Validation of the International Restless Legs Syndrome Study Group
rating scale for restless legs syndrome
The International Restless Legs Syndrome Study Group
Received 11 June 2002; received in revised form 22 October 2002; accepted 7 December 2002
Background: There is a need for an easily administered instrument which can be applied to all patients with restless legs syndrome (RLS)
to measure disease severity for clinical assessment, research, or therapeutic trials. The pathophysiology of RLS is not clear and no objective
measure so far devised can apply to all patients or accurately reflect severity. Moreover, RLS is primarily a subjective disorder. Therefore, a
subjective scale is at present the optimal instrument to meet this need.
Methods: Twenty centers from six countries participated in an initial reliability and validation study of a rating scale for the severity of
RLS designed by the International RLS study group (IRLSSG). A ten-question scale was developed on the basis of repeated expert evaluation
of potential items. This scale, the IRLSSG rating scale (IRLS), was administered to 196 RLS patients, most on some medication, and 209
Results: The IRLS was found to have high levels of internal consistency, inter-examiner reliability, test–retest reliability over a 2–4 week
period, and convergent validity. It also demonstrated criterion validity when tested against the current criterion of a clinical global impression
and readily discriminated patient from control groups. The scale was dominated by a single severity factor that explained at least 59% of the
pooled item variance.
Conclusions: This scale meets performance criteria for a brief, patient completed instrument that can be used to assess RLS severity for
purposes of clinical assessment, research, or therapeutic trials. It supports afinding thatRLS is a relatively uniformdisorder inwhich the severity
of the basic symptoms is strongly related to their impact on the patient’s life. In future studies, the IRLS should be tested against objective
measures of RLS severity and its sensitivity should be studied as RLS severity is systematically manipulated by therapeutic interventions.
q 2003 Elsevier Science B.V. All rights reserved.
Keywords: Rating scale; Restless legs syndrome; Assessment; Reliability; Validity; Factor analysis; Psycchometrics
Restless legs syndrome (RLS) is a common condition
which may affect as many as 15% of the general adult
population, at least in countries whose populations derive
from Western Europe [1–3]. In recent years, a number of
effective medications have been developed to treat this
condition [4,5]. Because this is both a common and treatable
disorder, it is necessary to have adequate means of assessing
its severity, both for clinical management and to guide the
development of further therapies. Measures of RLS severity
may also be quite useful in epidemiological and pathophy-
In the past, a variety of subjective [6–9] and objective
[10–12] means have been used to evaluate the severity of
RLS and its response to treatment . None of the subjective
instruments have been extensively tested for their psycho-
metric properties or their validity when used to assess
populations of RLS patients. The most common objective
measures – sleep efficiency as calculated from an overnight
sleep study or various indices of periodic limb movements
(PLM) – are well established in their fields as goodmeasures
of specific aspects of sleep. However, they have never been
shown to reliably measure the severity of RLS in all
individual patients. While some patients have major sleep
complaints, others have none. While some patients have
numerous periodic limb movements in sleep (PLMS), other
have few or none (a significant number of PLMS is regarded
as more than five per hour of sleep) [13,14]. In the absence of
validated, universally applicable objective measures, the
criterion for assessment of RLS remains the expert
clinician’s judgment or the clinical global impression
(CGI). However, the CGI may not always be available or
practicable and ameans of assessingRLS severity that can be
1389-9457/03/$ - see front matter q 2003 Elsevier Science B.V. All rights reserved.
Sleep Medicine 4 (2003) 121–132
* Corresponding author. Arthur S. Walters, New Jersey Neuroscience
Institute, JFK Medical Center, 65 James Street, Edison, NJ 08818, USA.
Tel.: þ1-732-321-7000x68177; fax: þ1-732-632-1584.
E-mail address: email@example.com (A.S. Walters).
used by trained, but not necessarily expert, interviewers
would be quite helpful. Such an instrument might also be
modified to be useful for self-assessment by patients. In order
to provide such an instrument, the International RLS Study
Group (IRLSSG) decided to develop a rating scale for
measuring severity (International Restless Legs Scale or
IRLS). Since RLS is a condition defined largely by its
subjective impact, such a subjective rating scale is an
appropriate instrument for examining different degrees of
severity of the disorder. The resulting ten-question instru-
ment was based, in large part, on the consensus clinical
features of RLS as previously delineated by the IRLSSG in
1995  (Table 1).
The scale (Appendix A) reflects both
subjective assessment of the primary features (diagnostic
features 1 through 3 reflected in questions 1 through 3 and 6
of the scale), intensity and frequency of the disorder
(questions 7 and 8 of the scale) and associated sleep problems
(features 5 and 6 reflected in questions 4 and 5 of the scale).
The scale also includes questions which probe the impact of
symptoms on the patients’ mood and daily functioning
(questions 9 and 10 of the scale).
In order to test the psychometric properties of the
scale and to begin assessing its validity, the IRLSSG
initiated an international, multi-center study of the scale.
We now report the results of that study.
Preliminary forms of this rating scale have already been
employed in published therapeutic studies [16,17]. The
current version of the rating scale was utilized in a large
multi-center and multi-national study of pergolide (Permax)
in RLS, which has been reported in abstract form .
Preliminary results from the current study have also been
reported in abstract form .
2.1. Development of the rating scale
The rating scale was developed on the basis of questions
proposed by members of the IRLSSG, who possess clinical
expertise with this condition (see list of contributors in
Appendix B). Numerous members of the group then
subjected the scale to several rounds of refinement with
commentary. An attempt was made to establish content
validity by having this large panel of RLS experts ensure
that no significant aspect of RLS was omitted from the scale.
This was balanced by the need to generate a scale
sufficiently brief to permit use in a clinical or interview
setting. Preliminary versions of the scale varied between 28
and six questions. The final scale is ten questions in length.
The number of questions was reduced by the decision not to
use questions in multiple formats to redundantly probe the
same aspects of the disorder. It was determined that all
questions should have a similar format and a similar
polarity. Each question had a set of five response options
graded from no RLS or impact (score ¼ 0) to very severe
RLS or impact (score ¼ 4). This produced a total scale
whose overall score could range from 0 to 40. During the
development process, the scale was expanded to include all
critical aspects of RLS designated by the expert group. The
period of development took 18 months. Besides those who
participated in the actual trial, many other centers and
individuals contributed to the formation of the scale (see list
of contributors in Appendix B). The final scale is
reproduced in Appendix A.
2.2. Centers and subjects
A total of 20 centers from six countries (Germany,
Ireland, Italy, Spain, Sweden, and the United States) were
included in the study. RLS patients were recruited from 17
centers and control subjects from 14 centers. The total
number of valid subjects recruited broken down into patient
and control subjects, overall and by country, is shown in
Table 2 together with their demographic information.
Overall, there were 405 subjects, 196 RLS patients and
2.3. Subject recruitment: inclusion and exclusion criteria
RLS patients were required to have a diagnosis of RLS
according to IRLSSG criteria . Diagnoses were made by
members of the IRLSSG involved in the study. Controls and
patients were excluded if they could not complete the
questionnaire for any reason, e.g. dementia or aphasia.
Controls were excluded if they met the criteria for RLS or
Features of RLS by IRLSSG consensus
(1) A desire to move the extremities usually associated with some
(2) Motor restlessness
(3) Worsening of symptoms at rest with at least temporary relief by
(4) Worsening of symptoms later in the day or at night
(5) Involuntary movements awake and asleep (PLM)
(6) Sleep disturbance and its consequences
(7) Normal neurological examination in idiopathic cases
(8) Variable age of onset with typical chronic, progressive course
(9) Frequent familial history of cases
These features were recently refined on the basis of a consensus
conference on RLS diagnosis held May 1–3, 2002 at the NIH.
These criteria have recently been refined on the basis of a consensus
conference held at the NIH (May 1–3, 2002). See this issue of Sleep
Medicine. ‘Allen RP, HeningWA, Montplaisir J, Picchietti D, Trenkwalder
C, Watters AS. Restless Legs Syndrome: Diagnostic criteria, special
considerations and epidemiology’. The new criteria delete criterion 2 on
motor restlessness (Table 1), because it has been found to be confusing. In
addition, criterion 3 is split into two separate criteria: provocation at rest
and relief with activity. These changes should not have any impact on the
design of the IRLS.
The International Restless Legs Syndrome Study Group / Sleep Medicine 4 (2003) 121–132122
had a history of neuroleptic exposure, neuroleptic-induced
akathisia, peripheral neuropathy, radiculopathy or any other
condition that could be confused with RLS. Patients with
RLS were excluded if they had a history of neuroleptic
exposure or neuroleptic-induced akathisia, but were not
excluded if they had ‘secondary’ forms of RLS, i.e. RLS
associated with peripheral neuropathy or radiculopathy. The
control subjects were broken down into two groups: those
with known or clinically suspected sleep disorders (N ¼ 99)
and those drawn from a normal population without known
or suspected sleep disorders (N ¼ 110).
2.4. Testing protocol
Prior to the study patients were asked to remain on stable
dosages of RLS medications and any other medications
known to affect the severity of RLS symptoms for 1 month
prior to day 1 (first administrations of the rating scales) and
for the 2 week interval between the two administrations of
the rating scale. All records were reviewed to ensure that
these conditions were met. In some cases, it was necessary
for patients to have their medications changed or follow-up
at the 2 week interval was not possible. We excluded
patients from test–retest evaluations if their medications
changed or the interval between the two tests was less than
12 or greater than 30 days.
On each of the testing days, patients were asked to rate
themselves twice on the ten-question rating scale (see
Appendix A) in the presence of different examiners. This
duplicate rating was performed in order to determine
whether differences in the responses, help, or instructions
of the examiner might influence ratings. Each examiner was
available throughout the entire time the patient was filling
out the rating scale in order to explain the rating scale and to
clarify any misunderstandings the patients might have
regarding the questions on the scale. The protocol dictated
two examiners at the first administration, but a second
examiner was optional at the second administration. If two
examiners were used, each remained blind to the answers
given by the patient to the other examiner. The patients were
also asked to give each examiner an overall rating of the
severity of their symptoms over the course of the previous 2
weeks, ranging from 0 (no symptoms) to 8 (most severe)
(patient global impression rating, PGI). A third, expert
examiner was also asked to conduct a general analysis of
patient symptoms and severity and to generate his or her
own CGI of the severity of the patient’s symptoms. This was
also scored on a scale from 0 (no symptoms) to 8 (most
severe). The CGI was required by the protocol on the first
day, but was optional for the second administration. This
third examiner was also required to be blind to all answers
given to the first two examiners by the patient. The
coordinating center audited all records to be sure that
these conditions were met. In some cases, where the
protocol was not followed exactly (e.g. as to rater blinds),
those scores were not used in analyses that required
independent scores. In other cases, not every rating was
completed. In that case, the patient was not included in
analyses requiring, for example, two PGI scores. In all cases
where patients were used for results related to the ten-
question scale, answers to every question were available.
Controls had only a single administration of the ten-
question rating scale on day 1 and were not asked to re-do
the rating scale on another date. No PGIs or CGIs were
generated for the controls.
2.5. Statistical analyses
For construct validity, we performed a factor analysis
and examined item convergent validity. Prior to the factor
extraction, we examined the dataset for the Kaiser–Meyer–
Olkin (KMO) value to see if the dataset supported valid
factor extraction. As a general rule, a KMO value greater
than 0.6 is considered adequate for extraction and a KMO
value greater than 0.9 is considered excellent [20,21]. The
factor analysis was first performed on the average ratings
obtained from the first administration (N ¼ 196). To avoid
spurious assignments of variance, we selected a principal
Demographics of patient and control groups
Patients All controls Normal controls Sleep disorder controls
Total no. 196 209 110 99
US 89 103 48 55
Germany 41 32 22 10
Italy 35 49 27 22
Spain 10 12 0 12
Sweden 5 0 0 0
Ireland 16 13 13 0
Mean (SD) 61.8 (11.9) 56.1 (15.0) 58.4 (15.0) 53.5 (14.7)
Range 34–90 22–91 39–91 22–88
Sex, n (%)
Male 70 (36) 109 (52) 52 (47) 57 (58)
Female 126 (64) 100 (48) 58 (53) 42 (42)
The International Restless Legs Syndrome Study Group / Sleep Medicine 4 (2003) 121–132 123
factor extraction using the Kaiser criterion of accepting only
those factors with an eigenvalue greater than 1 and also
evaluating the factors with a scree plot, including use of the
objective scree test . In order to confirm the validity of
this choice, we also explored other factor solutions
stipulating multiple factor solutions using both varimax
orthogonal and oblimin oblique solutions.
The factor results obtained from the first administration
scores were then compared with those obtained from the
second administration (N ¼ 187) using the same pro-
cedures. First, we correlated the factor loads derived from
the two sets of scores. Then, we calculated factor scores for
the second administration using the factor score coefficient
matrix generated for the first administration. We then
correlated those scores with the factor scores we extracted
directly from the second administration to determine the
similarity of the separate factors extracted from the two
administrations. We accepted as related to the factor all
those questions which loaded at levels greater than 0.4.
Our reliability analysis consisted of examination of
internal consistency, inter-examiner reliability, and test–
retest reliability. For internal consistency, we performed a
Cronbach alpha analysis . We used a criterion of 0.7 to
indicate adequate internal consistency [23,24]. For inter-
examiner reliability, we used an intra-class correlation
coefficient (ICC) equivalent to a weighted kappa analysis
. We have designated this an inter-examiner reliability
test since the patients themselves provided the ratings, but
did so in the presence of different examiners who might
conduct the testing differently, provide different infor-
mation, or simply influence the patients in different ways.
We used the same statistic to compute a test–retest
reliability. We used a criterion of an ICC of 0.7 as indicating
a satisfactory performance . We also compared scores
between the first and second administrations using a paired
t-test. Our hypothesis was that there would be no significant
difference between the scores at the two time points. We set
the significance level at 0.05.
Validity analysis consisted of criterion validity, con-
current validity, and discriminant validity. For criterion
validity, we regressed the IRLS scores against the CGI. For
concurrent validity we regressed the IRLS scores against the
PG1. For discriminant validity, we performed a one-way
ANOVA with three groups, patients and two types of
controls (sleep disorder and normal controls). We then did
post-hoc t-tests (Scheffe´, Bonferroni) to locate any signifi-
cant differences. Our hypothesis was that the ANOVA
would show significant group differences and that the
patients would be significantly different from either control
group, but that the control groups would not differ from each
Because differences between the raters were so low (see
Section 3.2.2), we averaged two ratings of a subject (IRLS
scores, PGI) for all other analyses of the scores, where two
ratings were available.
3.1. Construct validity
3.1.1. Factor analysis
Because the KMO value was in the excellent range
(0.908), we felt justified to proceed with the factor analysis
of the first administration scores. Only one factor had an
eigenvalue greater than 1 (6.28) using a principal factor
extraction on the dataset (N ¼ 196). The eigenvalue for the
next factor was 0.88. The scree plot showed a clear break at
the second factor. We therefore accepted a one factor
solution which accounted for 59.2% of the variance. We
therefore call this a general severity factor. All items except
question 3 had factor loads in excess of 0.7 (Table 3).
Further exploration of multiple factor solutions indicated
that, with rotation, there emerged two separate factors with
primary loading on symptom measures (questions 1, 2, 4, 6,
7, and 8) and disease impact measures (questions 5, 9, and
10). However, there was considerable overlap between
factors with variables contributing to one factor having
significant contributions to the other factors (loadings.0.4).
It was therefore concluded that the scale was truly unified
around one very strong factor and that the addition of another
factor only partially teased apart highly related variables. The
exploratory analysis also indicated that the two sleep items
(questions 4 and 5) and the two symptom prevalence
measures (questions 7 and 8) were highly related to each
other. Question 3 did not load well on either of the factors.
For the second administration (N ¼ 187; nine subjects lost
to follow-up between administrations, KMO ¼ 0:920), the
general severity factor was also seen with an eigenvalue of
6.88 and accounting for 65.0% of the variance. As in the first
administration, all items except question 3 had factor loads in
excess of 0.7 (Table 3). Further analysis also showed that a
two factor solution broke down into two factors representing
symptom measures and disease impact measures with much
overlap. The most distinct items for each factor (symptom
measures: questions1, 2, and6; impactmeasures, questions5,
Loadings on general severity factor for two administrations
Question number Administration
1 0.872 0.897
2 0.821 0.845
3 0.444 0.572
4 0.799 0.809
5 0.738 0.810
6 0.924 0.931
7 0.723 0.719
8 0.809 0.851
9 0.739 0.831
10 0.726 0.776
Single factor solution – principal factor extraction. Loads for two
administrations correlate 0.961 (P , 0:001).
The International Restless Legs Syndrome Study Group / Sleep Medicine 4 (2003) 121–132124
9, and 10) were common to the two administrations, but even
these showedmore than deminimus loads on the other factor
(all .0.29). The overall similarity of the weightings for the
two administration factor extractions can be seen from the
tabulated weightings (Table 3). In fact, the weights correlate
0.961 (P , 0:001).
We also further explored the similarity of the factors
extracted from the two administrations. We calculated
factor scores using data from the second administration with
the factor score coefficient matrix from the first adminis-
tration. The resulting scores correlated 0.982 (P , 0:001)
with the factor scores directly extracted from the second
administration, indicating an almost complete similarity of
the factors extracted from the two administrations. This
finding indicates that there is very little difference in the
factor structure of the scores on the two administrations.
3.1.2. Item convergent validity
Correlations between individual items (questions 1
through 10) and the total score of the questionnaire
(minus that item) were always significant and positive.
Except for question 3 correlations for the individual items
with the total score varied between 0.69 and 0.90 (day 1,
N ¼ 196; day 2, N ¼ 187). Items 1 and 6 had the highest
correlations on both days (day 1, 0.83, 0.88; day 2, 0.87,
0.90) (P , 0:001), while item 3 (response to movement)
had the lowest (day 1, 0.43; day 2, 0.56) (P , 0:01). It is
usually accepted that item convergent validities above 0.4
are acceptable for rating scales .
3.2. Reliability analyses
3.2.1. Internal consistency
Cronbach alpha measures for the two administrations
were 0.93 (N ¼ 196) and 0.95 (N ¼ 187), respectively
(P , 0:001). There was minimal change in this value when
each question was selectively removed. The only question
whose exclusion increased the alpha value was question 3,
concerning relief with walking.
3.2.2. Inter-examiner reliability
Inter-examiner reliability was measured by ICC, equiv-
alent to a weighted kappa analysis . Subjects were
accepted into this analysis if they had two ratings which
could be ordered in such a way that the scores for all
subjects could be divided into two distinct sets of raters (that
is, the same rater was not found in both sets of scores). For
the summed rating scale, the ICC was 0.93 for the first
administration (n ¼ 187 patients) and 0.97 for the second
administration (n ¼ 169 patients) (P , 0:001). Considering
the individual questions, ICCs for the individual questions
ranged between 0.68 and 0.93 for the first administration
and between 0.78 and 0.96 for the second (all P , 0:01).
For both days, the lowest reliability was for question 3
(relief with walking) and the highest was for question 7
(frequency of symptoms, days per week). For the PGI, the
ICCs were 0.95 (N ¼ 155) and 0.94 (N ¼ 130) respectively
(P , 0:001).
Because of the extremely high inter-rater reliability, the
two scores (either on the rating scale or PGI) were combined
for further analyses. In cases where only a single score was
available, that single score was used.
3.2.3. Test–retest reliability
A total of 145 patients met criteria for our test–retest
evaluation within the 12–30 day window. Of the remaining
subjects, 15 returned too soon for retest (2–11 days), five
returned too late (34–104 days), nine did not return at all, and
22 were not on constant medications. For those who met the
criteria, the mean period between testing was 15.0 days (SD
4.1). The ICC for these patients’ scoreswas0.87 (P , 0:001).
Mean scores went from 21.37 (SD 8.35) to 20.88 (SD 8.89)
between the two administrations. This difference was not
significant (t ¼ 1:35, d:f: ¼ 144, paired t-test, P . 0:05).
The mean difference in scores was20.49 (SD 4.37).
3.3. Validity analyses
3.3.1. Criterion validity
We assessed criterion validity by determining the
correlation of the summed questionnaire score to the CGI.
This yielded an r value of 0.74 on day 1 and 0.73 on day 2
(P , 0:001). The relationship for both days is plotted in Fig.
1. It is evident from these graphs that the intercepts are near
zero and that there is a tendency for the most extreme values
(low or high) of either measurement to be associated with
more moderate values of the other.
To examine this relationship further, we performed an
ANOVA after dividing the subjects into groups based on
their IRLS score, using a priori designations that mirrored
the proposed designations for levels of CGI (mild: scores
from 0 to 10; moderate: 11 to 20; severe: 21 to 30; very
severe: 31 to 40). An ANOVA using these groups with the
CGI as the dependent variable found that the F values for
the two administrations were 57.42 and 47.90, respectively
(P , 0:001), for those subjects with an independently rated
CGI. Post-hoc tests all revealed that for the first adminis-
tration there was a significant difference in CGI scores for
all comparisons of the four IRLS levels, while for the
second administration the same was true except that the
difference between the CGI scores for the severe and very
severe IRLS levels was not significant. Scores are given in
Table 4 for those subjects with both tests available and
independent CGI. As can be seen, the proposed severity
levels for the IRLS summed score in general correspond to
the same anchored levels of the CGI, except that the mean
CGI score for the very severe IRLS level falls into the
severe range rather than the very severe range.
The mean CGI for all subjects independently rated
(N ¼ 182, 153) was 4.03 on day 1 and 3.84 on day 2, also
near the middle of the scale of 4. Standard deviations were
1.98 and 2.04, respectively. The distribution of scores on the
The International Restless Legs Syndrome Study Group / Sleep Medicine 4 (2003) 121–132 125
CGI is given in Table 5. There was a fairly even distribution
of scores between 1 and 7 while few patients were scored as
8, most severe. A couple of patients who were without
current complaint were scored as 0.
3.3.2. Concurrent validity
Concurrent validity was also examined by comparing the
IRLS summed score to the PGI. This correlation was 0.82 on
day 1 and 0.78 on day 2 (for both, P , 0:001). In an
additional analysis, we correlated the PGI on the CGI. This
yielded an r value of 0.80 on day 1 and 0.84 on day 2
(P , 0:001).
3.3.3. Discriminant validity
The majority of the controls had scores of zero on the
IRLS. For those controls recruited in centers also recruiting
patients, there were four non-zero scores in controls drawn
from the normal population (normal controls, 4/105 non-zero
or 4%) and 13 non-zero scores in controls with a sleep
disorder (sleep disorder controls, 13/71 or 18%). Overall, 17
of 176 controls from these centers had non-zero scores
(9.7%). Five additional normal controls and 28 sleep disorder
controls were recruited at centers that did not recruit patients:
all of these controls had zero scores. For either control group
or all controls, the median and mode were both zero.
A one-way ANOVA found a highly significant F ratio for
the main factor of group (F ¼ 577:0, d:f: ¼ 2, P , 0:01).
Post-hoc tests between the patients and both the normal
control subjects and the sleep disordered control subjects
showed that the patients had significantly higher scores than
either group of controls (both P , 0:001), but that there was
no significant difference between the two control groups.
3.4. Distributions of scores
3.4.1. Total rating scale score
The mean averaged summed scores for the two
Fig. 1. The ratings assigned by the independent expert raters (CGI) are
plotted against the averaged rating scale summed scores for individual
subjects for the first (A) and second (B) administrations of the rating scale
for all cases with independent ratings. RS/1, averaged rating scale sum, first
administration; RS/2, second administration; CGI/1, clinical global rating
first administration; CGI/2, second administration.
Distribution of CGI scores by proposed severity levels of IRLS
IRLS level N Mean CGI SD CGI
Mild 22 1.77 1.27
Moderate 58 3.07 1.46
Severe 81 4.81 1.61
Very severe 29 6.07 1.46
Mild 29 1.64 1.47
Moderate 54 3.30 1.38
Severe 56 5.06 1.68
Very severe 22 5.82 1.44
IRLS, IRLS summed score levels: mild, 0–10; moderate, 11–20;
severe, 21–30; very severe, 31–40. CGI – clinical global impression: 0,
asymptomatic; 1–2, mild; 3–4, moderate; 5–6, severe; 7–8, very severe.
Distribution of CGI scores
CGI level Administration
Total 182 153
The International Restless Legs Syndrome Study Group / Sleep Medicine 4 (2003) 121–132126
administrations for all subjects were 21.91 and 20.27, near
the center of the possible range of scores. Standard
deviations were 8.39 and 9.24, respectively, and the full
range for both administrations was from 0 to 38.
Distributions of scores for the RLS patients are plotted in
Fig. 2. It is apparent that the scores are rather evenly
distributed from 12 to 32 with a longer tail towards the
lower scores. Median values for the two administrations
were 23.25 and 20.5, respectively, while modal values for
scores grouped into intervals of 4 were in the intervals 24–
27 and 28–31, respectively (Fig. 2).
3.4.2. Global impressions
The mean PGIs for all subjects with at least one rating
(N ¼ 188, 179) were 4.27 and 4.12, near the middle score of
4. Standard deviations were 2.01 and 2.08, respectively.
4.1. Summary of results
All the reliability and validity analyses revealed highly
significant results that met or exceeded minimum quality
standards for an instrument of this kind. Internal consistency
revealed that, with the possible exception of question 3, this
scale was very highly unified, a conclusion supported by the
emergence of a single strong factor with highly significant
loadings from each question except 3. This factor can be
termed a severity factor, and notably draws strong support not
only from primary measures of symptom severity (questions
1, 2, and 6) and intensity/frequency (questions 7 and 8) but
also from those which related to impact on sleep (questions 4
and 5) and impact on mood and daily functions (questions 9
and 10). Within each group of questions there were tighter
relations than across groups, but the overlap among groups
was so high that the one factor solution was the optimal one.
Similar trendsweremanifest in the high degree of convergent
validity found. Such a result argues that RLS may be a
relatively unified condition in which the severity of
diagnostic symptoms largely determines the impact on the
patients. This conclusion is also supported by the indication
in circadian studies that all features of the condition tend to
co-vary with the circadian cycle [29,30] and that, in
therapeutic trials, subjective and objective measures usually
indicate a similar result . This conclusion will need to be
explored in other studies. The degree of consistency observed
in the rating scale, however,maybe greater than optimal .
As a result, it may be feasible either to reduce the number of
questions or to add items that deal with other aspects of RLS
symptoms and impact.
Inter-examiner reliability was very high (0.93 and 0.97)
and suggested that there should be no difficulty in having this
scale administered by diverse raters. Test–retest reliability
measured on patientswith a consistentmedication profile and
at intervals up to 30 days revealed that the rating scale scores
Fig. 2. The averaged rating scale summed scores are displayed in 4 point bins for the first and second administrations. The data include scores from all RLS
patients with valid scores (N ¼ 196, first administration; N ¼ 187, second administration).
The International Restless Legs Syndrome Study Group / Sleep Medicine 4 (2003) 121–132 127
were quite stable over this time period. This was also true of
the CGI, currently the most generally accepted means of
assessing RLS severity.
In criterion validity, the rating scale score performed
well in comparison to the criterion of the CGI, with
correlations of 0.74 and 0.73 (P , 0:001) (Fig. 1). In
discriminant validity, it was clear that control groups, even
those with sleep disorders that might cause overlapping
symptoms, had mostly zero scores, dramatically different
from the RLS patients. This result is probably explained in
large part by the fact that all the questions were anchored by
a reference to RLS.
The analysis of the rating scale’s distribution showed that
it hadmost scores in a range that corresponded tomoderate to
severe CGI ratings (Fig. 1). The scale scores showed a good
spread of values from zero through the highest values of the
scale (.30). This should allow for adequate discrimination
of patients with a wide range of severities.
4.2. Utility of scale
Because the scale is brief and apparently posed few
problems to any of the patients, it offers the possibility of
ready use in clinical practice, in epidemiological and
pathophysiological research, and in clinical trials. It is
notable that, under the conditions of administration, all but
one questionnaire was completed with all questions
answered. However, since the current study had the scale
completed in the presence of a knowledgeable professional,
it is not clear from these results that the scale, if administered
by a telephone canvasser or in mailed questionnaires, would
perform equally well without such a professional being
available. However, the very high inter-examiner reliability
and strong test–retest stability suggest that it will be useful in
situations where scores need to be obtained by diverse
individuals. The results also suggest that changes in the scale
are likely to reflect true changes in the underlying condition.
Because it was well correlated to CGI, the scale is validated
as a satisfactory instrument for usewithout contribution from
a sustained, expert clinical interview.
Since the patients examined had by and large been
treated and were on a variety of medications, the scale
should be useful for assessment not only of untreated
patients, but of those who are on different medication
regimens. This study was conducted on the typical patient
populations regularly seen in RLS centers.
4.3. Comparison to other measures
This rating scale has an advantage over other measures
since it has been subjected to intensive evaluation of its
reliability and aspects of validity. Unlike objective measures
of RLS, it can be easily and effectively applied to all patients.
However, it does not examine all aspects of RLS. The Johns
Hopkins RLS severity scale (JHRLSS) takes a different
approach, examining severity by time of day of onset of
symptoms. That scale has been validated against objective
measures of RLS such as sleep efficiency and PLM index
. It has also proven useful in correlating severity to
biologicalmeasures such as serumferritin  or brain iron in
the substantia nigra . However, that scale, while
complementing the IRLS, does not cover as many aspects
of the RLS condition. Other areas not covered by either scale
include the number of involved limbs or the rapidity with
which symptoms develop when a patient first sits or lies
4.4. Future requirements and prospects
In further work, it will be necessary to establish the
relationship of the IRLS to such objective measures of RLS
as sleep efficiency and PLM indices. Some of this work is
currently under way: in a large parallel double blind
placebo/drug trial, changes in the scale were found to be
significantly related to changes in PLM indices, sleep
efficiency, and CGI (Trenkwalder, personal communi-
cation). This suggests that the scale is sensitive to changes
in or manipulations of the severity of RLS as is expected in
clinical trials. The scale should be able to discriminate
between different levels of RLS severity at different time
points within the same individual.
We have considered whether elimination of question 3
(relief with walking) from the scale would benefit its
psychometric properties. While the scale more than meets
all performance standards with this question included, it is
the one question that repeatedly stands out as less related to
the overall scale or the remainder of the scale items. It does
not contribute at a high level to the main factor, or even at
lower levels to any multiple factor rotated solution. This
may be due to its answers having a somewhat different
format (Appendix A) or to the fact that almost all patients
experience significant relief with walking, a possibility
supported by the low mean scores for this question and the
minimal standard deviation. However, several factors
mitigate against removing the item from the scale. First,
it measures relief with walking, a key diagnostic feature of
RLS. Second, it meets threshold standards for good
performance. Third, the scale, as presently constructed, is
more coherent than recommended . If question 3 were
eliminated, this coherence would only increase. The
authors are aware of ongoing studies using the scale in
other contexts, and if question 3 is repeatedly found to
suffer from these deficits, particularly an insensitivity to
changed clinical status within therapeutic trials, future
editions may decide that it should be eliminated.
Another question for future study is whether either a
shorter or a longer scale would be equally useful. By
eliminating some questions, a shorter scale with comparable
psychometric properties but a lesser degree of coherence
might be achieved. A longer scale could incorporate
additional aspects of the condition (such as time of day of
symptom onset or rapidity of symptom development at rest)
The International Restless Legs Syndrome Study Group / Sleep Medicine 4 (2003) 121–132128
and include additional quality of life measures. Such an
extended scale might better capture disease impact for
purposes of clinical evaluation or measurement of thera-
peutic response. It is also possible that in the future a more
complex scale or different scales aimed at different aspects
of RLS (e.g. symptom severity versus disease impact or
quality of life) will prove more useful for different
The scale is also flexible and can be used, with minor
modification, for either shorter (e.g. 1 week) or longer (e.g.
1 month) periods of assessment. While the IRLS was
administered with an examiner present in this study, in the
future it may prove possible to have patients fill out the
scales by themselves or be queried over the telephone to
expand the contexts in which the scale may be used.
We wish to thank Mark Atkinson, Linda Hirsch, David
Streiner, and Barbara Tabachnick for providing statistical
consultations during the analysis of data and development of
Appendix A. IRLSSG restless legs syndrome rating scale
for severity (IRLSSGRS)
This scale is copyrighted by the Interntional Restless
Legs Syndrome Study Group 2002 and this version IS NOT
TO BE USED OR DISTRIBUTED. A slightly modified
version of the scale that is re-worded for better clarity is
presented in an accompanying editorial in this issue of Sleep
Medicine. The English version of the modified scale and
translation into other languages can be obtained through
“Caroline Anfray, Information Resources Centre, MAPI
Research Institute, 27 rue de la Villette, 69003 Lyon,
France. Phone þ 33(0) 472 13 66 67. FAX þ 33 (0) 472 13
66 82. E-mail firstname.lastname@example.org or email@example.com”.
Rate your symptoms for the following ten questions.
Unless otherwise instructed, you should rate the average
symptoms that you have experienced for the most recent
two week period.
(1) Overall, how would you rate the
RLS discomfort in your legs or arms?
(4) Very severe
(2) Overall, how would you rate the need to move
around because of your RLS symptoms?
(4) Very severe
(3) Overall, how much relief of your RLS arm or leg
discomfort do you get from moving around?
(4) No relief
(3) Slight relief
(2) Moderate relief
(1) Either complete or almost complete relief
(0) No RLS symptoms and therefore question does not
(4) Overall, how severe is your sleep disturbance from
your RLS symptoms?
(4) Very severe
(5) How severe is your tiredness or sleepiness from your
(4) Very severe
(6) Overall, how severe is your RLS as a whole?
(4) Very severe
(7) How often do you get RLS symptoms?
(4) Very severe (This means 6 to 7 days a week)
(3) Severe (This means 4 to 5 days a week)
(2) Moderate (This means 2 to 3 days a week)
(1) Mild (This means 1 day a week or less)
(8) When you have RLS symptoms how severe are they
on an average day?
(4) Very severe (This means 8 hours per 24 hour day or
(3) Severe (This means 3 to 8 hours per 24 hour day)
(2) Moderate (This means 1 to 3 hours per 24 hour day)
The International Restless Legs Syndrome Study Group / Sleep Medicine 4 (2003) 121–132 129
(1) Mild (This means less than 1 hour per 24 hour day)
(9) Overall, how severe is the impact of your RLS
symptoms on your ability to carry out your daily affairs, for
example carrying out a satisfactory family, home, social,
school or work life?
(4) Very severe
(10) How severe is your mood disturbance from your
RLS symptoms – for example angry, depressed, sad,
anxious or irritable?
(4) Very severe
B.1. Writing, central data collection and data analysis
New Jersey Neuroscience Institute at JFK Medical
Center, Edison, NJ: Arthur S. Walters, MD, Cheryl
LeBrocq, Anjana Dhar, MD; UMDNJ-Robert Wood
Johnson Medical School, New Brunswick, NJ: Arthur
S. Walters, MD, Wayne Hening, MD, PhD, Ray Rosen,
PhD; Seton Hall University School of Graduate Medical
Education, South Orange, NJ: Arthur S. Walters, MD;
Department of Neurology, Johns Hopkins University,
Baltimore, MD: Wayne Hening, MD, PhD, Richard
P. Allen, PhD; Center for Molecular and Behavioral
Neuroscience, Rutgers University, Piscataway, NJ: Wayne
Hening, MD, PhD; Max Planck Institute of Psychiatry,
Munich, Germany: Claudia Trenkwalder, MD; Department
of Clinical Neurophysiology, University of Goettingen,
Goettingen, Germany: Claudia Trenkwalder, MD.
B.2. Members of the group and other contributors
Parkinson’s Disease and Movement Disorders Center,
Mayo Clinic, Scottsdale, AZ: Charles Adler*, Stephanie
Newman, Cynthia Reiners.
Department of Neurology, Erciyes University Medical
Faculty, Kayseri, Turkey: Murat Aksu.
Department of Neurology, Johns Hopkins University,
Baltimore, MD: Richard P. Allen*, David Buchholz*,
Wayne A. Hening*.
Sleep Disorders Center, St. Joseph Hospital, Orange, CA:
Melanie Anderson, Sarah Mosko*.
Department of Psychiatry, University of California, San
Diego, CA: Sonia Ancoli-Israel*.
National Institute of Neurological Disease and Stroke,
Bethesda, MD: William Bara Jimenez*, Mark Hallett*.
Neurologische Poliklinik, Universitatsspital Zurich, Zurich,
Switzerland: Claudio Bassetti*, Sandra Clavadetscher.
Department of Neurology, Emory University, Atlanta, GA:
Donald L. Bliwise*, Paul Gurecki, David B. Rye*.
SleepWake Disorders Center of the New York-Presbyterian
Hospital, New York, NY: Lauren L. Broch, Rochelle Zak*.
St. Vincent’s Medical Center, New York Medical College,
New York, NY: Sudhansu Chokroverty*.
Department of Neurological Sciences, University of
Bologna, Bologna, Italy: Giorgio Coccagna*, Elio
Lugaresi*, Filomena Miele, Pasquale Montagna*, Giuseppe
Plazzi*, Federica Provini*.
Psicobiologia Universidade Federal de Sao Paulo, Sao
Paulo, Brazil: Marco Tulio de Mello*, Sergio Tufik.
Centre for Sleep and Wake Disorders, Westeinde Hospital,
The Hague, The Netherlands: Al W. de Weerd*, Roselyne
New Jersey Neuroscience Institute at JFK Medical Center,
Edison, NJ: Anjana Dhar, Cheryl LeBrocq, Arthur
Department of Neurology, Tufts New England Medical
Center, Boston, MA: Bruce Ehrenberg*.
Department of Neurology, Ludwig Maximilians
Universitat, Munich, Germany: Ilonka Eisensehr*.
Department of Neurology, Huddinge University Hospital,
Huddinge, Sweden: Karl Ekbom Jr.*, Ake Ljungdahl.
Department of Neurology, Fundacion Jimenez Diaz,
Madrid, Spain: Diego Garcia-Borreguero*, Oscar Larrosa.
UMDNJ-Robert Wood Johnson Medical School, New
Brunswick, NJ: Wayne A. Hening*, Ray Rosen*, Arthur
Center for Molecular and Behavioral Neuroscience, Rutgers
University, Piscataway, NJ: Wayne A. Hening*, Linda
Universitatsklinik fur Neurologie, Innsbruck, Austria: Birgit
The International Restless Legs Syndrome Study Group / Sleep Medicine 4 (2003) 121–132130
Department of Psychiatry, Shimane Medical University,
Izumo City, Japan: Jun Horiguchi*.
Department of Psychiatry and Psychotherapy, Albert-
Ludwigs-University, Freiburg, Germany: Magdolna
Hornyak*, Ulrich Voderholzer*.
Sleep Disorders Center, St. Boniface Hospital-Research
Center, Winnipeg, Manitoba, Canada: Meir Kryger*,
Clinical Neuroscience Research Foundation, Concord, MA:
Joseph F. Lipinski*.
Department of Pulmonary and Critical Care Medicine,
University of Kentucky Medical Center, Lexington, KY:
Ahmed Masood, Barbara Phillips*.
Department of Neurology, Philipps University, Marburg,
Germany: Wolfgang H. Oertel*, Karin Stiasny*.
St. Michael’s Hospital, Dublin, Ireland: Shaun O’Keeffe*.
Sleep Disorders Center, San Raffaele Scientific Institute,
Milan, Italy: Alessandro Oldani, Marco Zucconi*.
Department of Neurology, Baylor College of Medicine,
Houston, TX: William G. Ondo*.
Carle Clinic, University of Illinois, Champaign-Urbana, IL:
Division of Neurology, Scripps Clinic, La Jolla, CA:
J. Steven Poceta*.
Department of Neurology, Pacific Sleep Program, Portland,
OR: Gerald B. Rich*.
Sleep Alertness Center, Aurora, CO: Larry Scrima*.
San Diego Sleep Disorders Center, San Diego, CA: Renata
Tulane University Hospital and Clinic, New Orleans, LA:
Mayo Clinic, Rochester, MN: Michael Silber*.
Department of Medicine, Michigan State University, East
Lansing, MI: Robert Smith*.
Max Planck Institute of Psychiatry, Munich, Germany:
Claudia Trenkwalder*, Thomas C. Wetter*, Juliane
Department of Clinical Neurophysiology, University of
Goettingen, Goettingen, Germany: Claudia Trenkwalder*.
Department of Neurology, UCLA School of Medicine, Los
Angeles, CA: Zeba Vanek.
Department of Pharmacy Practice, Rutgers University,
Piscataway, NJ: Mary Wagner.
(continued on next page)
Seton Hall University School of Graduate Medical
Education, South Orange, NJ: Arthur S. Walters.
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