a public hospital with a geographic catchment area of approximately 1.2 million inhabitants [Initial data are collected in the Spine Center’s electronic clinical registry, named the SpineData database, at the date of the first visit  [Patients aged between 18 and 65 years who had a new pain episode for more than 2 months and had their first outpatient visit at the Spine Center between November 1, 2012 and December 31, 2013, with no indications for acute spinal surgery intervention, are enrolled in the SPOC. Patient symptoms indicating serious systemic diseases such as cancer, infections, or inflammatory spondylopathies are excluded a priori. An individual patient is included in the SPOC if the baseline questionnaires has no missing data related to pain intensity. In accordance with the biopsychosocial model of health, information is collected across a set of broad health domains of pain, activity limitation, work participation, psychological factors, physical impairment, and contextual factors. Wherever possible, the choice of questions and questionnaires is based on evidence of their role in the diagnosis, prognosis, or treatment of spinal pain. The questions vary across the three spinal regions of principal complaint: neck pain, midback pain, and lower back pain .Patient-reported questions include a main pain chart (current pain) and other pain charts , the onset date of pain, any previous lower back pain or radiating episodes, pain intensity , extremity pain intensity , number of days per week with pain, cause of or reason for onset, morning stiffness, diurnal variation, movement-related pain, activity-related pain, the effect of physical rest on pain, and pain easily aggravated by movement .Patient-reported questions include the 23-item Roland-Morris Disability Questionnaire for lower back pain ,23 and tPatient-reported questions include type of employment, whether on sick leave due to back pain at any time in the last 3 months and for how long, expectation of working in 6 months, physically strenuous work, monotonous work, and work satisfaction .Levels of depressive symptoms and anxiety are measured with the Hospital Anxiety and Depression Scale (HADS) . The HADExposure to traumatic events is assessed by a modified version of the Life Events Checklist-5 (LEC-5) . Among aThe International Trauma Questionnaire (ITQ)  can be uAttachment security is measured on the Experiences in Close Relationship Scale-Short Form (ECR-S) . The ECRPatient-reported questions include height, weight, previous back surgery, prolonged corticosteroid use, exposure to prolonged mechanical vibration, handedness, level of recreational physical activity, allergies, cigarette use, alcohol consumption, serious lung disease, heart disease, or cancer, which are generated from SpineData .Quality of life is measured using a validated and recognized Danish self-perceived general health scale inspired by the EuroQoL Health Measure Thermometer (0-100) ,29.All patients are invited to complete two electronically provided follow-up questionnaires that contain representative questions from the baseline patient questionnaires. The first email follow-up questionnaire takes place 12 months after the date of the initial consultation. The second email-based follow-up questionnaires are sent to the patient as a 5-year follow up. All of the patients received an electronic questionnaire and have been asked to complete the questionnaire and return it via email. If a patient did not respond, they were contacted 2 more times after the first email communication.Linking data from the SPOC with data from the Danish national registers is possible using the Danish Identity Number (known as the CPR number) assigned to all citizens in Denmark at birth or immigration. We link the SPOC with information from the Danish Civil Registration System (CRS) , Danish The DNPR contains information on all inpatient and outpatient visits as well as emergency room visits in Denmark. From the DNPR we extract information on the patient’s comorbidities including 5 years ending on the date of the first visit in the Spine Center and using the diagnostic codes according to the disease-specific ICD-10 . Data inThe study also provides an opportunity to link the data to the many different national registers available in Denmark , and theBased on the diagnostic codes (ICD-10) from DNPR, we calculate comorbidity among the individuals at their first visit to the Spine Center. We use the Charlson Comorbidity Index to classify comorbidity among the individuals, which is based on 19 comorbid conditions ,35.The prescribed and dispensed medications from the NPR are classified according to the Anatomical Therapeutic Chemical (ATC) classification system . We extrThe Spine Center referral guidelines generally state that patients should not be referred to the center before having experienced pain for at least 2 months (mean 4.5 months) and for no longer than 12 months since symptom onset. As a consequence, for the present survey, we presume that the most intense pain period in the referred patients typically will begin and be concluded in the 364-day period minus 182 days (IndexDate) and up to 182 days following the day of the first visit in the Spine Center. In other words, the definition of the IndexDate is the date of the first visit to the Spine Center minus 182 days.From the IndexDate, we define a total of 9 observation intervals for a single patient, including 4 intervals before and 5 intervals after the IndexDate. Each interval consists of a period of 364 days. The numbers of dispensed prescriptions are counted in each interval.For most of the analyses, patients in the SPOC will be separated into the following 3 subgroups: (1) patients who do not receive any opioid prescriptions at any observation point (nonopioid users); (2) patients who have their very first opioid prescription following IndexDate (NaiveStarters); and patients who had one or more opioid prescriptions in the intervals before IndexDate (PreStarters).The separation between PreStarters and NaiveStarters allows us to differentiate patients with former spine pain episodes and related use of opioids before the IndexDate .This study’s primary outcome parameter was inspired by “The Copenhagen Criteria” definitions and methods to register LTOT . In our P<.05. Analyses are performed using STATA version 14.2 .Baseline, and 1- and 5-year follow-up characteristics of the SPOC will be reported either as proportions or median values with IQR. We use the median and IQR for continuous variables that are not normally distributed. Differences between the level of comorbidity according to the Charlson comorbidity index ,35 are aEthics approval for the collection and use of these data for quality assurance and research purposes has been provided by the Scientific Ethics Committee of the Region of Southern Denmark (project ID S-200112000-29). The database is also registered with the Danish Data Protection Agency (2008-58-0035). All patients have been invited to give two types of written informed consent. The first is for their patient data to be used for quality assurance and research purposes, including publications of anonymized group-level data, and the second is for the Spine Center to contact them requesting the completion of follow-up questionnaires.The project is ongoing. The gathering of questionnaire data among SPOC patients began in 2012. From 2013 to 2019, 1-year and 5-year follow-up data were collected and stored in the Spine Center’s SpineData database. It is estimated that more than 8000 patients fulfill the SPOC inclusion criteria. In 2019, we began the intellectual process of identifying the most relevant supplementary data available from the wide range of existing national registries available in Denmark. We have now begun merging SpineData with relevant opioid data from the NPR and will continue to extract data from this register as well as other relevant Danish national registers up to 2021-2022. We will be looking at data regarding somatic or psychiatric hospitalization patterns, patient usage of health care resources, as well as their working status and disability pensions. We have plans to run a 10-year follow-up study.This survey will be the first to document the scope of long-term problems regarding LTOT and opioid addiction following new spinal pain episodes and comparing opioid usage in PreStarters and NaiveStarters. Usage will be registered from 4 years before the new pain episode and for 5 years afterward. In parallel, we will be able to present continuously descriptive follow-up data from the SPOC.Previously published data indicate that Danes have relatively high utilization of opioids in patient groups experiencing benign pain in the musculoskeletal system ,11. A reThe structure of this study involves both methodological strengths and limitations. One of the absolute strengths in this study is the possible high number of included patients, all of whom experience pain from the same anatomical structures. Additionally, by using the links to national databases during the entire study period, we will be able to obtain a complete dataset for all of the patients regarding their ongoing use of opioids. All of the patients are included from the same regional health system and only one health care entity; in other words, a relatively homogeneous group of patients will be included in the study. A limitation is that we will not be able to gather general data regarding the stoppage of opioid utilization due to side effects or the frequency of side effects.From the long-term collection of relevant individual patient data, it will be possible to illuminate both the overall group data developments as well as long-term developments relating to individual usage of different opioids and to correlate these findings with the individual patient’s physical, psychological, and social data. Results obtained in this study may help to identify a subset of patients with chronic pain for whom long-term opioid use is both safe and effective ,37,38.We expect that SPOC data in the future will contribute research information qualifying the debate about the importance of the clinician undertaking a very thorough patient selection process before prescribing the very first opioid prescription and to provide patients with a solid understanding of the risks of developing addictive behavior and other potential side effects related to LTOT.


To develop a culturally adapted and validated Bengali Pain Catastrophizing Scale (BePCS).The English PCS was translated, adapted and back‐translated into and from Bengali, pre‐tested by 30 adult patients with chronic non‐malignant musculoskeletal pain. The BePCS was administered twice with 14 days interval to 90 patients. Convergent validity was measured by comparing the BePCS score with scores of the domains physical functioning and mental health of the Bengali Short Form 36, through Spearman's correlation coefficient. Test‐retest reliability was assessed by intraclass correlation coefficient (ICC) and Spearman's rank correlation coefficient and internal consistency by Cronbach's alpha. Content validity was assessed by index for content validity (ICV) and floor and ceiling effects.α = 0.92. Internal consistency for subscales rumination, magnification and helplessness, were Cronbach's α 0.903, 0.72 and 0.872 respectively. The test‐retest reliability of total BePCS was 0.78 (P < .001) and for the subscales rumination 0.872 (P < .001), magnification 797 (P < .001) and helplessness 0.927 (P < .001), showing excellent test‐retest reliability.The BePCS was well accepted by the patients in the pre‐test. The content validity was excellent, both item ICV and scale ICV were 1. Construct validity: the convergent validity was −0.424 for physical functioning and −0.413 for mental health, indicating a moderate negative correlation. Total BePCS score showed excellent internal consistency with a mean Cronbach's The interviewer‐administered BePCS appears to be an acceptable, reliable and valid instrument for measuring health‐related quality of life in Bengali speaking patients with chronic non‐malignant musculoskeletal pain. Further evaluation in the general population and in different medical conditions should be done. One of these factors is pain catastrophizing, characterized by patients magnifying their feelings about painful situations and continually thinking about these situations.The Pain Catastrophizing Scale (PCS) was developed in 1995 by Sullivan et al to measure the individual degree of pain catastrophizing. The PCS is a multidimensional questionnaire, consisting of 3 subscales: helplessness, magnification, and rumination. The English version of the PCS has been investigated extensively, and its psychometric properties are good.A culturally adapted and validated Bengali version of the PCS for the people of Bangladesh is not yet available. The purpose of this study is the translation of the PCS into Bengali, cultural adaptation of the Bengali version and to test its validity and reliability in adult patients with chronic non‐malignant musculoskeletal pain.22.1The study has been conducted in the Department of Rheumatology (inpatient and outpatient), of the Bangabandhu Sheikh Mujib Medical University.Consecutive adult female/male patients between 18 and 70 years of age, who visited the rheumatology outpatient and inpatient departments between September 2015 to August 2016, who suffered from chronic non‐malignant musculoskeletal pain (pain persisting ≥ 6 weeks) (10) at the spine or any part of the body and who consented to participate were enrolled in this study. Excluded were severely ill patients, patients with communication problems, patients who had a history of malignant disorders, those who suffered from alcohol or substance abuse and those who had acute pain or needed urgent surgery or other interventions. Substance/alcohol abuse was identified by taking histories and defined as: alcohol/substance used in amounts which are harmful to the individual or others.β = 0.2 (80% power) and α = 0.05, the sample size required was 22.873. Thus for the assessment of the test‐retest reliability of the questionnaire and considering drop‐out of some patients during retesting, a sample size of 32 was considered to be sufficient. These 32 patients were collected by simple random sampling from the 95 patients who were enrolled for the test.The sample size of the study was 95 patients, as calculated by Study Size 3.0, a validated statistical software developed by Creostat HB35 HB in Sweden.2.2The PCS was developed in 1995 at the University Center for Research on Pain and Disability of the McGill University of Canada by Michael JL Sullivan, in order to facilitate research on the mechanisms by which catastrophizing develops and its impact on pain experience.2.3For translation and validation of the Bengali version of PCS, we obtained permission from the original author (MJ Sullivan). For translation and cultural adaptation of the English PCS into Bengali, we followed the recommendations by Beaton et al2.4The 2 prefinal Bengali versions of the PCS were tested in a sample of 30 adult patients with chronic non‐malignant musculoskeletal pain. Each subject completing the questionnaires was interviewed to find out what he or she thought was meant by each questionnaire item, and about the response they gave, and whether they had any further suggestions. If a participant was able to understand both of the translations of the same item, he or she was asked which translation  he or she would prefer. Based on the response of these participants, the adapted version was prepared.The adapted version was administered twice with 14 days interval to 90 Bangladeshi patients who were suffering from chronic non‐malignant musculoskeletal pain.For measuring the physical functioning and mental health, these domains of the Bengali version of the Short Form‐36 (SF‐36) were applied.2.5The questionnaire was used as a self‐administered one for literate participants and an interviewer‐administered one in case of illiterate participants. The literate participants were allowed to read the questionnaire themselves and give the replies as per their own understanding. In case of illiterate participants, the interviewer read the questionnaire in a clearly audible voice, without giving explanation. The responses were recorded by the interviewer.2.6All data were assessed using SPSS 22.0 (SPSS Inc). All tests were 2‐tailed and conducted at a 5% level of significance. There were no missing data for any items. Both the content validity and construct validity were assessed. Reliability was assessed through three ways: internal consistency, test‐retest reliability and item to scale correlation. The internal consistency was measured using Cronbach's alpha. The internal consistency was considered acceptable when Cronbach's alpha was equal to or exceeded 0.70.U test , Wilcoxon rank‐sum test, or Wilcoxon–Mann–Whitney test) was used to compare between 2 groups with respect to a variable that does not follow a normal distribution.The Mann–Whitney The Kruskal‐Wallis test  is a rank‐based nonparametric test. It was used to determine whether there are statistically significant differences between 2 or more groups of an independent variable on a continuous or ordinal dependent variable.The ICC or the intraclass correlation, is a descriptive statistic. It was used when quantitative measurements are made on units that are organized into groups. It describes how strongly units in the same group resemble each other.2.7The Institutional Review Board (IRB) of Bangabandhu Sheikh Mujib Medical University provided clearance to conduct the study (No. BSMMU IRB 11606). All the participants were informed in details about the nature of the study. Only the individuals willing to participate in the study were included. Informed written consent was taken from the participants. Every participant enjoyed his/her right to participate or refuse to participate and to withdraw participation at any time. The principal investigator maintained the confidentiality of the information obtained from the participants. Data were intended to be used solely for this study.33.1P = .971), gender (Table P = .145). Although BePCS scores were lower in people with higher educational levels, the differences were insignificant as per the Kruskal‐Wallis test , 43 male (45.3%) and 52 female (54.7%). The rheumatological diagnoses are summarized in Table 3.01, 43 3.2The I‐CVI and the S‐CVI were the assessment tools of content validity. All items of the scale showed excellent content validity: both I‐CVI and S‐CVI were 1.3.3Convergent validity was measured by comparing the BePCS score with the scores of the physical functioning and mental health domain of the Bengali version of the SF‐363.4α 0.903, 0.72 and 0.872, respectively  and for the subscales rumination 0.872 (P < .001), magnification 797 (P < .001) and helplessness 0.927 (P < .001), indicative of a strong correlation between test and retest scores and hence showing excellent test‐retest reliability  as was done in some other studies A strength of the study is that it is the first study in the Bengal language and it will create opportunities to study this important field of catastrophizing and chronic pain in 265 million Bengali speaking people. Our study showed acceptable validity and excellent internal consistency, construct and content validity and reliability of the Bengali version of the PCS.In conclusion, the BePCS, being a valid and reliable tool, may be used to screen the probability of catastrophizing when suffering from chronic pain. The BePCS can be a valuable tool for patient education, treatment planning and to assess the need for psychological intervention.Bengali PCSClick here for additional data file.